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Federal
Reserve Bank of
New York
Quarterly Review
A utum n 1991
1
Volum e 16 N um ber 3
Do In te rn a tio n a l R eactions of S tock and Bond
M arkets R eflect M a cro e co no m ic F u n d a m e n ta ls?
14
The C ost of C a p ita l fo r S e cu ritie s
Firm s in the U nited S tates and
Japan
28
F inancial L ib e ra liza tio n and
M o n e ta ry C ontrol in Japan
47
E xpected Inflation and Real
Interest R ates Based on Indexlinked Bond P rices: The U.K.
E xperience
61
Tracking the E conom y w ith the
P urchasing M anagers’ Index
70
Treasury and Federal R eserve
Foreign E xchange O p e ra tio n s
Federal Reserve Bank of New York
Quarterly Review
Autumn 1991 Volume 16 Number 3
Table of Contents
1
Do International Reactions of Stock and Bond Markets Reflect
Macroeconomic Fundamentals?
Eli M. Remolona
Using quarterly data for the United States, Japan, and the United Kingdom,
this article investigates the relationship between changes in fundamental
economic conditions and the international reactions of stock and bond
markets. The author assesses the degree to which the transmission of
information about fundamentals prompts the stock or bond markets in
different countries to move together.
14
The Cost of Capital for Securities Firms in the United States and
Japan
R obert N. M cCauley and Steven A. Zim m er
The authors use stock market valuations to construct estimates of the cost
of capital for five U.S. and four Japanese securities firms in 1982-91. They
seek explanations for the observed capital cost differences in mac
roeconomic, risk, policy, and industrial organization factors. Their analysis
also contrasts the gap in capital costs between U.S. and Japanese
securities firms with the corresponding gap for industrial firms and banks.
28
Financial Liberalization and Monetary Control in Japan
Bruce Kasman and Anthony P. R odrigues
This article examines the effects of financial market reform on monetary
control in Japan during the past two decades. The authors identify changes
in the Bank of Japan’s operating strategy and evaluate the Bank’s ability to
influence key policy variables in the current liberalized environment.
Table of Contents
47
Expected Inflation and Real Interest Rates Based on Index-linked
Bond Prices: The U.K. Experience
G abriel de Kock
Some analysts have argued that indexed bonds convey important informa
tion for the formulation of monetary policy. This article investigates whether
a market measure of expected inflation derived from British indexed gilt
prices would be useful in predicting future inflation and real economic
activity.
61
Tracking the Economy with the Purchasing Managers’ Index
Ethan S. Harris
The purchasing managers’ index is a widely watched but virtually untested
indicator of manufacturing activity. This article examines how well the index
lives up to its billing as a leading indicator. The author also explores
whether the index supplies information about the economy beyond that
already provided by other indicators.
70
Treasury and Federal Reserve Foreign Exchange Operations
A report for the period May-July 1991.
76
List of Recent Research Papers
Do International Reactions of
Stock and Bond Markets
Reflect Macroeconomic
Fundamentals?
by Eli M. Remolona
Much of the movement in stock and bond markets
appears to be independent of changes in fundamental
economic conditions. Stock prices in particular often
fluctuate more sharply than would be justified by shifts
in the underlying fundamentals. Such market volatility
may distort the economy’s allocation of capital and on
occasion lead to liquidity crises and macroeconomic
instability. This article investigates one possible source
of the volatility: a tendency of market participants to
overreact to developments in other markets. The analy
sis yields some evidence that stock and bond markets
move together to an extent not easily explained by
changes in macroeconomic fundamentals.
Empirical work by Campbell and Shiller, Poterba and
Summers, West, and Bulkley and Tonks supports the
notion that the volatility of U.S. and U.K. stock prices
exceeds the volatility implied by fundam entals.1
Schwert finds that the timing of volatility in the U.S.
stock market often fails to coincide with that of mac
roeconomic fundamentals.2 A related finding by Bennett
and Kelleher is that high volatility in stock markets
1For U .S. stocks, see John Y. C am p b ell and Robert J. Shiller, "C o
integration and Tests of Present Value M odels," J o u rn a l o f P o litic a l
E co n om y, vol. 95 (1987), pp. 1062-88; Jam es M. Poterba and
Law rence H. Sum m ers, “M ean Reversion in Stock Prices: Evidence
and Im p lic atio n s ,” J o u rn a l o f F in a n c ia l E c o n o m ic s , vol. 22 (1988),
pp. 2 7 -5 9 ; and Kenneth D. West, “A Specification Test for
S p eculative B ubb les," Q u a rte rly J o u rn a l o f E c o n o m ic s , vol. 102
(1987), pp. 5 5 3 -8 0 . For U.K. stocks, see G eorge Bulkley and Ian
Tonks, "Are U.K. Stock Prices Excessively Volatile? Trading Rules
and V ariance Bound Tests," E c o n o m ic J o u rn a l, vol. 9 9 (1989),
pp. 1 08 3 -9 8.
2G. W illiam Schw ert, “Why D oes Stock M arket Volatility C hange over
Time?" J o u rn a l o f F in a n c e , vol. 44 (D e c e m b e r 1989), pp. 1115-53.
tends to be associated with high return correlations
across the markets.3 This result may arise because
stock markets have overreacted to one another. When
one market departs from fundamentals, the mistake is
propagated to other markets, thus giving rise to the
excess volatility. Moreover, Shiller and Beltratti find that
domestic stock markets often overreact to domestic
bond markets.4 Hence, unaccountable movements in
one market can be compounded as they spread to other
markets at home and abroad.
Consider an extreme example of high volatility: the
October 1987 stock market break. The global nature of
the event could be seen as evidence that stock markets
were overreacting to one another. Alternatively, some
would argue that stock investors at the time were simply
responding to information contained in price changes in
other markets, particularly information on expectations
of real activity abroad.5 But conclusions based on a
3Paul B ennett and Jea n ette Kelleher, "The International Transmission
of Stock Price D isruption in O c to b e r 1987," this Q u a rte rly R eview ,
Sum m er 1988, pp. 17-33. See also Paul K upiec, "Financial
Liberalisation and International Trends in Stock, C orp o ra te Bond,
and Foreign E xchange M arket Volatilities," O E C D D ep artm e n t of
Econom ics and Statistics Working Papers, no. 94, February 1991.
♦Robert J. Shiller and A ndrea E. B eltratti, "Stock Prices and Bond
Yields: Can Their C om ovem ents Be E xplained in Terms of Present
Value M odels?" National Bureau of Econom ic R esearch, Working
Paper no. 3 4 6 4 , 1990.
sSee R ichard W. Roll, "The International Crash of O c to b e r 1987," in
Robert Kam phuis, R oger K orm endi, and J.W. H en ry W atson, eds .,
B la c k M o n d a y a n d th e F u tu re o f F in a n c ia l M a rk e ts (M id -A m e ric a
Institute, 1988), pp. 3 5 -7 0 ; and M ervyn A. King and Sushil W adhani,
“Transmission of Volatility betw een Stock M arkets," R evie w o f
F in a n c ia l S tu d ie s , vol. 3 (1990), pp. 5 -36 .
FRBNY Quarterly Review/Autumn 1991
1
sin g le event are u n co n vin cin g , and for this reason, it is
im p o rta n t to co n sid er w h e th e r the broader experience
of sto ck and bond m arket co-m ovem ent is c o n siste n t
w ith m o ve m e n ts in real a c tiv ity , in te re s t rates, and
inflatio n.
T his a rtic le exam ines n e a rly tw o decades of m arket
expe rie n ce in the U nited S tates, Japan, and the U nited
K ingdom . The d iscu ssio n first focuses on the degree to
w hich m acroe co nom ic fu n d a m e n ta ls and foreign m arket
m o ve m e n ts e xp la in d o m e s tic m a rke t m o ve m e n ts. It
then asks w h e th e r a h yp o th e sis based on m arket m ove
m ents conveying in fo rm a tio n about fu n d a m e n ta ls can
explain the strong in te rn a tio n a l reactions m arkets have
to one another.
Of the d om e stic fu n d a m e n ta ls, future real a c tivity is
found to be the main force d rivin g the stock m arkets,
and future inflation, the main force driving the bond m ar
kets. Foreign fu n d a m e n ta ls a ppear to exe rt no direct
im pact on do m e stic m a rkets. A fte r taking account of the
e ffe cts of both do m e stic and foreign fu n d a m e n ta ls, the
a n a lysis finds tha t fo re ig n m arket returns still explain
m uch of the rem aining variance of do m e stic returns.
S u p e rfic ia lly , the s u b s ta n tia l re sid u a l e x p la n a to ry
pow er of foreign m arket m ovem ents m ight appear ipso
fa cto e vid e n c e th a t d o m e s tic m a rke ts o ve rre a ct to
m ovem ents in foreign m a rkets. But there is a fu rth e r
possibility, nam ely th a t the m ovem ents in foreign m ar
k e ts c o n v e y a d d itio n a l in fo rm a tio n re le v a n t to th e
d o m e stic m arkets. T h is in fo rm a tio n may bear on those
d o m e stic fu n d a m e n ta ls th a t influence d o m e stic m ar
kets. T hus, before a ve rd ict of “ o ve rre a ctio n ” by dom es-
Chart 1
Links among Markets and Fundamentals
Note: (A) and (B) represent the links between markets and
fundamentals, (C) represents the markets' reactions to each
other, and (D) represents the links between domestic and
foreign fundamentals.
Digitized for
2 FRASER
FRBNY Q uarterly Review/Autumn 1991
tic m arkets to fo re ig n m a rke ts can be b ro u g h t, this
p o s s ib ility m ust be exam ined.
The sig n ific a n t re la tio n s h ip s found to exist am ong
do m e stic and foreign fu n d a m e n ta ls and d o m e s tic and
foreign s e c u ritie s m a rke ts are illu s tra te d in C h a rt 1. The
link (A) represents the im pact of d o m e s tic fu n d a m e n ta ls
on the d o m e stic m a rke t and (B) the im p a ct of foreign
fu n d a m e n ta ls on the foreign m arket. T he lin k (C) rep re
se n ts the re a c tio n s of th e m a rk e ts to e ach other.
B ecause fu n d a m e n ta ls in one c o u n try are found to have
no direct bearing on the m a rke t in the o th e r country,
a p p ro p ria te reactions of d o m e stic m a rke ts to foreign
m arkets should reflect links betw een d o m e s tic and fo r
eign fundamentals, shown as (D). The hypothesis implied
by these relationships is that the domestic market tries to
infer information about foreign fundam entals from foreign
markets to better predict domestic fundamentals.
To d e te rm in e w h e th e r the actual m a rke t re a ctio n s are
co n siste n t w ith th is h yp o th e sis, each of th e id e n tified
links is estim a te d . Testing the h y p o th e s is involves c o m
paring e stim a te s of the a ctual m a rke t re a ctio n s w ith the
reactions im plied by the e stim a te s of the links betw een
fu n d a m e n ta ls and the links betw een m a rke ts and fu n
d am entals. The a n a ly s is fo cu se s on real a c tiv ity as the
main fu n d a m e n ta l fo r sto ck m a rke ts and on in fla tio n as
the co rre sp o nd in g fu n d a m e n ta l fo r bond m a rke ts. The
estim ates su g g e st th a t the U.S. and J a p a n e se sto ck
m arkets react to each o th e r’s m o ve m e n ts on average
more than w ould be ju s tifie d by the in fo rm a tio n th e se
m arkets convey a bout real activity, w h ile the U .S ., J a p
anese, and U.K. bond m a rke ts react to one a n o th e r’s
m ovem ents more than w ould be ju s tifie d by the in fo rm a
tion these m a rke ts convey a b o u t in fla tio n.
Explaining stock and bond returns
Id e n tify in g the fu n d a m e n ta ls
M arket p rices of sto cks and bonds can be sp e cifie d as
the present d is c o u n te d values of stre a m s of expected
cash flow s. H ence sto ck and bond re tu rn s sh o u ld be
affected by m a cro e co n o m ic fa cto rs th a t re fle ct d is c o u n t
rates or expected cash flow s. A t the sam e tim e , the
strength of th e se e ffe cts sh ould d iffe r betw e e n sto ck
and bond m a rkets, p a rtic u la rly the e ffe c ts of exp ecte d
real a c tivity and expected in fla tio n. D iffe re n ce s b etw een
the m arkets allow us to iso la te the e ffe c ts of one fu n d a
m ental on one m arket by using the o th e r m a rke t to
control for the e ffe cts of o th e r fu n d a m e n ta ls .
For stocks, cash flow s m ay ta ke the form of d ivid e n d
paym ents, pro ce e d s from sto ck bu yb a cks by the issu ing
firm , or p roceeds from sto ck p u rch a se s by ta ke o ve r
fir m s .6 P re v io u s s tu d ie s s h o w th a t s e v e ra l m a c 6Cash flows from stock bu ybacks and takeovers becam e im po rtan t
in the U nited States in the 1980s.
ro e cono m ic va ria b le s — the d ivid e n d yield, the spread
betw een long and sh o rt inte re st rates, inflation, and real
a c tiv ity — s ig n ifica n tly influ e nce re tu rn s.7 The d ivid e n d
yield and term spread should reflect d isco u nt rates and
b u siness co n d itio n s. E xpected inflation should affect
real d isco u n t rates, but it may also affect cash flow s. Of
th e se m a c ro e c o n o m ic v a ria b le s , fu tu re real a c tiv ity
sh o u ld e xe rt the s tro n g e s t in flu e n c e on sto ck cash
flow s.
U nlike stocks, bonds in th is study have cash flow s
th a t are fixed in nom inal term s. C hanges in the s h o rt
term real in te re st rate a ffectin g the d isco u nt rate should
have a s tro n g e r e ffe ct on bond returns than on stock
re tu rn s .8 E x p e c te d in fla tio n s h o u ld e x e rt an even
stro n g e r in flu e nce on bond returns b e ca u se of its direct
effect on the real value of in te re s t and p rin c ip a l pa y
m ents. By co n tra st, real a c tiv ity sh o u ld have re la tive ly
little e ffe ct on bond cash flow s, p a rtic u la rly in the case
of g o v e rn m e n t b o n d s not s u b je c t to d e fa u lt ris k ,
although it may affect d is c o u n t rates.
D e s c rip tio n o f key va ria b le s
The key v a ria b le s fo r th is s tu d y are e x c e s s s to c k
returns, excess bond returns, real a c tiv ity g row th, and
inflation rates (Table 1). E xcess returns are the returns
over a q u a rte r m inus a th re e -m o n th in te re s t rate at the
b e g inning of the q u a rte r.9 The a n a ly s is relies on quar-
7See in p a rticu la r Nai-Fu Chen, Richard Roll, and Stephen A. Ross,
"E conom ic Forces and the Stock M arket," Journal of Business,
vol. 59 (1986), pp. 383-403; and Eugene F. Fama, "Stock Returns,
E xpected Returns, and Real A ctivity," Journal of Finance, vol. 45
(1990), pp. 1089-1108 Chen, Roll, and Ross also identify the yield
spread between low -grade and high -grade bonds as a significant
variable, but Fama finds that the d ivide nd yield does just as well
as the the risk spread. The estim ates presented here will use the
divide nd yield because data for this variable are available for
foreign markets. Shiller som etim es uses the price-earnings ratio
instead of the d ivide nd yield.
8See John Y. C am pbell and John Ammer, "W hat Moves the Stock
and Bond Markets? A Variance D ecom position for Long-Term Asset
Returns," paper presented at the annual m eeting of the A m erican
Finance A ssociation, W ashington, D C.. D ecem ber 1990.
9Stock returns are measured as the divid e n d yield plus the change
in the log stock price index Bond returns are ap proxim ated by the
yield at the beginning of the qu arte r minus the cha nge in log yields.
Table 1
Statistical C haracteristics of Key Variables
(Q uarterly Data at an Annual Rate, June 1973 to S eptem ber 1989)
Excess Stock Returns*
U nited States
Japan
U nited K ingdom
Excess Bond Returns*
Mean
Standard
Deviation
Mean
Standard
Deviation
5.27
15.65
19.98
37.56
32.64
50.30
0.23
3,97
0 82
30.30
38.66
40.10
U nited States
Japan
U nited States
0.62
0.38
0.35
0.45
0.20
Stock Return C orrelations
Japan
U nited K ingdom
B ond Return C orrelations
0.42
Real Growths
Mean
U nited States
Japan
United Kingdom
Inflation
Standard
Deviation
2,50
3.79
1.88
4.08
3.24
8.90
United States
0,30
0 05
S tandard
Deviation
Mean
6.42
5.26
9.84
Real Growth C orrelations
Japan
U nited Kingdom
Japan
3.89
6.19
7.26
Inflation C orrelations
Japan
0.07
U nited States
0.57
0.55
Japan
0.52
^Excess stock returns are stock returns minus a three-m onth interest rate.
^Excess bond returns are bond returns minus a three-m onth interest rate.
§ lo g change in real GNP or GDP.
FRBNY Q uarterly R eview/Autum n 1991
3
te rly d ata from June 1973 to S eptem ber 1990 for the
U nited S tates, Japan, and the U nited K in g d o m .10 C hart
2 show s the beh avior of excess stock returns and C hart
3 the beh avior of excess bond returns over the p e rio d .11
U sing excess returns a llow s us to abstract from possi10The analysis looks at stock and bond returns only up to the third
q u a rte r of 1989 because these returns are related to leads of
m acroeconom ic variables that go up to the third qu arter of 1990.
11The lack of movement in excess bond returns in Japan in the early
1970s seems to reflect a m arket sub je ct to “ g u id a n c e ” by the
m onetary authorities. Efforts to take acco unt of this period with the
use of dum m y variables did not alter the analysis.
4 FRBNY Q uarterly Review/Autumn 1991
ble e ffe c ts of c ro s s b o rd e r s h o rt-te r m in te re s t-ra te
a rb itra g e and to focus on the m a cro e co n o m ic v a ria b le s
m ost relevant to sto ck and bond m a rke ts. Real grow th is
m easured by GNP or GDP, in fla tio n by th e co n su m e r
price index.
As m easured by th e ir sta n d a rd d e v ia tio n s , excess
stock returns are o n ly so m e w h a t m ore vo la tile than
excess bond re tu rn s . S to c k re tu rn s , how ever, seem
m ore c o rre la te d a cro ss m a rk e ts th a n bon d re tu rn s,
except betw een Japan and the U nited K in g d o m . Real
grow th and in flation seem a b o u t e q u a lly v o la tile , but
they are c le a rly less v o la tile th a n s to c k and bond
returns. Real grow th also te n d s to be m uch less corre-
lated across co u n trie s than does inflation. S tock returns
tend to be sig n ific a n tly more co rrelated than real grow th
across co u n trie s, w hile bond returns tend to be less
co rrelated than inflation.
E x p la in in g s to c k re tu rn s
To id e n tify the s ig n ifica n t links am ong sto ck m arkets
and fu n d a m e n ta ls, th is a n a lysis evaluates the degree to
w hich m a cro e co no m ic varia b le s and foreign sto ck m ar
ket m ovem ents explain do m e stic stock m arket m ove
m ents. Table 2 rep orts a d ju ste d R -squared s ta tis tic s as
m easures of the e xp la na to ry pow er of reduced-form
sto ck retu rn e q u a tio n s . The d e p e n d e n t v a ria b le is
excess stock returns, or the returns m inus a three-
m onth in te re st rate. The e x p la n a to ry va ria b le s com m on
to all th e e q u a tio n s are th e la g g e d d iv id e n d y ie ld ,
excess bond returns, and fo u r q u a rte rly leads of real
grow th. The a d ju ste d R -squared s ta tis tic s fo r the basic
equation are 20 p e rce n t fo r the U.S. m a rke t, 16 p e rce n t
for the Ja p a n e se m arket, and 25 p e rc e n t fo r the U.K.
m arket. A lthough th e se s ta tis tic s show th a t a large
p ro p o rtio n of the m ovem ent of sto ck re tu rn s is left
u nexplained, the e x p la n a to ry va ria b le s used here are
som ew hat more su c c e s s fu l than va ria b le s used in p re
v ious s tu d ie s .12
12Using U.S. data for 1953-87, Fama estim ates a sim ilar equation that
gives an unadjusted R-squared of 23 pe rcent for q u a rte rly stock
Chart 3
Bond Market Excess Returns
Quarterly Returns at an Annual Rate
Percent
Source: Japanese and U.K. data, Bank for International Settlements.
Note: Excess returns are bond returns over a quarter minus a three-month interest rate at the start of the quarter.
FRBNY Q uarterly Review/Autum n 1991
5
To evaluate the im p o rta n ce of future real a c tiv ity re la
tive to oth e r fu n d a m e n ta ls, a second equation is e s ti
m ated fo r each m arket, th is tim e including the change
in the sh o rt-te rm real in te re st rate and fo u r leads of
inflatio n. The sh o rt-te rm real rate is based on ex post
inflation. The a d d ition a l va ria b le s produce a sig n ifica n t
gain in e xp la na to ry pow er fo r the Japanese and U.K.
m a rkets. But for U.S. sto ck returns, the in clu sion of
in fla tio n reduces the a d ju ste d R -squared s ta tistic, in d i
c a tin g o n ly a w e a k e ffe c t. S u b s e q u e n t e s tim a te s
inclu de in flation leads for the Japanese and U.K. m ar
kets but not for the U.S. m arket.
The use of leads fo r real grow th and inflation allow s
fo r th e e x tra in fo rm a tio n m a rk e t p a rtic ip a n ts m ay
have.13 However, the e q u a tio n s in effect leave out fo re
cast errors by p a rtic ip a n ts , an om itted va ria b le problem
that w ould bias the c o e fficie n t e stim ates. A lthough this
om issio n w ould not a ffe ct the evaluation of e xp la na to ry
power, the a n a lysis of the in fo rm a tio na l c o n siste n cy of
m arket reactions in the next section requires a p ro c e
dure tha t gives unbiased co e ffic ie n t e s tim a te s .14
Footnote 12 (continued)
returns, com pared with 27 percent here. The explanatory power is
lower for m onthly returns and higher for annual returns. See Fama,
"S tock Returns "
13See Fama, “ Stock Returns," and C am pbell and Ammer, “ What
Moves the Stock and Bond M arkets?"
14Suppose we have the relationship r, = a + bE,(yl t t ), where the
return r, depends on the current expectation of a future value y, + (.
The realized value would be y , = E,(y,t ,) + e, ,,, where e ,+ , is
the expectational error. Since expectations are unobservable, we
m ight run the regression on the realizations, estim ating r, = a +
b y ,t , + u,. Unfortunately, doing so would give a biased estim ate of
Table 2
Explaining Stock M arket Excess Returns:
Adjusted R-Squared Statistics
Japan
United States
Equation with real
growth ieadsf
United Kingdom
0.42 (4.5**)
0.18 (0.8)
0.18 (0 5)
0.45 (1.3)
A ddition of foreign
excess returns
0.42 (12.2**)
o
b
0.24 (2.3**)
A ddition of foreign
real growth
leads
o
0.25
0.20
iv>
0.16
A ddition of inflation
0.18 (0.1)
leads*
) 0.46 (1.8)
Notes: Parentheses contain F-statistics indicating how each addi
tion affects the explanatory power. Double asterisks indicate
significant addition to explanatory power at the 5 percent level
com pared with the previous equation.
^Equations include lagged dividend yield, domestic excess bond
returns, and four leads of real growth.
6 FRBNY Q uarterly Review/Autumn 1991
Next let us te st the p o ssib le relevance of foreign
fu n d a m e n ta ls to d o m e s tic sto ck returns. Foreign real
a ctivity, fo r e xa m p le , m ay in flu e n c e d o m e s tic s to c k
returns to the e xtent th a t som e of the firm s tra d e d in the
m arket c o n d u ct an im p o rta n t p a rt of th e ir b u sin ess
abroad. Table 2 re p o rts the a d ju ste d R -squared s ta tis
tics for eq u a tio n s th a t ta ke a cco u n t of fo re ig n fu n d a
m entals, here co n s is tin g of fo u r le a d s of real g ro w th in
each of the tw o o th e r co u n trie s . In no in s ta n c e is the
gain in e xp la n a to ry pow er s ta tis tic a lly s ig n ific a n t, a fin d
ing that in d ica tes little d ire ct relevance of fo re ig n real
a c tivity to d o m e stic m a rk e ts .15 T h is m eans th a t if fo r
eign m arkets do convey in fo rm a tio n relevant to the
fu n d a m e n ta l d e te rm in a n ts of m ovem ents in d o m e stic
m arkets, the path of in flu e nce m ust link m o ve m e nts in
foreign fu n d a m e n ta ls to d o m e stic fu n d a m e n ta ls , w ith
m ovem ents in foreign m a rke ts se rv in g as the cha n n e l
through w hich th is in fo rm a tio n is conveyed.
A p re lim in a ry te s t fo r m a rke t o ve rre a ction is to d e te r
mine w h e th e r returns in fo re ig n m a rke ts add s ig n ific a n t
e x p la n a to ry p o w e r to s to c k re tu rn e q u a tio n s th a t
alre a d y ta ke a c c o u n t of m a c ro e c o n o m ic fu n d a m e n
ta ls .16 Table 2 re p o rts a d ju ste d R -sq u a re d s ta tis tic s for
e q uations that in clu d e fo re ig n excess sto ck re tu rn s. The
gain is q uite im pressive fo r the U.S. and Ja p a n e se
stock m arkets, s u g g e s tin g th a t p a rtic ip a n ts in the se
m arkets may often overreact to d e ve lo p m e n ts in m ar
kets a b ro a d .17
E xp la in in g b o n d re tu rn s
To id e n tify the s ig n ific a n t links am ong bond m arkets
and fu n d a m e n ta ls , th e a n a ly s is now e v a lu a te s the
degree to w hich m a cro e co no m ic v a ria b le s and fore ign
bond m arket m ovem ents explain d o m e s tic bond m a rke t
m ovem ents. Table 3 re p o rts a d ju ste d R -sq u a re d s ta tis
tics as m easures of e x p la n a to ry pow er fo r re d u ce d -form
bond re tu rn e q u a tio n s . T h e d e p e n d e n t v a ria b le is
Footnote 14 (continued)
b, because the residual term u, contains e(+J, w hich w ould be
correlated with yl +1.
1®The finding is consistent with the con clu sion that do m estic factors
dom inate international factors in explaining stock returns of
individual firms, even of m ultinationals. See Bruno Solnik,
International Investm ents, 2d ed. (Reading, M assachusetts:
Addison-W esley, 1991), pp. 135-40.
16Robert P indyck and Julio R otem berg use this approach to co n clu d e
that there is overreaction in m arkets for inte rnatio nally tra ded
com m odities. See “ The Excess C om ovem ent of C om m odity P rices,"
Econom ic Journal, vol. 100 (1990), pp. 1173-89.
17As to the effects of the individ ual foreign stock m arkets, the
Japanese m arket is significa nt for the U.S. m arket, as is the U.S.
market for the Japanese m arket. In the case of bond m arkets, the
Japanese market matters sig n ifica n tly for the U.S. m arket, the U.S.
and U.K. m arkets for the Japanese m arket, and the Japanese
market for the U.K. market.
excess returns on lon g-term g o vernm ent bonds, th a t is,
the returns m inus a th re e -m o n th rate. The com m on
e x p la n a to ry v a ria b le s are e xce ss sto c k re tu rn s, the
change in the real s h o rt-te rm rate, and fo u r q u a rte rly
leads of inflation. As w ith stocks, the e xp la na to ry power
of the se va ria b le s is good by the standards of the
lite rature . The a d d ition of fo u r leads of real grow th does
not help sig n ific a n tly for any of the bond m arkets. S ub
se q u e n t e stim a te s of bond return e q uations leave out
real grow th leads. The a n a lysis confirm s th a t future
in fla tion is a m ore im p o rta n t fu n d a m e n ta l for bond m ar
kets than is future real activity, w h ile future real a c tiv ity
is m ore im p o rta n t fo r stock m arkets.
W hen foreign fu n d a m e n ta ls in the form of fo u r in fla
tio n le a d s in each of the tw o o th e r c o u n trie s are
in clu d e d , the gain in e xp la n a to ry power is s ta tis tic a lly
in s ig n ific a n t in every case. A gain foreign fu n d a m e n ta ls
seem to have little d ire ct relevance to d o m e stic m ar
kets. T his finding m eans th a t if foreign m arkets convey
in fo rm a tio n a b o u t fu n d a m e n ta ls , th e re le v a n c e to
d o m e stic m arkets is like ly to arise from links betw een
foreign fu n d a m e n ta ls and d o m e stic fu n dam entals.
Table 3 also reports the a d ju ste d R -squared s ta tistics
fo r e q u a tio n s in co rp o ra tin g foreign excess bond returns.
T he su b sta n tia l gain fo r all three bond m arkets s u g
g e sts th a t p a rtic ip a n ts in th e se m arkets could be over
reacting to m arket m ovem ents abroad.
In te rp re ta tio n
The strong e xp la n a to ry pow er of foreign m arket returns
by its e lf does not m ean th a t m arket overreaction exists.
It is easy to im agine the m arkets m oving to g e th e r in
Table 3
Explaining Bond M arket Excess Returns:
Adjusted R-Squared Statistics
United States
Japan
United Kingdom
Equation with
inflation leads*
0.26
0.14
0.24
Addition of real
growth leads*
0.23 (0.6)
0.11 (0.6)
0.23 (0.7)
Addition of foreign
inflation leads
0.28 (1.2)
0.12 (0.9)
0.19 (0.5)
Addition of foreign
excess returns
0.37 (4.6**)
0.33 (9.0**)
0.25 (3.0**)
Notes: Parentheses contain F-statistics indicating how each a d di
tion affects the explanatory power. Double asterisks indicate
significant addition to explanatory power at the 5 percent level
compared with the previous equation.
^Equations include change in real short-term rate, domestic
excess stock returns, and four leads of inflation.
response to com m on or c o rre la te d in fo rm a tio n a bout
fu n d a m e n ta ls not ca ptured by o u r m a cro e co n o m ic v a ri
ables. Such in fo rm a tio n m ay so m e tim e s cause the d if
fe re n t m a rk e ts to m ake s im ila r m is ta k e s a b o u t th e
future, so th a t the e x p la n a to ry pow er of fo re ig n returns
may a rise sim p ly from the co rre la tio n of fo re ca st errors.
In particular, foreign sto c k and bond m a rke ts m ay throw
out s ig n a ls a b o u t fo re ig n m a c ro e c o n o m ic fu n d a m e n ta ls
th a t in turn help p re d ict d o m e s tic fu n d a m e n ta ls . The
s ig n a ls may at tim e s turn o u t to be fa lse , so th a t the
im plied future d e ve lop m e n ts do not show up in the data,
b u t th e d o m e s tic m a rk e ts w ill have b e e n rig h t to
respond to the sig n a ls. The a n a ly s is in th e next se ctio n
a s k s w h e th e r a c tu a l m a rk e t b e h a v io r c a n be so
ju s tifie d .
Testing for inform ational consistency
A n a ly tic a l a p p ro a c h
If m arkets are in fact reacting to in fo rm a tio n related to
fu n d a m e n ta ls and if, as the d ata in d ica te , fo re ig n fu n d a
m entals have little d ire ct relevance to d o m e s tic m arkets,
then the m arkets m ust be reacting to one a n o th e r’s
m ovem ents la rg e ly b ecause of re co g n ize d lin ks betw een
d o m e stic and fo re ig n fu n d a m e n ta ls . T he d o m e s tic m ar
ket w ill be try in g to in fe r in fo rm a tio n a b o u t foreign
fu n d a m e n ta ls from the o th e r m a rke ts in o rd e r to m ake
b e tte r fo re ca sts of d o m e s tic fu n d a m e n ta ls . If the actual
reaction is c o n s is te n t w ith the links betw een fu n d a m e n
ta ls and w ith the links betw een fu n d a m e n ta ls and m ar
ket returns, then in fo rm a tio n a l c o n s is te n c y h o ld s .18
The e m p irica l a n a ly s is below p ro ce e d s by e s tim a tin g
the various s p e cifie d links am ong m a rke ts and fu n d a
m entals to te st fo r in fo rm a tio n a l c o n s is te n c y .19 The d is
c ussion fo cu se s on real a c tiv ity as the fu n d a m e n ta l
m ost im p o rta n t to sto c k m a rke ts and on in fla tio n as the
fu n d a m e n ta l m ost im p o rta n t to bond m a rke ts. Three
q u e stio n s are a d d re sse d : How c lo s e ly w ould sto ck m ar
kets move to g e th e r if the d o m e s tic m a rke ts were relying
on foreign sto ck m a rke ts s im p ly fo r in fo rm a tio n a b o ut
foreign real a c tiv ity ? How c lo s e ly w ould bond m a rke ts
move to g e th e r if the d o m e s tic m a rke ts were relying on
foreign bond m a rke ts sim p ly fo r in fo rm a tio n a b o u t fo r
eign in fla tio n? How c lo s e ly do th e m a rke ts a ctu a lly
move to g e th e r?
S to ck re tu rn s a n d re a l g ro w th
To o btain unbiased e s tim a te s of the e ffe c t of real grow th
on stock returns, we can re e stim a te the sto ck return
e q u a tio n s by replacing the fo u r leads of real grow th and
18The appendix offers a form al m odel for this analysis.
1®Unlike a test based on statistica l exp la nato ry power, this test does
not require com plete data on the markets' inform ation, only enough
data to produce good coe fficients on the various links.
FRBNY Q uarterly Review/Autum n 1991
7
in fla tio n w ith th e ir p redicted va lu e s.20 The d ivid e n d yield
and excess bond returns are kept in the eq u a tio n s to
control fo r oth e r fu n d a m e n ta ls, p a rtic u la rly the effects
of chan ges in the real d is c o u n t rate. Inflation is om itted
from the U.S. e stim a te s because it lacks a d d ition a l
e xp la n a to ry power.
The e stim a te s in Table 4 su g g e st th a t the Japanese
stock m arket is the m ost sensitive to predicted real
grow th and the U.K. m arket the least sensitive. The sum
of the c o e fficie n ts on real grow th indicates that w hen
exp ecte d d o m e stic real grow th over the next fo u r q u a r
te rs increases by a point, U.S. excess sto ck returns rise
2.5 p e rce n ta g e points on average, Japanese excess
returns rise 4.3 points, and U.K. excess returns rise 0.6
of a point.
In te rn a tio n a l re a l g ro w th lin ks
If real grow th abroad has no d ire ct effect on d o m e stic
sto ck returns, a reaction to foreign m arkets should in d i
cate a link betw een grow th abroad and grow th at home.
To e stim a te th is link, an index of real grow th leads is
co n stru cte d fo r each m arket using w eights p ro p o rtio n a l
to the real grow th c o e ffic ie n ts in the e stim ated stock
20The instrum ents used to predict real growth and inflation are the
divide nd yield, excess bond returns, contem poraneous real growth
and inflation, and four lags each of dom estic inflation, real growth
in each of the three countries, the change in log dollar oil prices,
and the changes in the two relevant log exchange rates.
return e q u a tio n s .21 The m ovem ents in th is index for
other c o u n trie s are p re su m a bly w hat d o m e s tic m arket
p a rtic ip a n ts can in fe r from m ovem ents in fo re ig n sto ck
prices.
The upper se ctio n of Table 5 re p o rts the e stim a te d
effe cts of foreign fu tu re real grow th on d o m e s tic fu tu re
real g ro w th .22 The e stim a te s are based on the c o n
stru cte d indexes and co n tro l fo r c u rre n tly o b s e rva b le
v a ria b le s th a t m ay a lso h e lp p re d ic t d o m e s tic real
g ro w th . To a llo w fo r the jo in t d e te rm in a tio n of real
grow th in d iffe re n t co u n trie s , in s tru m e n ts are used fo r
the foreign real grow th in d e xe s.23 The e s tim a te s in d i
cate th a t w hen U.S. real grow th over the next fo u r
q u a rte rs rises a p e rce n ta g e point, J a p a n e se real grow th
can be expected to rise 0.15 of a p o in t and U.K. real
grow th 0.60 of a point. W hen fu tu re J a p a n e se real
grow th rises a point, U.S. real grow th can be expected
to rise 0.60 of a p oint and U.K. real grow th to fa ll 0.3 6 of
21For exam ple, the index for U.S. real growth w ould have a 60
percent w eight for the first lead because the co e fficie n t on this
lead is 60 percent of the sum of the four lead coe fficients
(Table 4).
22The control variables are the current values and four lags each of
dom estic real growth and inflation.
23The instrum ents are four lags of real growth and inflation in each of
the three countries.
Table 4
Stock Return Equations Estimated by Two-Stage Least Squares
(D ependent Variable Is Excess Stock Return)
United States
C onstant
D ividend yield
E xcess bo nd return
P redicted real growth
First lead
Second lead
Third lead
Fourth lead
Sum of c oe fficients
U nited K ingdom
- 1 9 .5 0
2.88
0.38*
( - 1 .2 9 )
(1 31)
(2.71)
- 2 0 .2 5
2.73
0.15
( -0 .5 8 )
(1.09)
(1.21)
17.66
0.87
0.49*
1.51
3.76
- 3 .3 9 *
0.64
(0.74)
(1-55)
(2 69)
(0.32)
3.32
1.54
- 0 .9 4
0.34
(1.45)
(0.74)
( - 0 .4 5 )
(0.17)
0.79
0.51
-0 .9 1
0.24
4.27
2.52
P redicted inflation
First lead
Second lead
Third lead
Fourth lead
R-squared
A djuste d R-squared
Japan
- 1 .1 2
0.14
0.05
(
1.83
-3 .4 1 *
1.81
0.28
0.15
(0.51)
(1.21)
(3.02)
(0.79)
(0.51)
(-0 .9 8 )
(0 24)
0.63
- 0 .8 7 )
(1.31)
( -2 .4 4 )
(1.14)
2.10
2.76*
- 4 .7 2 *
-1 .8 9
(1.47)
(1.70)
( - 3 .0 4 )
( - 1 .3 2 )
0.48
0.38
Notes: Instrum ents are d ivide nd yield, excess bond returns, contem poraneous real growth and inflation, and four lags each of dom estic
inflation, real growth in each of the three countries, percentage change in d o lla r oil prices, and pe rce n ta g e cha nge in the two d o lla r
exchange rates. T-statistics are in parentheses. Asterisks indica te significa nce at 10 pe rcent level.
8 FRBNY Q uarterly Review/Autumn 1991
a p o in t.24 U.K. real grow th seem s to have little e ffe ct on
real grow th in the other econom ies.
The large stan dard erro rs indicate th a t these e s ti
m ates are not very precise. The usual sig n ifica n ce tests
w ould su g g e st that there are no real grow th links. N o n e
th e less, this a n a lysis m ust give the m arkets the benefit
of the dou b t by a llow in g them a reason to react to one
a n o th e r’s m ovem ents. M arket p a rtic ip a n ts presum ably
do not lim it th e ir responses to only those influences that
su rvive strin g e n t s ta tis tic a l te sts. Hence the a n a lysis
w ill proceed on the a ssu m p tio n th a t the e stim a te s of
real grow th links in Table 5 are our best e stim ates.
Im p lie d s to c k m a rk e t re a c tio n s
The m iddle section of Table 5 show s the m agnitudes of
sto ck m arket in te ra ctio n s im plied by the links betw een
m a rke ts and real a c tiv ity and the links betw een d o m e s
tic and foreign real activity. The stro n g e st im plied reac
tio n s are betw een the U.S. and Japanese m arkets. The
various links am ong m a rke ts and fu n d a m e n ta ls im ply
th a t if U.S. m arket p a rtic ip a n ts saw Ja panese stock
re tu rn s rise 1 p e rc e n ta g e p o in t w hile o th e r fa c to rs
re m a in e d u n c h a n g e d , th e y w o u ld in fe r a ris e in
expected Jap anese real grow th of 0.23 of a point (1
d ivid e d by 4.3) and thus a rise in expected U.S. real
grow th of 0.14 of a point (0.23 tim e s 0.60), so th a t U.S.
24Because real growth is correlated across countries, the sum of the
coe fficients provides better estim ates than do the individual
coefficients. In the case of U.K. growth, for example, the sum of
the estim ated effects of U.S. and Japanese growth of 0.24 is more
reliable than the individ ual effects of 0.60 and -0.36, respectively.
stock returns w ould rise 0 .3 5 of a p o in t (0.14 tim e s 2.5).
S im ila rly, if J a p a n e s e m a rk e t p a rtic ip a n ts saw U.S.
stock returns rise 1 point, th e y w ould react so that
Japanese sto ck returns rise 0 .25 of a p oint. The various
links do not im ply strong p o sitive re a ctio n s in the case
of the U.K. m arket.
A c tu a l s to c k m a rk e t re a c tio n s
T h e te s t o f in fo r m a tio n a l c o n s is te n c y u s e d h e re
involves e xtra ctin g th a t p a rt of fo re ig n sto ck m arket
returns reflecting m ovem ents in fo re ig n exp e cte d real
activity. To th is end, the e stim a te d fo re ig n sto ck return
e q u a tio n s are cle a n se d of the e ffe cts of o th e r fu n d a
m e n ta ls to cre a te p re d ic to rs of fo re ig n real a c tiv ity
g ro w th .25 The p re d icto rs of fo re ig n real a c tiv ity grow th
are s u b stitu te d into the real grow th link e q u a tio n s, and
the resulting p re d ictio n s of d o m e s tic real grow th are in
turn s u b s titu te d into the d o m e s tic sto ck return e q u a
tions. In p rin cip le , th e se e q u a tio n s co n tro l fo r the m ove
m e n ts o f fu n d a m e n ta ls o th e r th a n re a l a c tiv ity ,
p a rtic u la rly those reflected in excess bond returns. The
estim ated c o e ffic ie n ts on the fo re ig n real grow th p re d ic
to rs in th e m o d ifie d s to c k re tu rn e q u a tio n s p ro v id e
m easures of sto ck m a rke t re a ctio n s to the relevant
foreign m arket m o ve m e n ts.26
25From the excess stock return equations in Table 4, the terms
involving the divide nd yield, excess bond returns, and inflation are
subtracted, so that only the real growth term s are left
26The regression is based on equation A .5 of the model developed in
the appendix
Table 5
Real Growth Links and Implied and Actual Stock Market Reactions
United States
Japan
U nited
Kingdom
Effect on dom estic real
growth of real growth in:
United States
Japan
United Kingdom
0.60
0.03
(0.50)
(0.08)
0.35
0.12
(0.30)
(0.32)
0.62*
0.04
(0.13)
(0.09)
0.15
(0.13)
- 0 .0 8
(0 05)
0.60
- 0 .3 6
(0.78)
(1.72)
0.15
- 0 .0 5
(0.20)
(0.25)
0.14
0 29
(0.17)
(0.21)
Im plied dom estic stock
m arket reaction to stock m arket in:
U nited States
Japan
United Kingdom
0.25
(0.22)
-0 .5 4
(0.34)
Actual dom estic stock
market reaction to stock market in:
U nited States
Japan
United K ingdom
0.38*
(0.08)
0.08*
(0.07)
Notes: S tandard errors are in parentheses. Asterisks indica te significa ntly greater actual over im p lie d reaction at the 10 pe rcent level.
FRBNY Q uarterly R eview/Autum n 1991
9
The bottom sectio n of Table 5 reports the e stim a te s of
actual sto ck m arket reactions. The estim ates show s ig
n ifica n t o ve rrea ction by the U.S. and Ja panese stock
m a rkets to each other. In response to a 1 point rise in
Ja p a n e se sto ck returns, U.S. sto ck returns rise 0.62 of
a p oint on average, an increase nearly tw ice the m a g n i
tude ju s tifie d by the in fo rm a tio n about real a c tiv ity co n
veyed by the Jap anese sto ck m arket. S im ilarly, the size
of the Japa nese sto ck m arket reaction to U.S. stock
m arket m ovem ents is ha lf again as great as the size
im plied by in fo rm a tio n a l consistency. A lthough the J a p
anese sto ck m arket is show n to have a s ta tis tic a lly
s ig n ific a n t overreaction to the U.K. m arket, th is result is
b a s e d on an im p la u s ib ly la rg e n e g a tiv e im p lie d
reaction.
B o n d re tu rn s a n d in fla tio n
To o b ta in unbiased e stim a te s of the e ffe ct of inflation,
the bond return eq ua tio n s are reestim ated by replacing
the inflation leads w ith th e ir predicted va lu e s.27 Excess
sto ck returns are kept as a se parate variable in the
eq u a tio n s to co ntrol fo r o th e r fu n d a m e n ta ls, p a rtic u la rly
for the po ssib le e ffe cts of future real grow th on the
d isco u n t rate. The estim a te s reported in Table 6 show
that the U.S. and Ja p a n e se bond m arkets are very
s e n sitive to inflatio n, w hile the U.K. m arket in e xp lica b ly
re s p o n d s p o s itiv e ly to in fla tio n . A 1 p o in t ris e in
expected inflation over the next fo u r q u a rte rs reduces
U .S . b o nd re tu rn s 3 .8 p o in ts and J a p a n e s e bond
returns ne arly 3.0 points.
27The instrum ents used to predict inflation are excess stock returns,
current real growth and inflation, and four lags each of dom estic
real growth, inflation in each of the countries, oil price inflation, and
currency deprecia tion rates.
In te rn a tio n a l in fla tio n lin ks
To estim a te the in fla tio n links, an index of in fla tio n leads
is co n stru cte d fo r each country. The w e ig h ts are d erived
from the bond return e q u a tio n s in the sam e way th a t
they were draw n from the sto ck return e q u a tio n s fo r the
real grow th indexes. The m ovem ents in th is in fla tio n
index are w hat can be inferred from bond m a rke t m ove
m ents. Based on th e se indexes, the e s tim a te s in the
upper se ctio n of Table 7 m easure the e ffe c ts of foreig n
on d o m e stic fu tu re in fla tio n. T he e s tim a te s co n tro l for
o th e r fa cto rs and fo r the jo in t d e te rm in a tio n of inflation
in the d iffe re n t c o u n trie s .28 Here a 1 p o in t rise in the
J a p a n e se in fla tio n rate o ve r th e next fo u r q u a rte rs
raises the expected U.S. rate 0 .29 of a p oint, and a 1
point rise in the U.S. rate ra ise s the J a p a n e se rate 0.43
of a point. A 1 p oint rise in both the U.S. and Ja p a n e se
rates raises the U.K. rate 1.46 p o in ts. T h e se e stim a ted
in te rn a tio n a l in fla tio n links tend to be s ta tis tic a lly s ig n ifi
cant and thus more re lia b le than the e stim a te d real
grow th links.
Im p lie d b o n d m a rk e t re a c tio n s
The various lin ks am ong bond m a rke ts and expe cted
inflation rates im p ly the m a rke t re a ctio n s c a lc u la te d in
the m iddle p a rt of Table 7. H ence, if U.S. bond m arke t
p a rtic ip a n ts saw J a p a n e se bond returns rise a p e rc e n t
age point, they w ould in fe r a fa ll in the e xp e cte d J a p
anese in flation rate of 0 .34 of a p o in t (1 d ivid e d by
2.98) and a fall in the expected U.S. rate of n early
0.10 of a p oint (0.34 tim e s 0.29), so th a t U.S. bond
28The control variables are current and four lags each of dom estic
real growth and inflation in the three countries. To allow for the joint
determ ination of inflation, four lags each of real growth and inflation
in the three countries are used as instrum ents.
Table 6
Bond Return Equations Estimated by Two-Stage Least Squares
(D ependent Variable Is Excess Bond Return)
United States
23.39
0.17
(2 45)
(1.45)
Japan
12.86
0.16
U nited K ingdom
(1.47)
(0.92)
-1 8 .7 9
0.47*
( - 1 .4 8 )
(4.02)
(0.54)
(0 20)
(0.21)
(--2 .3 6 )
1.79*
- 3 .7 5 *
3.38*
-0 .4 1
(1.74)
( - 3 .6 4 )
(2.81)
( - 0 .4 3 )
C onstant
Excess stock return
Predicted inflation
First lead
S econd lead
Third lead
Fourth lead
-4 .4 9 *
3.19
- 2 .0 8
- 0 .4 4
Sum of c oe fficients
- 3 .8 3
- 2 .9 8
1.01
0.32
0.26
0.18
0.11
0.21
0.14
R-squared
A djuste d R-squared
( - 2 .3 2 )
d 62)
( - 1 00)
(-0 2 1 )
0.84
0.32
0.40
- 4 .5 4 *
Notes: Instrum ents are excess stock returns, contem poraneous real growth and inflation, and four lags each of d o m estic real growth,
inflation in each of the three countries, percentage change in dollar oil prices, and pe rcentage cha nge in the two do lla r exch ange rates.
T-statistics are in parentheses. A sterisks indica te significance at the 10 percent level.
10 FRBNY Q uarterly Review/Autumn 1991
returns w ould rise 0.37 of a p oint (about 0.10 tim e s 3.8).
T he im plied reaction of the Japanese bond m arket to
the U.S. m arket of 0 .33 is a lm o st as strong. The im plied
re a ctions involvin g the U.K. m arket are much weaker, if
not negative.
A c tu a l b o n d m a rk e t re a ctio n s
The bottom section of Table 7 reports e stim a te s of the
relevant actual bond m arket reactions. To extract the
p a rt of foreign bond m arket returns that reflects m ove
m ents in foreign expected inflation, the e stim ated bond
m arket eq u a tio n s are used to c o n stru ct p redictors of
foreign in fla tio n .29 These pre d icto rs are then su b stitu te d
into the inflation link e qua tio n s, w hich in turn are s u b
stitu te d into the d o m e stic bond return equations. The
d o m e stic bond return eq u a tio n s are then reestim ated,
w ith the foreign bond m arket m ovem ents in e ffe ct h e lp
ing predict d o m e stic infla tio n. The equations co n tro l for
the m ovem ents of fu n d a m e n ta ls other than inflation,
p a rtic u la rly those reflected in excess stock returns. The
e stim a ted co e fficie n ts on the foreign bond m arket va ri
ables then m easure the actual bond m arket reactions to
the relevant m ovem ents in the foreign m arkets.
T he e stim a te s show a sig n ific a n t overreaction by the
J a p anese bond m arket to m ovem ents in the U.S. m ar
ket. Japan ese returns rise 0.54 of a point instead of
0.33 of a p oint in response to a 1 point rise in U.S.
“ S pecifically, we subtract from the excess bond return equations in
Table 6 all the term s but those for inflation
returns. The e stim a te s also show s ig n ific a n t o ve rre a c
tio n s by the U.K. and Ja p a n e se bond m a rke ts to each
o th e r’s m ovem ents. The a p p a re n t o v e rre a ctio n s in v o lv
ing the U.K. m arket sh ould be tre a te d w ith m ore s k e p
ticism , however, because they re fle ct an in e x p lic a b ly
positive e ffe c t of U.K. in fla tio n on U .K. bond returns.
Conclusion
The beh a vio r of sto ck and bond m a rke ts is of co n ce rn
to e co n o m ists because th e se m a rke ts set p rice s a ffe c t
ing the co st of ca p ita l fo r the c o rp o ra te s e c to r and
because excess m a rke t v o la tility m ay lead to fin a n c ial
s tra in s and m a cro e co no m ic in sta b ility. T h o se w ho w o rry
a bout excess v o la tility have reco m m e n d e d such p o lic ie s
as ta xin g se c u ritie s tra n s a c tio n s , ta x in g s h o rt-te rm c a p
ital ga in s more than lo n g -te rm c a p ita l g a in s, and ra isin g
m argin re quirem ents on sto ck p u rc h a s e s .30 T h is stu dy
asks w h e th e r excess c o rre la tio n s a cro ss m a rke ts are a
like ly source of excess vo latility.
The evid e nce presented s u g g e sts som e te n d e n c y by
p a rtic ip a n ts in the U.S. and J a p a n e se sto ck m arkets
and in the U .S., Ja p a n e se , and U.K. bond m a rke ts to
overreact to one a n o th e r’s m a rke t m ovem ents. A lth o u g h
the e stim a te s are im p re cise , th e y in d ic a te th a t the tw o
“ Lawrence Summers and V ictoria Summ ers s u p p o rt the securities
transactions tax; see "When Financial M arkets Work Too Well: A
Cautious Case for a S ecurities Transactions Tax,” Jou rnal of
Financial Services Research, vol. 3 (1989), pp. 261-86. Gikas
Hardouvelis advocates raising m argin requirem ents; see "M argin
Requirements and Stock Market V olatility," this Q uarterly Review,
Summer 1988, pp. 80-89
Table 7
Inflation Links and Implied and Actual Bond Market Reactions
United States
Japan
U nited K ingdom
Effect on dom estic
inflation of inflation in:
U nited States
Japan
United K ingdom
(0.18)
0,43*
0,29*
- 0 .0 6
(0.12)
(0.02)
0.37
0.23
(0.15)
(0.08)
0.42
- 0 .0 7
(0.10)
(0,10)
0.00
(0.03)
0.33
(0.14)
-0 .0 1
(0.09)
- 0 .5 4
2.00*
(1 04)
(0.96)
0.14
-0 .6 8
(0 27)
(0.33)
- 0 .0 8
0.24*
(0.14)
(0.11)
Im plied dom estic bond
market reaction to bond market in:
U nited States
Japan
United Kingdom
A ctual dom estic bond
m arket reaction to bond m arket in:
U nited States
Japan
U nited Kingdom
0.54*
(0.13)
0.22*
(0.11)
Notes: S tandard errors are given in parentheses. Asterisks on inflation links indica te significa ntly positive coe fficients, and asterisks on
m arket reactions indica te s ignifica ntly greater actual over im plied reaction at the 10 pe rcent level,
FRBNY Q uarterly Review/Autum n 1991
11
stock markets move together more closely than would
be expected from the information the markets convey
about future real activity and from the links between
domestic and foreign real activity. The evidence on bond
markets is less consistent, but the three markets seem
to move together more closely than would be expected
from the information they provide on inflation and from
the links between domestic and foreign inflation.
An important limitation of the present study is that it
analyzes market reactions on the basis of average
behavior over the period. In fact, the stock markets
sometimes move very closely together, while at other
times they move independently. When the Japanese
stock market plunged in the spring of 1990, the U.S.
and U.K. stock markets shrugged off the event; by
contrast, in October 1987 the three markets fell as one.
It is as if the markets have “moods,” so that a shock is
sometimes transmitted to other markets more forcefully
than at other times.
If overreaction helps drive market prices away from
fundamental values with some frequency, the resulting
market volatility may pose needless risks to investors
12 FRBNY Quarterly Review/Autumn 1991
and raise the cost of financing in the form of publicly
traded debt or equity.31 International comparisons of the
cost of capital suggest that U.S. corporations are
placed at a competitive disadvantage by the relatively
high costs of equity financing in the United States,
costs that some observers attribute in part to stock
market volatility.32 Worse, the high volatility may make
markets vulnerable to a global crash. In a world where
market prices can occasionally take on a life of their
own, the various markets may at times inflate together
and then burst like enormous bubbles.
31The underlying behavior m ight be c h a ra c te riz e d as a form of
international noise trad ing . B radford D e Long, A ndrei Shleifer,
Lawrence Sum m ers, and R obert W aldm ann show that in g e n e ra l the
presence of noise trad ers can m ake the m arkets too risky for
investors who rely on fun dam entals, so that prices m ay deviate
from fun dam entals for e x te n d e d periods of tim e. See "N oise Trader
Risk in Financial M a rk e ts ,” J o u rn a l o f P o litic a l E c o n o m y , vol. 98
(1990), pp. 7 0 3 -3 8 .
“ R obert N. M cC auley and Steven Zim m er, “ Explaining International
D ifferences in the Cost of C a p ita l,” this Q u a rte rly R e vie w , Sum m er
1989, pp. 7-28 .
Appendix: A Model of Domestic Stock Markets’ Reactions to information from Foreign Markets
This model formalizes a possible role for foreign stock
markets as conveyors of information about real activity
relevant to domestic stock markets. Real activity can be
thought of as determining the cash flows of publicly
traded firms. With stock prices assumed to be the pres
ent values of the future streams of cash flows, we can
write the stock market return as a function of future real
activity growth and of variables tracking the discount
rate:
(A.1)
r\
N1
= Y4 +
k=1
^ 'k Y 't + k
+
V 't<
where z't is the vector of discount rate variables, y ‘t+k is
the kth lead of real activity growth, and v'\ is noise in
returns. The number of leads is N‘.
Real activity in one country could affect real activity in
other countries through international trade. To construct
an index of real activity that reflects stock market behav
ior, we can derive the weights from the lead structure
implicit in the returns equation. The index collapses real
growth in several periods into a single variable:
(A.2)
# =
XT' w*
>
& ky't+k,
param eters in equation
reduces to
( A .3 )
r\
= y Z ' + 8fyi +
A .1 .
The returns equation
v\,
and the co-variation of returns and real activity is mea
sured by a single parameter, 81.
(A.4)
y ‘t = a‘(L)Xt +
.
&Jy { + <4.
where y ‘t is our index of future real activity growth in
country / as of time t, y{ our index of future real activity
growth in country j, ot!(L)x‘t a vector polynomial in the lag
operator L, X, a vector of observable variables helping to
predict y ‘t, and ift the unpredictable part of real activity
growth. The 0,y coefficients measure the co-movement
with real activity growth in other countries after we have
controlled for other macroeconomic variables.
The hypothesis of rational expectations allows us to
assume that stock market investors in country / know
equations A.2, A.3, and A.4. At time t, they observe the
returns r{ and other fundamentals t t in the other coun
tries as well as x\ and 4 in their own country. The
hypothesis gives
(A .5) r 1, = Y 4 + h'a'(L)x‘, + S'
in which investors infer y{ from — (rj
5*
hi come from the
where 0L = 81/8* and 8j
We now measure the links in real activity across coun
tries by estimating
Pf(r{ - 7
'>*< 8i
+ v'„
iA )-
Once 8' and S' are estimated from equation A.1 and p* is
estimated from equation A.4, a regression of stock mar
ket returns in country / on stock market returns in other
countries, as in equation A.5, should yield a coefficient
not significantly different from 8,3'//8/ for country /. Other
wise, we can conclude that international stock market
correlations fail to reflect macroeconomic fundamentals.
FRBNY Q uarterly R eview/Autum n 1991
The Cost of Capital for
Securities Firms in the
United States and Japan
by Robert N. McCauley and Steven A. Zimmer
Recent studies of international differences in capital
costs have focused on industry and banks. In the 1980s
U.S. firms seemed to be losing ground internationally,
whether measured by semiconductor trade, industrial
investment, manufacturing trade, or market share in
U.S. commercial lending. This slipping competitiveness
prompted economists to investigate whether U.S.
industry and banks were laboring under a cost of capital
disadvantage.
By contrast, the cost of capital for U.S. securities
firms received little attention during the last several
years because these firms appeared to perform more
creditably. They defended their home turf, mounted
major expansions into foreign markets, and staked out
market share and profit in trading government bonds
and equities abroad.1 U.S. securities firms invested
much more abroad that their foreign competitors
invested in the United States. In other industries, espe
cially banking, foreign direct investment into the United
States dominated U.S. investment abroad (Table 1).2
’ The firms showed best results in trad ing Jap anese and G erm an
governm ent bonds in Tokyo and London, respectively, and in
trad ing J a p a n e se equities and equity derivatives in Tokyo. See John
J. R uocco, M aureen LeB lanc, and Patrick D ignan, "C om petitiveness
in G overnm ent Bond M a rk e ts ,” and M artin Mair, M ichael Kaufman,
and Steven Saeger, “C om petitiveness in Equity M a rk e ts ,” in
In te rn a tio n a l C o m p e titiv e n e s s o f U.S. F in a n c ia l F irm s : A S ta ff S tu d y
(N ew York: Federal R eserve Bank of New York, 1991), pp. 130-72.
2Perhaps as a result, the pu blic policy discussion of the securities
industry has focused on ensuring that U .S. firms enjoy equal
a c c e s s to foreign financial m arkets. S ee, for instance, D ep artm ent
of the Treasury, N a tio n a l T re a tm e n t S tudy, 1990 U p d a te , pp. 225-41;
Staff of the Board of G overnors of the Federal R eserve System and
the Federal Reserve Bank of New York, R e p o rt o n Im p le m e n ta tio n o f
th e P rim a ry D e a le rs A ct, m em orandum , August 15, 1989; and
14 FRBNY Quarterly Review/Autumn 1991
Reversing the procedure of earlier studies, this article
takes the respectable performance of U.S. securities
firms as its rationale for exploring cost of capital differ
ences between countries. If U.S. firms achieved some
degree of success in spite of higher capital costs, then
this disadvantage is clearly not a decisive one. But if
the disadvantage faced by securities firms is smaller
than that faced by U.S. industry and banks, then capital
costs may help to explain differences in competitive
outcomes.
Our investigation begins with a comparison of the
capital costs faced by U.S. and Japanese securities
firms in the 1982-91 period. We measure the cost of
capital to five U.S. and four Japanese securities firms
as the multiple that their respective stock exchanges
assigned to the economic earnings of the firms. Our
findings indicate that U.S. equity investors placed a
lower value on a given stream of earnings of U.S.
securities firms than Japanese equity investors placed
on a comparable stream of earnings of Japanese secu
rities firms. As a result, U.S. securities firms needed to
earn more on a given sum of capital underpinning any
line of business.
The gap in valuation of securities firms’ earnings in
the New York and Tokyo stock exchanges nevertheless
appears to be narrower than the gaps we found
between U.S. and Japanese industries and banks in our
own earlier studies of cost of capital differences.3 If
F o o tn o te 2 (c o n tin u e d )
"Japan, U .K . and Sw itzerland: Prim ary D ea le rs A ct U p d a te ,"
m em orandum , D e c e m b e r 3, 1990.
3R obert N. M cC auley and Steven A. Zim m er, "Explaining
International D ifferences in the C ost of C a p ita l," this Q u a rte rly
U.S. se cu ritie s firm s c a rry a s m a lle r d is a b ility in ca p ita l
co sts than o th e r U.S. firm s, then it m akes sense that
any adva ntage s in oth e r d im e n sio n s of co m p e titio n ,
such as e xpe rien ce w ith derivative products or a p p lic a
tion of technology, could be d e cisive in overall c o m p e ti
tive outcom es.
In see king to explain ca p ita l cost diffe re n ce s, we
e m p ha size ge neral fa cto rs a cco u n tin g fo r a lower J a p
anese cost of e q u ity in the la tte r 1980s. These include
h ig h e r h o u s e h o ld s a v in g s and s m o o th e r e c o n o m ic
grow th.
O u r a n a ly s is also c la rifie s w h y the gap betw een
m easured e q u ity costs in New York and Tokyo m ight be
s m a lle r fo r se cu ritie s firm s than fo r banking and other
in d u strie s . On the one hand, Japanese se c u ritie s firm s’
co st of e q u ity may be h ig h er than that of Ja panese
n o n fin a n cia l firm s or banks because the m arket per
ceives a re la tively severe th re a t to the se c u ritie s firm s’
revenues and e arnings in the ongoing trend toward
fin a n cial d eregulation . On the o th e r hand, the lower
e q u ity costs for U.S. s e cu ritie s firm s relative to other
U .S. com p a n ie s may be influenced by the choice of
s a m p le p eriod. The m id-1980s were boom years for the
s e c u ritie s business, and U.S. investors, seized w ith the
g row th p o s s ib ilitie s created by the fin a n cial innovators
and e nginee rs of Wall S treet in in cre a sin g ly g lobal m ar
kets, m ay have priced U.S. s e c u ritie s firm s’ e a rnings at
a prem ium .
Footnote 3 (continued)
Review, vol. 14 (Summer 1989), pp. 7-28; and Steven A. Zimmer
and Robert N. McCauley, "Bank Cost of C apital and International
C o m p e titio n ,” this Q uarterly Review, vol. 15 (Winter 1991),
pp. 33-59.
Measuring the cost of capital
S e cu ritie s firm s p rovide p ro d u cts and e n gage in a c tiv i
ties of varying risk a g a in s t w hich th e y m ust hold e q u ity
ca p ita l. The required return on th is e q u ity c a p ita l w ill be
im p o rta n t in d e te rm in in g the c o m m issio n or fee th a t a
firm m ust ch a rg e fo r a s e rv ic e or the return it m ust earn
a rb itra g in g m a rke ts or in ve stin g on its own a cco u n t. We
define the co st of ca p ita l fo r a s e c u ritie s firm as the
m inim um required fee the firm m ust ch a rg e , or the
return it m ust m ake, to cover the required return on the
e q u ity ca p ita l a llo tte d to an activity.
O ur d e fin itio n of cost of ca p ita l fo r a s e c u ritie s firm ,
like our d e fin itio n of the co st of ca p ita l fo r a bank,
excludes d ebt co sts. T he reason fo r th is e xclu sio n is
that in te rn a tio n a lly active s e c u ritie s firm s face s im ila r
borrow ing costs. For in sta n ce , J a p a n e se firm s’ s u b s id i
aries in New York sh o u ld be able to fin a n ce th e m se lve s
at m uch the sam e rates as U.S. firm s. Indeed, th is
arg u m e nt may be more firm ly g ro u n d e d fo r s e c u ritie s
firm s than fo r banks. The m ost im p o rta n t so u rce of
borrow ed fu n d s fo r a large s e c u ritie s firm is the sale
and fo rw a rd re p u rc h a s e of s e c u ritie s . T he se cu re d
nature of th is fin a n c in g te c h n iq u e le s s e n s c re d ito r
dem ands fo r su b s ta n tia l d iffe re n c e s in in te re s t rates
b a s e d on th e c r e d itw o r t h in e s s o f th e b o rro w e r.
R e p u rc h a s e a g re e m e n ts have g e n e ra lly p e rm itte d
se c u ritie s firm s in the U nited S tates to fin a n ce th e m
selves at rates below in te rb a n k ra te s .4
O ur d e fin itio n of co st of c a p ita l fo r s e c u ritie s firm s
fo llo w s the d e fin itio n of bank c o st of c a p ita l p re se n te d in
our e a rlie r stu d ie s, and we w ill p roceed in a s im ila r
fashion. The first step in asse ssin g c o st of c a p ita l differ4For the last year, the overnight repurchase rate has on average
exceeded the federal funds rate in the U S money market.
Table 1
Foreign Direct Investment Flows into and out of the United States, 1985-89
(In B illions of Dollars)
Total
M anufacturing
Inflow
Outflow
232.9
90.5
2.6
110.4
45.2
2.4
Ratio of Inflow to Outflow
Banking
9.1
0.1
109.6
Finance (except banking)
6.9
12.6
0.5
Source: "Foreign Direct Investm ent in the United States" and “ U.S. Direct Investment A b ro a d ,” Survey o f C urrent Business, vol. 70 (A ugust
1990), pp. 54, 55, 97, 98.
Notes: M anufacturing, banking, and finance do not sum to total. Direct investment flows relating to the N etherlands A ntilles and to the U.K.
C aribbean Isles are sub tracte d from U.S. direct investm ent abroad and foreign direct investm ent in the U nited States, respectively. These
adjustm ents are made because outflows to the N etherlands A ntilles in this period essentially reflect repaym ents of E urobonds sold through
shell finance affiliates and because outflows to the U.K. C aribbean Isles reflect onle ndin g of the proceed s of com m ercial p a p e r and bond
sales by U.S. finance affiliates of nonfinancial foreign corporations via tax havens in the C aribbean. The removal of these flows reduces
cum ulative U.S. dire ct investm ent outflows by $20.3 billion and boosts foreign direct investm ent inflows by $2.2 b illion for both the total and
the finance com ponent.
FRBNY Q uarterly Review/Autum n 1991
15
ences is to estimate the required return on equity— the
“cost of equity”— to securities firms in the United States
and Japan. Our analysis of a small sample of key
publicly traded firms suggests that Japanese securities
firms enjoy a substantial cost of equity advantage over
U.S. firms.
The second step is to show how differences in the
cost of equity translate into differences in the cost of
capital. Because securities firms, unlike banks, do not
have uniform international capital requirements, this
step requires care. One complication is that both
observed and required shareholder-equity-to-asset
ratios of Japanese securities firms are higher than
those of U.S. securities firms.
The co st of equity
We define the cost of equity as the ratio of a firm’s
sustainable profits to the market value of its equity. We
cannot observe sustainable profits, but we can observe
reported profits for a sample of firms and make adjust
ments to them. In addition to making reported profits
better reflect economic profits, these adjustments make
the stated profit measures internationally comparable.
Our sample of firms for the United States is neces
sarily limited to those whose shares have been publicly
traded throughout the sample period. First Boston and
Shearson-Lehman are thus excluded because their
public shareholders were bought out by their respective
parents, Credit Suisse and American Express; Goldman
Sachs, Drexel, and Prudential-Bache are excluded by
virtue of their private ownership. That leaves Merrill
Lynch, Morgan Stanley, and Salomon Brothers of the
“bulge bracket,” or lead underwriter, firms and Bear
Stearns and Paine Webber of the remaining top ten
firms. The selection of Japanese securities firms is
quite obvious in light of their dominant status: Daiwa,
Nikko, Nomura, and Yamaichi, the so-called Big Four.
The sample period runs from 1982 to 1991. The nine
and one-half fiscal years covered cannot be syn
chronized across the two countries. For all U.S. firms
except Bear Stearns and Paine Webber, fiscal years
correspond to calendar years and the 1991 observation
covers only the first half.5 For the Japanese firms, the
half year covers October 1988 to March 1989, an
accounting period that permitted their fiscal years to be
aligned with general practice in Japan. Because Bear
Stearns and Morgan Stanley made their initial public
offerings in October 1985 and March 1986, respectively,
5For B ear Stearns, d a ta for fiscal years ending in April through 1987
and in June from 1988 on are a g g re g a te d with the other firms' data
for the previous D ecem ber. For Paine W ebber, d a ta for the fiscal
year ending in S e p te m b e r are a g g re g a te d with the other firms’ data
for the following D ec e m b e r through 1986; in 1987 the firm switched
to fiscal years ending in D ecem ber.
16 FRBNY Quarterly Review/Autumn 1991
1985 is the first sample year for each (Morgan Stanley’s
public offering price is taken to be its December 1985
price). Altogether, this study’s cost of equity calculations
rely on forty-three observations of U.S. securities firms’
share prices, earnings statements, and balance sheets
and forty corresponding observations for Japanese
securities firms.
We adjust reported profits for the following:6
d epreciation— stated earnings are lowered to offset
the upward bias introduced when depreciation
expenses are based on historical cost during a
period of inflation;
equity/inflation— the increase in the nominal value
of shareholder equity necessary to maintain the
real value of shareholder equity is subtracted from
stated profits;
crossholding— the undistributed profits associated
with equity shares held by Japanese firms are con
solidated into income; and
restru ctu rin g c h a rg e s — U.S. firms’ restructuring
charges are spread out over three years.
The crossholding adjustment is performed for Jap
anese securities firms but not for U.S. securities firms
even though both hold significant amounts of equities.
The reason for the asymmetry in this adjustment is that
U.S. firms mark their equities to market, while Japanese
firms do not. Over time, U.S. firms’ marked-to-market
equity values reflect retained earnings on equity hold
ings insofar as these earnings are embodied in share
prices. Japanese firms not only value their equity hold
ings at historical cost, but also hold and rarely realize
large and growing stakes in their investment accounts
for strategic purposes. It is this combination of low
turnover and historical cost accounting that requires the
crossholding adjustment.
Taken together, the adjustments performed on the
raw observed ratios of after-tax earnings to market
capitalization narrow the differences between the U.S.
and Japanese firms significantly (Table 2 )7 Making
«Com pare the adjustm ents to bank profits in Z im m er and M cCauley,
“Bank Cost of C a p ita l,” pp. 3 6 -4 2 .
^The rows do not sum for U.S. firms in the years 1984, 1 9 8 8 -9 0 , and
the average owing to our constraining the cost of e quity to be non
negative. This constraint a dds 0 .4 p e rc e n ta g e point to the average
cost of equity. Excluding firm -years of com p u te d negative cost of
equity would yield an average cost of equity of 8 .6 pe rce n t.
Treatm ent of the industry as a single firm — a d d in g earn in g s across
firms in a given year and c o m paring the total with sum m ed m arket
c a p ita liza tio n s — results in an ave ra g e cost of e quity of 7 .4 p ercen t.
a llo w a n c e s fo r in f la t io n ’s e ro s io n o f d e p re c ia tio n
e xp e n se s and of s h a re h o ld e rs ’ e q u ity re d u ce s U.S.
se c u ritie s firm s’ e a rn in g s to a gre a te r extent than th e ir
Ja p anese co u n te rp a rts ’ ea rn in g s, la rg e ly because of
the h ig h er rate of inflation e xperienced in the U.S.
econom y in the sam ple period. S preading out U.S.
firm s’ e x tra o rd in a ry reserves should in p rin cip le sim ply
sm ooth th e ir cost of e q u ity but in practice th is a d ju s t
m ent in te ra cts w ith share price m ovem ents to w iden the
gap a bit.
The cro ssh o ld in g a d ju stm e n t narrow s the gap s u b
stantially, a fin ding in line w ith previous w ork on d iffe r
ences in e q u ity va lu a tion s in the tw o m a rke ts.8 The
cro s sh o ld in g a d ju stm e n t fo r Japanese se c u ritie s firm s
in the late 1980s is more c o n s iste n t than fo r Japanese
b a n ks, e s p e c ia lly c ity b a n ks, in the sam e p eriod.
B ecause the city banks cam e u nder pressure to m eet
new in te rn a tio n a l ca p ita l s ta n d a rd s and responded in
»See James M. Poterba, "C om parin g the Cost of C apital in the
United States and Japan: A Survey of M ethods," this Quarterly
Review, vol. 15 (W inter 1991), pp. 20-32, and references contained
therein.
p a rt by re a lizin g m assive g a in s on c ro ssh e ld shares,
the cro ssh o ld in g a d ju s tm e n t a c tu a lly su b tra c te d e a rn
ings in the three fisca l years to M arch 1990.9 In the
sam e period, Ja p a n e se s e c u ritie s firm s, a ctin g like th e ir
co rp o ra te cu sto m e rs, te n d e d to e sch e w re a lizin g ga in s
on e q u itie s in th e ir in ve stm e n t p o rtfo lio s — and th e re b y
avoided the ta xe s a sso cia te d w ith such re a liza tio n s.
The resulting co st of e q u ity se rie s show som e v o l
a tility but c a rry a c le a r m e ssa g e (C h a rt 1). The J a p
anese se c u ritie s firm s in o u r sa m p le face an average
cost of e q u ity of 5.1 p e rce n t in the sa m p le p e riod as
ag a in st 7.8 p e rce n t fo r the U.S. s e c u ritie s firm s. Such a
diffe re n ce is u n like ly to be w ith o u t im p lic a tio n s fo r in te r
n ational co m p e titio n . At the sam e tim e , th e adva n ta g e
of J a p a n e s e s e c u ritie s firm s is s m a lle r th a n th a t
enjoyed by Ja p a n e se b anks (3.1 p e rc e n t co m pared w ith
11.9 pe rce n t fo r U.S. b a n k s )10 or J a p a n e se in d u s tria l
firm s (4.5 pe rce n t co m pared w ith 11.2 p e rc e n t fo r U.S.
aZimmer and McCauley, "B ank Cost of C a p ita l,” p. 40.
10Zim mer and McCauley, "B ank Cost of C a p ita l,” p. 42.
Table 2
Summ ary of Adjustm ents to Cost of Equity
(Cross-Firm Averages in Percent)
tt tp * lr m s
10.01
11.57
3.67
8.30
10.74
13.55
12.37
8.28
3.31
19.30
1982
1983
1984
1985
1986
1987
1988
1989
1990
1991
Averages
Japanese Firms
A djustm ents
Profit/
Market
C apitalization
10.11
Profit/
M arket
C apita lization
1982
1983
1984
1985
1986
1987
1988
1989
1990
1991
Averages
E quity/
D epreciation
Inflation
- 0 .9 8
- 0 .9 7
- 1 .0 0
- 0 .6 9
- 0 .7 3
-1 .5 7
- 1 .4 4
- 1 .3 9
-1 .7 1
-1 .2 1
-1 .1 7
CrossH olding
R estructuring
Cost of
Equity
- 1 .7 2
- 1 .5 5
- 1 .7 7
-1 .1 4
- 1 .2 3
- 1 .9 6
- 2 .2 5
- 2 .3 5
-2 .1 6
- 1 .9 4
0
0
0
0
0
0
0
0
0
0
0.21
0 02
0.94
- 0 .2 5
- 0 .1 7
0.78
- 0 .2 6
0.97
3.02
- 2 .3 4
7.51
9.07
1.90
6.22
8.62
10.80
8.57
6.79
4.77
13.82
-1 .8 1
0
0.29
7.81
A djustm ents
D epreciation
Equity/
Inflation
CrossH olding
R estructuring
5.12
5.43
6.31
5.28
3.81
3.98
3.61
4.51
7.37
3.06
- 0 .1 0
- 0 .0 8
-0 .0 8
- 0 .0 5
- 0 .0 2
- 0 .0 2
-0 .0 2
- 0 .0 2
- 0 .0 3
- 0 .0 5
- 0 .7 7
- 0 .2 2
- 0 .7 2
- 0 .6 5
- 0 .0 4
- 0 .0 4
- 0 .1 6
- 0 .4 8
- 1 .1 6
—1.17
1.66
1.18
1.15
0.84
0.43
0.56
0.48
0.34
0.60
0.76
0
0
0
0
0
0
0
0
0
0
4.85
- 0 .0 5
-0 .5 4
0.80
0
Cost of
Equity
5.91
6.30
6.66
5.42
4.18
4.47
3.91
4.35
6.78
2.60
5.06
Sources: Annual reports; Toyo Keizei Inc., Japan Com pany H andbook; Federal Reserve Bank of New York staff estim ates.
FRBNY Q uarterly Review/Autum n 1991
17
in d u s try ).11
T he se fin d in gs are c o n s is te n t w ith m anagers’ actions
in the 1980s. C o n sid e r the m atch betw een the observed
p attern of fu n d -ra isin g in the e q u ity m arkets and the
pattern of a b solute and relative advantage in equity
co sts of U.S. and Jap a n e se firm s across industry. First,
the a b so lute advantage of Japanese firm s in equity
co sts in 1985-89 was reflected in the c o n tra s tin g behav
ior of n on fin a n cia l co rp o ra tio n s in the U nited S tates and
Japan: U.S. no n fin a n cia l c o rp o ra tio n s retired (net) $500
b illio n w h ile th e ir Jap a n e se co u n te rp a rts issued ¥ 11.4
trillio n , or $80 billio n , n e t.12 S econd, p a rtic u la rly low
e q u ity costs help e xplain w hy Ja panese banks raised
m o re e q u ity th a n a n y o th e r in d u s try in J a p a n ,13
alth ough ca p ita l regu la tio n also played a role. (U.S.
banks were co n stra in e d by regulation from jo in in g th e ir
co rp o ra te cu sto m e rs in share re p urchases.) Finally, the
"M c C a u le y and Zimmer, "E x p la in in g ," p. 12.
12M argaret H astings P ickering, “A Review of C orporate R estructuring
Activity, 1980-90," Board of Governors of the Federal Reserve
System Staff Study, no. 161, May 1991; and Bank of Japan, flow of
funds data in Econom ic S tatistics Monthly.
13Robert Zielinski and Nigel Holloway, U nequal E quities: Power and
Risk in J a p a n ’s Stock Market (Tokyo: Kodansha International, 1991),
pp. 184-86.
18 FRBNY Q uarterly Review/Autum n 1991
U.S. se c u ritie s in d u s try stood out as an is s u e r of new
eq u ity in the 1980s: B ear S te a rn s, M organ Stanley, and
o th e rs m ade in itia l p u b lic o ffe rin g s ,14 and G o ld m a n
S achs, S h e a rso n L e h m a n , and P aine W e b b e r so ld
e q u ity to S u m ito m o Bank, N ippon Life, and Y asuda
Trust, respectively. Moreover, the issu e s of th e U.S.
firm s clu ste re d in the m id-1980s, w hen o u r m easured
co st of e q u ity was m ost favorable.
A llo c a tin g e q u ity to fin a n c ia l a c tiv itie s
The required fee or return on a given p ro d u c t o r a c tiv ity
is de te rm in ed by the required return on e q u ity and by
the am ount of e q u ity a llo tte d to th e p ro d u c t o r activity. If
both a U.S. and a J a p a n e se s e c u ritie s firm a llo t the
sam e e q u ity to a given p ro d u ct or activity, th en the
required fee or return w ill be an equal fra c tio n of each
firm ’s cost of equity. A ny d iffe re n ce in th e co s t of e q u ity
is then reproduced in the c o st of c a p ita l fo r the p ro d u ct
or activity.
If U.S. s e c u ritie s firm s lever up th e ir s h a re h o ld e rs’
e q u ity w ith more a sse ts than J a p a n e se s e c u ritie s firm s,
it m ight seem safe to co n c lu d e th a t th e y a llo t less e q u ity
to a given a c tiv ity than d oes th e ir c o m p e titio n . T his
c o n clu sio n does not follow , however. At the o u ts e t, it is
easy to overstate the d iffe re n ce in le ve ra g e b ecause
U.S. acco u n tin g sta n d a rd s leave s e c u ritie s sold un der
agreem ents to re p u rch ase on the b a la n ce s h e e t, w h ile
J a p a n e s e a c c o u n tin g ta k e s th e m o ff. E ve n if o n e
a d ju sts fo r th is d iscrepancy, however, J a p a n e se s e c u ri
tie s firm s rem ain less leveraged, w h e th e r m easured at
book or m arket value (Table 3, lin e s 5 and 7).
To som e extent, Ja p a n e se s e c u ritie s firm s’ low er le v
erage o ffse ts the h ig h e r risk of th e ir a sse ts. By h is to r
ical acco u n tin g , U.S. and Ja p a n e se s e c u ritie s firm s
have 3 to 4 p e rce n t of th e ir asse ts invested in e q u itie s.
By m arket value, however, the J a p a n e se firm s have
a lm ost tw ice the e q u ity (Table 3, lin e s 1 and 2). S till, a
d iffe re n t m ix of e q u itie s in a sse ts d oes not p ro vid e a fu ll
a ccount of the le verage d iffe re n ce . If e q u ity h o ld in g s
are s u b tra c te d fro m s h a re h o ld e rs ’ e q u ity, J a p a n e s e
firm s rem ain s ig n ific a n tly le ss le v e ra g e d (T able 3,
line 6).
Even the rem aining d iffe re n ce in le ve ra g e need not
im ply that J a p a n e se firm s a llo c a te m ore e q u ity to a
given a c tivity in a g ive n m arket. T he lack of in te rn a
tio n a l c o o rd in a tio n in the re g u la tio n of th e s e c u ritie s
business m ust be re co g n ize d . S e c u ritie s firm s in Japan
m ust hold s h a re h o ld e rs’ e q u ity eq u a l to 10 p e rce n t of
assets. D espite the a p p lic a tio n of th is s ta n d a rd to both
d o m e stic and fo re ig n firm s o p e ra tin g in Tokyo, U.S.
firm s have co m p la in e d th a t so high a c a p ita l require14Chris J. M uscarella and Michael R. Vetsuypens, "A S im ple Test of
Baron's Model of IPO U n d e rp ric in g ,” Jou rnal o f F inancial
E conom ics, vol. 24 (1989), pp. 125-35.
m ent is re strictive .15 W hatever the w eight of th is c o n te n
tio n , U.S. and Ja pane se firm s in Tokyo require the sam e
e q u ity in a given a ctiv ity to, say, a rb itra g e betw een cash
and fu tu re s m arke ts in stock.
In New York the s u b s id ia rie s of Japanese s e cu ritie s
firm s are not bound by Jap a n e se ca p ita l s ta n d a rd s but
need o n ly s a tis fy U.S. Treasury and S e cu ritie s and
E xchange C o m m ission ca p ita l requirem ents. Indeed,
the Big F o u r’s U.S. s u b sid ia rie s operate w ith leverage
m ore like that of U.S. firm s than that of th e ir parents
(Table 4; Table 3, line 5). W hen in New York, th e se firm s
do as New Y orkers do.
T he overall d iffe re n ce in leverage, therefore, can be
a scrib ed la rg e ly to d iffe re n ce s in capital requirem ents
and in the g e o g ra p h ica l m ix of business. Indeed, capital
15Foreign securities firm s have faced the same capital requirements
as Japanese firm s since the mid-1980s. See U.S. Treasury, N ational
Treatment Study: 1986 U pdate, p. 78; and Report on Primary
Dealers Act, A ttachm ent 3, Summ ary of Public Comm ents, pp. 7-8.
requirem ents b e tte r e xplain the d iffe re n c e s in le verage
than the degree of leverage of e ith e r U.S. or Ja p a n e se
firm s since firm s in both c o u n trie s te n d to hold ca p ita l in
excess of re q uirem ents.
S im ila r leverage w ith in a m a rke t m akes fo r c o st of
ca p ita l d iffe re n ce s th a t re fle ct co st of e q u ity d iffe re n ce s.
G iven th a t a 10 p e rc e n t e q u ity -to -a s s e t ra tio is required
in Japan, if U.S. firm s face a required return on e q u ity of
10 pe rce n t w h ile J a p a n e se firm s face a required return
of 5 percent, then the fo rm e r need to earn 1 p e rc e n t on
assets in Tokyo w h ile the la tte r can get away w ith
1/2 p e rc e n t. If th e c a p ita l re q u ire m e n t w o rk s o u t to
2 p e rce n t in the U.S. m arket, th e n th e U.S. firm needs
to earn 20 basis p o in ts per annum on its a sse ts w hile
the Ja p a n e se firm needs to earn o nly 10 b asis p o in ts. In
th is m anner the c o st of e q u ity d iffe re n c e s c a rry over
into cost of ca p ita l d iffe re n ce s.
Explaining cost of capital differences for
securities firms
The fin d in gs so fa r raise tw o q u e s tio n s : W hy do J a p
anese s e c u ritie s firm s cla im an a d va n ta g e in the co s t of
e q u ity over th e ir U.S. c o u n te rp a rts ? A nd w h y is the
advantage s m a lle r than th a t fo u n d fo r J a p a n e se nonfinancial firm s and banks?
Table 3
Selected Balance Sheet Characteristics of
U.S. and Japanese Securities Firms
(Percent)
Japanese
Firms
U.S.
Firms
Equity holdings in perspective
1. Equity portfolio/total assets
(security holdings at book value)
3.0
3.7
2. Equity portfolio/total assets
(security holdings at market value)
6.9
3.7
26.3
87.4
44.5
87.4
11.7
4.3
3. Equity portfolio/shareholder equity
(security holdings at book value)
4. Equity portfolio/shareholder equity
(security holdings at market value)
Leverage
5. Shareholder equity/total assets
(security holdings at book value)
6. Shareholder equity less equity holdings/
total assets less equity holdings
(security holdings at book value)
7. Shareholder equity/total assets
(security holdings at market value)
9.0
14.7
0.38
4.3
Sources: Annual reports; Toyo Keizai Inc., Japan Com pany
H an d b o o k; Federal Reserve Bank of New York staff estim ates.
Notes: Data are averages for 1986-89. Assets for Japanese
firm s includ e gensaki and repurchase agreem ents. For Daiwa,
Nikko, and Yamaichi, the market value of securities p o rtfolio is
estim ated from net assets at market value less unconsolidated
shareholder equity from the Japan Com pany Handbook. For
Nomura, whose annual reports detail the market value of
secu rities, this difference overstates unrealized gains on secu
rities by an average of 6 percent, with a range of 1 to 9
percent. U nrealized gains on D aiw a’s, Nikko's, and Yam aichi’s
eq uity ho ld ings alone are estim ated as the product of the
diffe ren ce above and .905.
M a c ro e c o n o m ic e x p la n a tio n s fo r U .S .-J a p a n e s e
d iffe re n c e s 16
Ja panese se c u ritie s firm s share in the re la tive ly low
e q u ity costs th a t c h a ra c te riz e d the w h o le Ja p a n e se
co rp o ra te s e cto r in the la tte r 1980s. T h e se low co sts
are tra ce a b le in la rg e p a rt to m a c ro e c o n o m ic fa cto rs.
E ve n th o u g h th e in te r n a tio n a l m o b ility o f c a p ita l
increased in the 1980s (as e vid e n ce d by s u b s ta n tia l
c ro s sb o rd e r tra n s a c tio n s in e q u ity), c a p ita l co s ts were
fa r from e q u a lize d a cro ss c o u n trie s and n a tio n a l fa cto rs
still played a p re d o m in a n t role. In Japan, h ig h e r h o u s e
hold savings m ade fo r low er e q u ity co sts. In a d d ition ,
sm o o th e r grow th in Japan, re su ltin g in p a rt from s u c
ce ssfu l m a cro e co no m ic policy, m eant low er risk in p ro f
its, and low er risk in profits m eant low er c o st of equity.
S a fe ty n e t d iffe re n c e s b e tw e e n U.S. a n d J a p a n e se
s e c u ritie s firm s
We have argued e lse w h e re th a t the risk faced by in v e s
tors in the e q u ity of banks d e p e n d s on th e nature of the
safety net provided by o ffic ia ls of various c o u n trie s to
th e ir banks. Investors in the shares of s e c u ritie s firm s
also face s y s te m a tic a lly d iffe re n t risks ow ing to n a tio n a l
diffe re n ce s in s a fe ty -n e t c h a ra c te ris tic s . In p a rticu la r,
investors in J a p a n e se s e c u ritie s firm s have m ore reason
16M acroeconom ic explanations of U .S .-Japanese cost of ca p ita l
differences are discusse d at length in M cCauley and Zimmer,
"E xp la in in g ,” pp. 16-20.
FRBNY Q uarterly Review/Autum n 1991
19
to su p pose th a t th e ir d ow nside risk is s u b s ta n tia lly
le sse ned by the p o s s ib ility of governm ent in te rve n tio n
than do investors in the shares of U.S. se c u ritie s firm s.
P otential investo rs try in g to im agine the w orst that
m ight happen to the value of th e ir shares in a se cu ritie s
firm are liable to co n ju re up diffe re n t sce n a rio s for
losses in Ja pane se and U.S. se c u ritie s firm s. If they are
co n sid e rin g in vesting in shares of a Japanese firm , they
m ay w ell call to m ind the d istre ss of Yam aichi S e cu ritie s
in the 1960s; if the y are co n sid e rin g investm ent in a
U .S. firm , th ey may re a d ily recall the ba n kru p tcy filing of
Drexel in 1990.
The e sse n tia l fe atu re s of Y a m a ich i’s d iffic u ltie s may
be re la te d b rie fly : lo s s e s on sto c k m a rke t h o ld in g s
im paired the firm ’s ca p ita l; cu sto m e rs w ithdrew liq u id ity ;
the B ank of Japan w orked w ith the M in is try of F inance
to p ursue a rescue plan in volving la rg e ly unsecured
advances by the B ank of Japan; eventually Yam aichi
recovered and repaid the loans over fo u r ye a rs.17
The e sse n tia l featu re s of D re xe l’s d iffic u ltie s may be
related w ith equal bre vity: losses on ju n k bonds and
bridge loans im paired the firm ’s ca p ita l; providers of
w h o le sa le fun d in g w ith d re w liq u id ity ; the S e cu ritie s and
E x c h a n g e C o m m is s io n w o rk e d w ith th e F e d e ra l
R eserve Bank of New York to achieve an o rd e rly reduc
tion of the bala nce sh e e ts of the registered brokerde a le r and the g o vern m en t se cu ritie s su b s id ia rie s ; the
firm so u g h t pro te ctio n from its cre d ito rs u nder C hapter
11 of the B a n kru p tcy C ode; and the fate of unsecured
17A ppe ndix to statem ent of E. Gerald Corrigan, President of the
Federal Reserve Bank of New York, in D eposit Insurance Reform
a n d F inancial M odernization, H earings before the Senate Comm ittee
on Banking, Housing, and Urban Affairs, 101st Cong., 2d sess
(W ashington, D.C.: Governm ent Printing Office, 1990), pp 82-86,
reprinted as "How Safety Nets W ork," Central Banking, Autumn
1990, pp. 61-63.
creditors, like th a t of h o ld ers of the firm ’s (u n tra d e d )
equity, rem ains u n c le a r at th is ju n c tu re .18
The s trikin g co n tra s t betw een th e se tw o e p is o d e s, of
course, p rovides no c e rta in g u id e to how a tro u b le d
s e cu ritie s firm w ould be han d le d in the fu tu re . C e rta in ly
the contexts of the o fficia l a ctio n s d iffe re d : g e n e ra lly
low share p rices reflected g e n e ra l e co n o m ic w e a kness
in Japan in 1962, w h ile D re xe l’s d iffic u ltie s cam e late in
an e co n o m ic upsw ing. N e ve rth e le ss, m a rke t p a rtic i
pants may well view the e q u ity of a m a jo r U.S. s e c u ri
ties firm as s u b je ct to one more risk than th a t of a m ajor
Japanese se c u ritie s firm .
M a rk e t m easures o f ris k
M arket m easures of risk show Ja p a n e se s e c u ritie s firm s
to be, if anyth in g , a bit ris k ie r than th e ir U.S. c o u n te r
parts. B ecause Ja p a n e se s e c u ritie s firm s are m uch less
leveraged than U.S. firm s, th e y sh ould e x h ib it low er
stock betas, given equal ris k in e s s of a s s e ts .19 B ut in
fact the sto ck betas of J a p a n e se s e c u rity firm s have
averaged 1.46 over the p e riod 1987-91, as co m pare d
w ith 1.29 fo r U .S. s e c u ritie s firm s o ve r th e p e rio d
1986-91, and the d iffe re n ce is even m ore s trik in g for
18C hristopher Byron, “ Drexel's Fall: The Final D ays," New York, M arch
19, 1990, pp 32-38.
19S tarting with the relationship
ba = w x be + ( 1 - w ) x bd,
where
ba =
w =
be =
0d =
asset beta
equity/asset ratio
equity beta
bond beta,
we have dt>e/dw = w ~ 1 x { b a - be + [(1 - w) x dt>d/dw]}.
Given that ba and d£>d/dw are small and of opposite sign, we have
dbe/dw < 0. If we further assum e that bond betas are ge nera lly
n egligible, we have dbe/dw = - b j w.
Table 4
Shareholders’ Equity as a Share of Total Assets for U.S. Affiliates of Japanese Securities Firms
(Percent)
Date
Daiwa
S eptem ber 1985
S eptem ber 1986
S eptem ber 1987
S eptem ber 1988
M arch 1989
M arch 1990
M arch 1991
1 38
1.05
0.92
0,99
0.85
1.17
Period average
1.06
Nikko
Yamaichi
Average
2.96
2.90
2.22
1.64
1.90
5.91
2.30
1.48
2.00
1.57
1.38
1,41
Nomura
2.45
1.84
0.90
0.92
0.96
5.91
1.84
1.99
1.92
1.42
1.20
1.36
2.32
2.29
1.41
1.77
Source: Annual reports.
Note: For Nomura and Yamaichi, M arch figures for 1989 and 1990 are averages of S eptem ber 1988 and S eptem ber 1989 and S eptem ber
1989 and S eptem ber 1990, respectively.
20 FRBNY Q uarterly Review/Autumn 1991
years oth er than 1990 (Tables 5 and 6).
F in a n c ia l d e re g u la tio n a n d the in s e c u rity o f Ja panese
s e c u ritie s firm s’ e a rn in g s
Investors in the Big F ou r’s shares may well perceive a
risk of more con cern than b a n kru p tcy or the shares’
e xa ggerated response to g e neral m arket m ovem ents.
P rospective dere g u la tio n is w id e ly viewed as a threat to
the firm s’ revenues, and the risk of an adverse change
in the rules can boost the m easured cost of e q u ity for
Ja p anese se cu ritie s firm s relative to Japanese firm s in
g e n eral. In ad d ition , if inve sto rs a n ticip ate a d e clin e in
the p ro fita b ility of Jap ane se se cu ritie s firm s, then the
cu rre nt relation of th e ir e a rn in g s to the m arket valuation
of th e ir shares w ill tend to ove rsta te th e ir cost of equity
unless the stock m arket is ve ry m yopic. E vidence s u g
g ests that inve sto rs in the shares of the Big Four s e c u ri
ties firm s do fe a r lower p ro fita b ility going forward.
Japan ese se cu ritie s firm s resem ble U.S. secu ritie s
firm s in the m id-1970s in th e ir d e p e n d e n c e on e q u ity
co m m issio n s as a so u rce of revenue. U.S. s e c u ritie s
firm s drew about half of all revenues from e q u ity c o m
m issions w hen th e y were lib e ra liz e d in May 1975 (C h a rt
2). S ince then, the share of co m m is s io n s in in d u s try
revenues has fa lle n below a fifth . By co n tra s t, the large
Japanese s e c u ritie s firm s have d e p e n d e d and co n tin u e
to depend on e q u ity co m m is s io n s fo r a b o u t h a lf of th e ir
revenue (C h a rt 3).
Investors need o n ly e x tra p o la te a trend to foresee
th a t th e se revenues w ill sh rin k over the m edium term .
The Japanese a u th o ritie s have been re ducing e q u ity
co m m issio n rates g ra d u a lly (C h a rt 4). O ver the last
decade, co m m issio n rates fell at an an n u a l rate of 1
perce n t fo r tra d e s of 1 m illio n yen (a b o u t $7000), 1.6
perce n t for tra d e s of 10 m illio n yen ($7 0 ,0 0 0 ), 5.5 per
cent for tra d e s of 100 m illio n yen ($70 0 ,0 0 0 ), 13.4 per
cent fo r tra d e s of 1 b illio n yen ($7 m illion), and 18.9
perce n t fo r tra d e s of 10 b illio n ($70 m illion).
Table 5
Relation of U.S. Securities Firms’ Share Returns to Returns on the Standard and Poor’s 500 Index
M errill Lynch
Morgan Stanley
Salom on Brothers
Period
Beta
S tandard Error
R2
Beta
S tandard Error
R2
Beta
S tandard Error
R2
1986-91
1986
1987
1988
1989
1990
1991 (26 weeks)
1.30*
0.81
1.29
0.88
2.09*
1.45
1.67
0.085
0.19
0.15
0 15
0.23
0.24
0.34
45
.26
60
40
.61
.41
.50
1.14
1.17
1.30
0.87
0 88
1.06
1.37
0.088
0.21
0.17
0 20
0.25
0.22
0 36
38
.45
.53
.27
.19
.32
.38
1 44*
1.64*
1 55*
1.33
1.06
1.31
1.61
0.089
0.21
0.20
0.21
0.22
0.20
0.38
48
.56
.55
.45
.33
.47
.43
Source: S tandard and Poor’s
Note: Data are weekly.
'B e ta is significa ntly different from one on a tw o-tailed test at 5 percent significance
Table 6
Relation of Japanese Securities Firms’ Share Returns to Returns on the TOPIX Index
Daiwa
Nikko
Yamaichi
Nomura
Period
Beta
S tandard Error
R2
Beta
Standard Error
R2
Beta
S tandard Error
R2
Beta
S tandard Error
R2
1987-91
1987
1988
1989
1990
1991 (25 w eeks)
1.62*
2 07*
2.7 9’
2.03*
1.05
2.05*
0.09
0.21
0.27
0.27
0.13
0.28
.57
.68
69
.54
.58
71
1.34*
1.53*
2.44*
1.99*
.91
1.33
0.09
0.21
0.24
0.25
0.11
0.31
.50
.53
.66
.55
56
.45
1.42*
1.63*
1 99*
1.68*
1.10
1.95*
0.07
0.16
0.21
0.22
0.12
0.19
.62
.70
.63
.53
64
82
1.44*
1.79*
2.53*
1.89’
.98
1 39
0.09
0 22
0.23
0 26
0.11
0.24
.55
59
.71
.52
61
.60
Source: Daiwa and Dow Jones Tradeline International.
Note: Data are weekly.
’ Beta is significa ntly different from one on a two-tailed test at 5 percent significance.
FRBNY Q uarterly R eview/Autum n 1991
21
N ote th a t lib e ra liz a tio n of c o m m is s io n s h u rts the
s e c u ritie s firm s more than the lib e ra liza tio n of interest
rates ever h u rt Japa n e se banks. C om p e titio n am ong
th e b an ks fo r borrow ers kept the spread betw een aver
age d e p o sit rates and prim e lending rates fa irly narrow
by in te rn a tio n a l stand a rd s. R egulation of com m ission
rates proved m uch more e ffe ctive in protecting the reve
nues of the s e cu ritie s firm s.
R e in forcing the trend tow ard co m m issio n d eregulation
was a 1988 regulation th a t sh ra n k the Big F our’s share
of e q u ity b ro ke ra g e . The M in is try of F in a n ce was
re p orted to have a dvised s e c u ritie s firm s not to perform
m ore than 30 percen t of d a ily trading in any single
share. T h is g uid ance , aim ed at excesses a ssociated
w ith th e m a tic pro m otio ns of the Big Four, co n trib u te d to
a d e clin e in th e ir share of e q u ity brokerage from 60
p e rce n t in 1981 to 46 p e rce n t in the m iddle of the
d ecade to 33 percent at the end of the d e ca d e .20 As a
^ S a to sh i Takeuchi, "B ig Four's Transaction Share No Longer So Big;
30% Cap on Trade Volume H obbles Strategy to Promote Selected
Issues,” Japan E conom ic Journal, O ctober 28, 1989, p. 2. The
article notes that “ the gu id elines em erged after the U.S.
governm ent's special body on stock trading, the Brady Comm ission,
sharply c ritic iz e d the Big Four’s o lig o p o lis tic control [and] accused
the Big Four of m anipulating stock price s by con ducting con certe d
result of co m m issio n cu ts and lo st m a rke t share, Big
Four c o m m issio n s show ed little of the bu o ya n cy of the
trading value of J a p a n e se e q u itie s (C h a rt 5). N ote th a t
th e v a lu e o f s h a re tu r n o v e r on th e T o k y o S to c k
E xchange reflected not o n ly the p e rfo rm a n c e of share
prices but also the c le a r dow nw ard trend in share v o l
ume from the b e g inn in g of 1988.
F u rth e r a n a lysis of the Big Four co m m is s io n incom e
confirm s the e rosion of th e ir revenue base in the m id st
of the boom m a rke t of the late 1980s. We relate the log
of annual co m m issio n incom e fo r each of th e Big Four
for 1983-91 to a tim e trend and to the log of the value of
shares tra d e d on the Tokyo S to ck E xch a n g e . The e s ti
m ated co e ffic ie n t fo r tim e s u g g e sts th a t w hen the value
of tra d e s on the Tokyo S to ck E xch a n g e is held co n sta n t,
c o m m issio n revenue te n d s to d e clin e 4.7 p e rc e n t per
year. T his rate lies w ith in the sp e ctru m of ra tes of
d e clin e fo r regulated co m m is s io n s over the d e ca d e , as
o u tlin ed a b o ve — 1 p e rce n t to 18 p e rc e n t— and is close
to the rate of d e clin e fo r co m m is s io n s a s s o c ia te d w ith
Footnote 20 (continued)
buying operations based on sp e cific them es." See R eport o f the
P residential Task Force on M arket M echanism (W ashington, D C.:
Government Printing Office, 1988), p. I-8.
Chart 2
Composition of Revenues of the U.S. Securities
Industry
Commissions
q
I------1— I------1------1------1------1------ 1------ 1------ 1— I------ 1------1------ 1------1------1— 1
1975
77
79
81
83
85
87
89
91
1984
85
86
87
88
89
90
Source: Securities Industry Association.
Note: Data for 1991 cover first half of year only.
22 FRBNY Q uarterly Review/Autumn 1991
Sources: Daiwa, Nikko, Nomura, and Yamaichi annual reports.
91
an 80 m illio n yen trade. A llo w in g fo r 10 perce n t grow th
in tra d in g value and other, no n co m m issio n revenues,
in ve sto rs may readily foresee co m m issio n incom e d ro p
ping to less than a q u a rte r of the Big F our’s revenues
over the next fifte e n ye a rs.21
The Big F our’s co m m issio n incom e is q uite re sp o n
sive to the sto ck m a rk e t’s p e rform ance. O ur regression
a n a ly sis su g g e sts th a t a 10 p e rce n t rise in the value of
sto ck m arket tra d in g yie ld s a 7.1 pe rce n t increase in Big
Four co m m issio n revenues (Table 7). Big Four c o m m is
sions did not respond one-fo r-o n e to the value of tra d in g
because rising share p rice s tended to push tra n sa ctio n
values along the d e clin in g sch e d u le of co m m issio n s
and b ecause th e ir m a rket share was d e clin in g .
In vestors in the shares of Ja p a n e se se c u ritie s firm s
m ust pay a tte n tio n to the la rg e r agenda of dere g u la tio n
th a t in clu d e s a re co n sid e ra tio n of the A rtic le 65 b a rrie rs
betw een se cu ritie s and b anking businesses. A lready
21lf tra d in g value rises at 10 pe rcent per annum, if the elasticity of
com m issions with respect to tra ding value is .71, and if
deregulation continues to put a 4.7 percent per annum drag on
com m issions, then com m issions will grow at 2.1 pe rcent per
annum. If other revenues start off equal to com m issions and grow
at 10 pe rcent per annum, then it will take fifteen years for
com m issions to fall to a qu arte r of revenues. In other words,
(1.021 )x = (1.1 )x/3; solving for x, we have fifteen.
the J a p a n e s e c ity b a n ks have e q u ity s ta k e -o u ts in
s m a lle r s e c u ritie s firm s th a t co u ld be c a p ita liz e d upon
were A rtic le 65 repealed or m odified to p e rm it bank
e n try in to b ro k e rin g J a p a n e s e s h a re s . Even if the
change in the law now th o u g h t m ost lik e ly w ill not
perm it banks to b ro ke r shares, in ve sto rs n e ve rth e le s s
w ill have borne the risk th a t a m ore sw eeping d e re g u la
tion poses to s e c u ritie s firm s’ revenues and p ro fits .22
Finally, in ve sto rs may p e rce ive th a t th e e n try of fo r
eign se c u ritie s firm s may p re se n t a th re a t to the c o m
m ission revenue of the large J a p a n e se s e c u ritie s firm s.
Foreign firm s have b ro u g h t w e ll-d e v e lo p e d te c h n ic a l
tra d in g ta c tic s and more c ritic a l re search to th e ir b id
ding for in s titu tio n a l tra d e s . W ith th e s e a d vantages,
they have raised th e ir share of tra d in g on the Tokyo
S tock E xchange from 1.5 p e rc e n t in 1986 to 5.4 p e rce n t
in 1989 and 7.3 p e rce n t in the first h a lf of 1990.23
22“ While the entry of the banks into ce rta in areas of secu rities
business is now a foregone conclusion, the speed with w hich such
reforms will be im plem ented, the scope of the banks' new b u si
nesses, the form w hich en try will take, and the new questions
surrounding the banks’ ab ility to expand aggressively w hile
international capital ad equa cy requirem ents still seem a problem for
them, all com bine to sug gest a pictu re w hich is not as b la ck as
orig inally pe rce ive d ” (A licia Ogawa, "D aiw a S e cu ritie s ,” S. G.
W arburg Securities, M arch 26, 1990, p. 12).
23Business Week, July 9, 1990, p. 60. In N ational Treatment Study,
1990, p. 236, the U.S. Treasury cites the "m arket pow er” of the Big
Chart 4
Equity Commission Rates in Japan
Effective Dates
----------
April 1 ,1 9 7 7
----------
April 15, 1985
---------- November 25, 1986
—— -
October 5, 1987
June 4, 1990
1
10
100
1,000
10,000
Value of trade in millions of yen
Source:
Tokyo S tock E xchange Fa ct Book, various issues.
FRBNY Q uarterly Review/Autum n 1991
23
O ne m easure of the loss of fra n ch ise value of the
Ja p a n e se s e cu ritie s firm s is the ratio of m arket value to
b o o k v a lu e . T h e s e firm s ’ m a rk e t-to -b o o k ra tio has
de clin e d as com m issio n s have been reduced (C h art 6).
Note th a t the spate of p u b lic share o ffe rin g s in late 1985
and ea rly 1986 by U.S. s e c u ritie s firm s, in cluding B ear
S te arn s and M organ Stanley, were well tim ed by this
m easure.
The p o s s ib ilitie s of a d d itio n a l com m ission cuts, Ja p
anese bank co m p e titio n , and fu rth e r pen e tra tio n by
foreign firm s all represent risks that investors in Big
Four shares take into account. It is u n d e rsta n d a b le if
inve sto rs in the shares of the Big Four d isco u nt current
ea rn in g s som e w hat to allow for ch e a p e r sto ck trading
for Japa nese househ o ld s and in s titu tio n s .24 As a result
of the Big F ou r’s p ro b le m a tic grow th prospects, the
m easured cost of eq u ity for th e se firm s may be higher
than th a t of Jap anese firm s in general.
In d u s tria l o rg a n iza tio n
A n o th e r fa cto r je o p a rd iz in g the e a rnings of the J a p
anese se cu rity firm s is the perip h e ra l p o sition of the
J a p a n e s e s e c u rity firm s in th e c o u n try ’s in d u s tria l
org a n iza tio n . A Japa n e se c ity bank is at or near the
Footnote 23 (continued)
Four to acco unt for the minim al shares acco rded foreign firms in
un derw riting syndicates in Tokyo. In underw riting carve-outs of U.S.
firms, however, U .S.-based underw riters have played im portant
roles. See Ted Fikre, “ Equity Carve-Outs in Tokyo,” this Quarterly
Review, vol. 15 (W inter 1991), pp. 60-64.
24A major rating firm cited “ concerns about future profitability in light
of structural changes that are currently taking place in the
dom estic Japanese financial m a rket,” including “ lower dom estic
equity brokerage com m ission rates and ongoing discussions about
financial reform s,” in w arning investors of possible dow ngradings.
Standard and Poor's C redit Week, May 13, 1991, p. 19.
c e n te r of a k e ire ts u , a n e tw o rk of firm a ffilia tio n s tha t
a p p ro x im a te a c ro s s -s e c tio n of th e e co n o m y. T h is
arra n g em e n t a ffe cts the co st of e q u ity d ire c tly th ro u g h
the sto ck m arket: extensive cross sh a re h o ld in g w ith in
the keiretsu may s ta b iliz e and p e rh a p s even elevate
share prices. Indirectly, the keiretsu stru c tu re assures
ste a d ie r bu sin ess flow s and p ro vid e s im p lic it g u a ra n
tees of a ssista n ce to tro u b le d m em bers, b e n e fits th a t in
turn help to s ta b iliz e profit flow s.
The p e rip h e ra l p o sitio n of the Big Four s e c u ritie s
firm s is e vid e nt in the reference w ork In d u s tria l G ro u p
in g s in Ja p a n .25 Three of the Big Four a p p e a r o n ly once
each and no group a ffilia tio n is given. By co n tra s t, nine
of the eleven Ja p a n e se banks e xam ined in o u r s tu d y of
bank c o s t of c a p ita l a n c h o r w e ll-d e fin e d in d u s tria l
groups.
E ven th e e x c e p tio n a l J a p a n e s e s e c u r itie s firm
broadly co n fo rm s to the patte rn . N ikko S e c u ritie s is
listed as a sso cia te d w ith the M itsu b ish i group, but the
a ffilia tio n is d e scrib e d as w eak. The a g g re g a te e q u ity
stake in N ikko held by M its u b is h i g ro u p c o m p a n ie s,
m easured a g a inst the o ve ra ll c o n c e n tra tio n of s h a re
h o ldings in the s e c u ritie s firm , s u p p o rts th a t ch a ra c25Eighth ed. (Tokyo: Dodwell M arketing C onsultants, S eptem ber
1988), pp. 34-35, 49, 128, 304, 306, 506, 512.
Chart 6
Ratio of Market to Book Value for U.S. and Japanese
Securities Firms
Ratio
7 -------------------------------------------------------------------------------------------------Japanese firms
with equity holdings
at book value
/
«
/
\ ~
%
\
Table 7
Regression Analysis of Japanese Securities
Firms’ Com m ission income
September, 1984-91
Dependent variable
Natural Log of Commission
Independent variables
Time
Natural log of Tokyo Stock
Exchange trading value
Intercept
-.0 4 7
(.012)
.713
(0 5 0 )
9.31
(.146)
.87
R2
Degrees of freedom
0 I I I I 1i 1 1 I I I I I I 1I I I 1I M 1I I I 1I I I 1I I 1■ 1-L-J
1982
83
84
85
86
87
88
89
90
91
33
Note: Standard error of coefficients is given in parentheses.
24 FRBNY Q uarterly Review/Autum n 1991
Sources: Annual reports; Toyo Keizai Inc., Ja pa n C om pany
Handbook-, and Federal Reserve Bank of New York staff estimates.
te riza tio n . The M itsu b ish i g ro u p ’s aggregate holding of
N ik ko ’s e q u ity am o unts to no more than a third of the
top ten s h a re h o ld e rs’ co lle ctive stake. By co n tra st, the
M itsu bish i g ro u p ’s h o lding of M itsubishi B a n k ’s shares
bulks m uch large r: alm o st tw o -th ird s of the top ten
sh a re h o ld e rs’ s ta k e .26 O nly 13 of the 128 firm s in the
M itsu bish i group show lower group “ in flu e n c e ” ratios
than does N ikko S e cu ritie s. Moreover, N ikko has no
d ire cto rs from M itsubishi group com panies w hile M it
subishi Bank has two.
R eversing the p e rsp e ctive to exam ine fin a n cial firm s’
h o ld in g s o f e q u itie s c o n firm s th a t s e c u ritie s firm s
rem ain m uch less well co n n e cte d than Ja panese banks.
The se cu ritie s firm s chan n e le d p a rt of th e ir strong flow
of retained ea rn in g s during the boom years of the 1980s
into accu m u la tin g e q u ity stakes. As a result, se cu ritie s
firm s increased th e ir stra te g ic share of e xch a n g e -liste d
firm s fa ste r than banks did in the 1980s, e sp e cia lly if
“ m ost of the increase in bank e q u ity o w n e rs h ip ” was
“ not .. . fo r stab le share-o w n in g purposes [but rather]
fo r sh o rt-te rm investm en t p u rp o se s.” 27 In M arch 1990,
alm o st fo u r-fifth s of N o m u ra ’s e q u ity holdings by value
were held in the in ve stm e n t a ccount; such shares “ are
acquired for the C o m p a n y’s o p e ra tin g purposes and are
rarely sold under a C om pany policy.”28 S till, Japanese
b a n k s ’ s ta k e in firm s lis te d on th e T o kyo S to c k
E xchange rem ains ab out ten tim e s d e e p e r than th a t of
Ja p ane se s e cu ritie s firm s (Table 8).
At the firm level, exam in a tio n of the se c u ritie s firm s’
m ajor ho ld in g s in J a p a n ’s top com panies show s the
ho ldin gs to be few er and more c o n ce n tra te d than those
of the banks. A lth o u g h som e o b se rve rs contend that
“ N om ura is a ctive ly bu ild in g its own keiretsu of n o n in
d u s tria l co m p a n ie s in a va rie ty of sectors in clu d in g real
estate, in su ra n ce , d is trib u tio n , research, tra in in g , and
a d v e rtiz in g ,” 29 N om ura has not broken into the to p tie r
of o w n e rsh ip of firm s tra d e d in the firs t s e ctio n of the
Tokyo S tock E xchange. A se a rch of the to p e ig h t o r ten
sh a re h o ld e rs in each of the 1254 firm s listed on the
Tokyo S tock E x c h a n g e ’s firs t se ctio n fo u n d o n ly th irty three s h a re h o ld in g s of the Big Four s e c u ritie s firm s
(Table 9). N om ura a cco u n te d fo r half of the se , but its
h o ld in g s w ere q u ite c o n c e n tra te d in fin a n c ia l firm s ,
in cluding the shares of tw o of its own m a jo r s h a re
holders, Daiwa Bank and Toyo Trust. By co n tra st, the
oth e r 3 se c u ritie s firm s were not re p re se nte d am ong the
top sh a re h o ld e rs of any of th e ir own top sh a re h o ld e rs.
W hatever the d iffe re n ce s am ong th e m a jo r s e c u ritie s
firm s, none of them has h o ld in g s a p p ro a ch in g the near
c ro ss-se ctio n of co rp o ra te Japan ow ned by the city
banks.
The u se fu ln e ss of the lim ite d e q u ity sta ke s th a t the
Big Four do p o sse ss is s u g g e ste d by th e ir role as
un d e rw rite rs fo r 22 out of 24 of the firm s in w h ich th e y
hold m ajor sh a re h o ld in g s. In all but tw o ca se s fo r w hich
an u n d e rw rite r is liste d , the s e c u ritie s firm w ith the
e q u ity stake is at least co -le a d u n d e rw rite r, u su a lly
m ain u n d e rw rite r, and o fte n s o le u n d e rw rite r. T h is
strong pattern su g g e sts th a t e q u ity s ta ke s ce m e n t b u s i
ness relations and c o n s e q u e n tly u n d e rs c o re s the th re a t
to u n d e rw ritin g incom e a risin g from exp a n d e d pow ers
fo r banks.
C om bined w ith p ro sp e ctive d e re g u la tio n , the more
ce n tra l p o sition of banks in the stru c tu re of co rp o ra te
n etw orks renders the e a rn in g s of the s e c u ritie s firm s
insecure. If banks are allow ed to e n te r the w h o le sa le
se c u ritie s m a rkets, c o rp o ra tio n s m ay well favor th e ir
banks in the face of ro u g h ly co m p a ra b le p ricin g of
p rospective deals. For th is reason, u n d e rw ritin g reve
nues could be p a rtic u la rly at risk.
A co m p a riso n of the re sp o n se s to Y a m a ic h i’s d istre ss
26M itsubishi group com panies held 8.8 percent of Nikko Securities'
shares, w hile the top ten held 26.4 percent. M itsubishi group
com panies held 18.8 percent of M itsubishi Bank's shares, while the
top ten held 29.5 percent.
MRichard W. W right and G unter A. Pauli, The S econd Wave (New
York: St. Martin's Press, 1987), p. 71. M artin French, “ Japan's Great
Finance Plan," Aslamoney, July-A ugust 1991, p. 35, also suggests
that Nomura might establish itself at the cen ter of a m ajor industrial
group. The a rticle also associates Daiwa S ecurities with the
Sumitomo group and Yamaichi S ecurities with the Fuyo group.
27W. Carl Kester, Japanese Takeovers (Boston: Harvard Business
School Press, 1991), p. 207.
28Nomura Securities Company, Annual Report 1990, p. 23.
Table 8
Share of Tokyo Stock Market Owned by Japanese Securities Firms and Banks
.
S ecurities firms
Banks
; " :
. :a:
1982
1983
1984
1.6
1.7
1.7
17.5
18.0
17.7
; 4 ||
....
.
.
:,
: " .
. :
. >. , ...
:
■: ^
1985
1986
1987
1.8
17.4
1.9
2.1
2.3
2.3
2.0
18.4
19.3
19.8
21.3
21.3
1988
1989
1990
Source: Tokyo Stock Exchange.
FRBNY Q uarterly R eview/Autum n 1991
25
and the tro u b le s of a w e ll-co n n e cte d a u tom obile m aker
h ig h lig h ts the gre a te r risk a tte n d a n t on the se cu ritie s
firm s’ re lative ly perip h e ra l p o sitio n (although the sheer
size of Y am a ichi’s problem may have had so m e th in g to
do w ith the diffe re n ce in h a n d lin g the tw o cases). W hile
the a u to m o b ile firm M azda was helped through a period
of d istre ss by its main bank and affiliated c o m p a n ie s,30
Yam aichi had to resort d ire ctly to the governm ent.
Conclusions
U.S. s e cu ritie s firm s m ust c le a r a higher cost of equity
hu rd le in pricing th e ir p ro d u cts and s e rv ic e s than th e ir
“ R ichard Pascale and Thomas P. Rohlen, “ The Mazda Turnaround,”
Journal o f Japanese Studies, vol. 9 (Summer 1983), pp. 219-63.
Japanese c o u n te rp a rts . H ig h e r c a p ita l re q u ire m e n ts in
Japan may put U.S. firm s at a p a rtic u la r d is a d v a n ta g e
in com p e tin g there.
Factors c o n trib u tin g to low er co sts fo r J a p a n e se firm s
in th e 1 9 8 0 s w e re h ig h e r h o u s e h o ld s a v in g s and
sm o o th e r e co n o m ic g row th. In a d d itio n , a c o m p a riso n
of the exp e rie n ce of tro u b le d s e c u ritie s firm s in the
U nited S tates and Japan s u g g e sts a w id e r sa fe ty net in
Japan th a t m ay low er e q u ity costs.
Ja panese s e c u ritie s firm s seem to have a s m a lle r
cost of e q u ity a dvantage over th e ir U.S. c o u n te rp a rts
than J a p a n e se n o n fin a n cia l firm s and banks have over
th e ir respective c o u n te rp a rts . In p a rt, J a p a n e se in ve s
tors bear a risk of low er e a rn in g s fo r Ja p a n e s e s e c u ri
ties firm s in a d e re g u la te d e n v iro n m e n t, and th is risk
Table 9
Japanese Securities Firms’ Equity Stakes in Firms Listed on the Tokyo Stock Exchange First Section
U nde rw rite r Status
S ecurities Firm
S ector
Firm
Nomura
Financial
Daiwa Bank
Toyo Trust
Dai-Tokyo Fire & M arine
C hiba Bank
Osaka Securities Finance
Japan S ecurities Finance
Kokusai S ecurities
Sanyo S ecurities
N onfinancial
M anufacturing
Retail trade
C om m unications
C onstruction
Nikko
Transport
Fishing
Financial
M anufacturing
Daiwa
Yamaichi
M anufacturing
Retailing
C onstruction
Financial
Percent Stake
Hokko C hem ical
Nissho (m edical equipm ent)
Toyo Denki, Seizo
(railroad equipm ent)
Sogo (d epartm ent store)
N ippon Television Network
Nissan C onstruction
Daiwa Danchi
H itachi Transport
Hoko
Tokyo Securities
Toyo S ecurities
Maruman S ecurities
Kosei S ecurities
Japan Securities Finance
Tateho Chem ical
Ikegai (m achine tools)
Nissan Nohrin Kogyo
(plyw ood)
Kyodo Printing
N ippon Conveyor
Nihon Matai
(food containers)
Senshukai
Morimoto
N ippon Trust Bank
K ita-N ippon Bank
Taiheiyo S ecurities
3.1
6.9
9.2
1.7
17.0
3.4
32.5
8.1
Sole
Main
Co
Sub
Not
X
X
X
X
X
4.9
1.4
X
2.4
3.9
4.3
2.1
6.5
0.8
5.1
33.6
6.4
4.9
4.1
5.0
4.2
1.7
X
X
X
x
X
X
X
X
X
x
4.2
2.5
1.8
X
3.8
3.5
3.5
1.6
4.1
4.1
X
X
X
X
x
X
X
Source: Toyo Keizai Inc., Japan Com pany H a n d b o o k -F irs t Section, W inter 1990.
Notes: Nomura com prises Nomura S ecurities and Nomura Land and B uilding, and Nikko com prises N ikko S ecurities, N ikko B uilding , and
N ikko Investm ent Trust. No underw riters are listed for the securities firms in w hich the Big Four own stakes.
26 FRBNY Q uarterly Review/Autumn 1991
boosts their measured cost of equity. In addition, the
distance of Japanese securities firms from corporate
networks of mutual support may render their shares
more risky than the shares of firms secure within such
networks.
FRBNY Quarterly Review/Autumn 1991
27
Financial Liberalization and
Monetary Control in Japan
by Bruce Kasman and Anthony P. Rodrigues
The last fifteen years have witnessed a substantial
liberalization of Japan’s financial markets. Controls on
cross-border capital flows have been gradually dis
mantled and restrictions affecting competition and price
flexibility in domestic financial markets have been
relaxed. As a result, the range of free market assets has
grown significantly, as has the range of credit sources
available to domestic borrowers.
The experience of other industrial countries indicates
that changes in financial structure can have important
implications for the conduct of monetary policy. A num
ber of countries substantially revised their operating
procedures during the past decade as financial market
changes altered the relationships between policy tools
and objectives.
This article examines the effects of financial reforms
on Japanese monetary policy. In the first section of the
article we discuss how the Bank of Japan has altered its
operating strategy in response to the evolving financial
environment. We focus in particular on changes in the
intermediate objectives of monetary policy and in the
instruments used to implement policy. Our analysis sug
gests that the complex system of controls prevailing in
the m id -1970s supported an operating strategy
designed to influence the supply of bank credit. With the
relaxation of these controls, monetary policy authorities
shifted their strategy away from the control of credit
aggregates and, in recent years, have increasingly
em phasized interest rates as an agent of policy
transmission.
The a rticle ’s second section offers an empirical
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28 FRBNY Quarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
assessment of the monetary control mechanism in the
current liberalized environment. Specifically, we evalu
ate the degree to which the Bank of Japan has been
able to influence market interest rates and broad money
through interbank interest rates, its chief operating tar
get. We find that monetary policy changes have elicited
strong and consistent interest rate responses across
the term structure in recent years. In particular, long
term bond yields are much more responsive to mone
tary policy actions than in the past. In contrast, our
analysis of the relationship between money and interest
rates indicates that as financial reform has reduced
policy makers’ direct influence over banks, the link
between policy and broad money may have weakened.
Our results do not address the extent to which mone
tary policy actions have been transmitted to real activity
or prices. Nonetheless, our findings suggest that the
Bank of Japan has successfully adapted its operating
strategy to the changing financial environment.
Evolution of monetary control in Japan
In Japan, as in most countries, the ultimate goals of
monetary policy are output growth and inflation man
agement. The authorities typically tighten monetary pol
icy to reduce inflationary pressures and ease policy to
stimulate activity. Output and prices are controlled only
indirectly and with lags, however. Policy actions first
affect financial markets and only over time can be
expected to influence real activity and prices.
Because financial markets play a central role in trans
mitting monetary policy, policy makers generally base
th e ir op e ra tin g stra te g y on fin a n cia l v a ria b le s .1 In par
ticular, a fin a n cial varia ble su b je ct to a high degree of
co n tro l by a u th o ritie s u sua lly se rve s as the ta rg e t for
d a y-to-day o p e ra tio n s. Borrow ed reserves and the fed
funds rate are g e n e ra lly view ed as the cu rre n t o perating
ta rg e ts em ployed by the Federal R eserve in the U nited
S ta te s.2 In Japan, the reserve progress ratio, the ratio
of reserves accu m ula ted w ith in a m onthly m aintenance
period to tota l required reserves, and in te rb a n k interest
ra te s h a v e s e rv e d a s im ila r fu n c tio n s in c e th e
m id-1970s.
F in ancia l varia b le s are also em ployed as in te rm e d ia te
ta rg e ts or in d ica to rs. As the term “ in te rm e d ia te “ s u g
g ests, th e se varia b le s fit betw een the in stru m e n ts and
o p e ra tin g ta rg e ts of policy, on the one hand, and the
u ltim ate po licy goals, on the other. To be e ffective, an
in te rm e d ia te va riab le should provide inform ation about
p o licy goals and bear som e relation to op e ra tin g ta r
g ets. In the late 1970s and ea rly 1980s a num ber of
c e n tra l banks used a m o n e ta ry aggregate as a key
in te rm e d ia te variable, in m any cases se ttin g e xp licit
ta rg e ts fo r its annual grow th. In recent years, reliance
on m o n e ta ry aggregates as e xp licit in te rm e d ia te ta rg e ts
has d im in ish e d and a tte ntio n has sh ifte d to a w id e r set
of fin a n cial m arket variables.
1A more com plete d e scription of the role of targets and indicators in
the im plem entation of m onetary p o licy can be found in Richard G.
Davis, "In term ediate Targets and Indicators for M onetary Policy: An
Introduction to the Issues,” this Q uarterly Review, Summer 1990.
2Borrowed reserves are obtained by banks directly from the Federal
Reserve discou nt window. The federal funds rate is the rate
d e pository institutions cha rge one another to borrow reserves.
The m ovem ent away from m o n e ta ry ta rg e tin g has
la rg e ly stem m ed from c h a n g e s in the fin a n c ia l e n v iro n
m ent. The re m a in d e r of th is s e ctio n c o n s id e rs how
dere g u la tio n , g lo b a liz a tio n , and in n o vatio n in J a p a n ’s
fin a n c ia l m a rk e ts o ve r th e p a st tw o d e c a d e s have
s h a p e d th e B a n k of J a p a n ’s p o lic y a n d o p e ra tin g
strategy.
The m on e ta ry c o n tro l m e c h a n is m : m id-1970s
Until the m id-1970s, the J a p a n e se fin a n c ia l syste m was
highly regulated. A com plex system of c o n tro ls had
evolved, lim itin g in te re st rate m ovem ents and the a c tiv i
ties of m arket p a rtic ip a n ts .3 T h is syste m ensured that
la rg e p e rs o n a l s e c to r s u r p lu s e s w e re tr a n s fe r r e d
through banks to large c o rp o ra tio n s to prom ote high
rates of d o m e stic ca p ita l fo rm a tio n .4
In th is highly regulated e n v iro n m e n t, the B ank of
J a p a n ’s o p e ra tin g s tra te g y was d e s ig n e d to c o n tro l
bank cre d it to the n o n fin a n cia l private sector. The m o n
e ta ry policy co n tro l m e ch a n ism d u rin g the m id-1970s is
sum m arized in C h a rt 1. D a y-to -d a y p o licy o p e ra tio n s
3For a detailed discussion of the structure
Japanese financial system, see Robert A.
Financial M arkets: D eficits, Dilemm a, an d
MIT Press, 1986): and Yoshio Suzuki, ed.,
System (London: IFR Books, 1987).
and evolution of the
Feldman, Japanese
D eregulation (C am bridge:
The Japanese Financial
♦Generally, households were able to invest the ir savings only in bank
deposits, and banks had few alternatives to lending these funds to
corporations. N either the corpora te nor the banking sector had
significant direct recourse to raising funds in open cap ital markets,
which consequently remained undeveloped. Because interest rates
were adm inistratively controlled and often held below m arketclearing levels, major corpora tions cou ld borrow cheaply, while
sm aller firms and individ uals faced strin gent cre dit constraints.
Chart 1
Japan’s Monetary Control Mechanism: Mid-1970s
Instruments
Intermediate Objectives
Operating Targets
Policy Goals
Bank of Japan lending
Interbank market operations
Discount rate
I
Reserve requirements
Window guidance
Interest rate controls
FRBNY Q uarterly Review/Autum n 1991
29
too k the form of inte rb a n k m arket a ctivitie s or Bank of
Japan lending to banks. These o p e ra tio n s affected the
rate at w hich banks accu m u la te d reserves during a
m a in te n a n ce period as m easured by the reserve p ro g
ress ra tio .5 B ecause le nding by the Bank of Japan m ade
up a large co m p o n e n t of bank reserves and banks were
a lm ost e xclu sive ly lim ite d to the interbank m arket as an
a lte rn a tive source of fun d s, the response of in te rb a n k
in te re st rates to chan g e s in bank reserves was strong
and h ighly p re d icta b le .6
C h anges in in te rb a n k in te re st rates, in turn, in flu
enced the q u a n tity of cre d it provided by banks. A d m in
istra tive c o n tro ls on loan (and deposit) rate m ovem ents
lim ite d banks’ a b ility to pass on in te rb a n k rate changes
to th e ir c u s to m e rs .7 Thus, h ig h er in te rb a n k rates led
banks to ration credit and, given the heavy d e p e ndence
of the c o rp o ra te se cto r on bank lending, prom pted c u t
backs in e xpend itures. The Bank of Japan actively used
se ve ra l o th e r s u p p le m e n ta ry in s tru m e n ts , in c lu d in g
q u a n tita tive lending lim its on in d ivid u a l banks (w indow
guidance), d isco u nt rate ch a n g e s, and a d ju stm e nts in
reserve re quirem ents, to secure a desired level of bank
le ndin g, p a rtic u la rly in p e rio d s of tig h te n in g .
F in a n c ia l lib e ra liz a tio n : 1974-89
E con om ic grow th slow ed m a rke d ly a fte r 1973 and was
a cco m pa nied by a sharp d e clin e in the share of output
devoted to investm ent. Net c o rp o ra te borrow ing as a
share of GNP fell by m ore than half from its 7 percent
average share over 1965-74 (Table 1). At the sam e tim e,
the dem and fo r bo rrow ing by the p ublic se cto r more
than d o ubled du ring the 1970s, and J a p a n ’s te n d e n cy to
run p e rs is te n t cu rre n t acco u n t surp lu se s, in te rru p te d
o nly by oil price sho cks, becam e more pronounced.
T hese m a cro e co no m ic ch a n g e s d ra m a tica lly altered
the flow of fu n d s in the Ja panese econom y, creating
pressures that eroded the tig h t re strictio n s on fin a n cial
5lf banks fulfill their requirements along an average path, the reserve
progress ratio increases by 3.3 pe rcentage points each day. The
Bank of Japan adjusts aggregate reserves to determ ine this ratio
and transm it actions to the interbank market.
6Banks and securities corpora tions exchange funds in two interbank
markets: the call market, a short-term market analogous to the U.S.
federal funds m arket; and the bill discount market, where
com m ercial bills are rediscounted. Interest rates in the interbank
market are theoretically free from control. Nevertheless, because
money market brokers have until recently set interbank rates in
close consultation with the Bank of Japan, the Bank has had
con sid era ble short-term influence on interbank rates.
7H igher interbank interest rates were passed on to corporate
borrow ers in the form of higher deposit-to-loan ratios and of
increases in loan rates tied to the Bank of Japan's discount rate.
A lthough these rate movements allowed policy to affect expenditure
de cisio ns through financial price changes, they were less
significa nt than the effects of credit rationing.
http://fraser.stlouisfed.org/
30 FRBNY Q uarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
activity.8 In p a rticu la r, the large increase in g o v e rn m en t
borrow ing was pivotal in the d e ve lo p m e n t of active s e c
o n d a ry m a rke ts in se c u ritie s . D uring th e 1960s and
early 1970s, in itia l issues of g o ve rn m en t b onds were
bought by a syn d ica te of fin a n c ia l in s titu tio n s at p rice s
fixed by the Bank of Japan. T h e se in itia l fixed prices,
c o m b in e d w ith th e B a n k o f J a p a n ’s p r o m is e to
repurchase the bonds, s ig n ific a n tly lim ite d the d e v e lo p
m ent of s e c o n d a ry s e c u ritie s m a rk e ts .9 A fte r 1975,
however, the large scale of g o ve rn m en t bond issues
th reatened to u n d e rm in e m o n e ta ry co n tro l (b e ca u se of
the Bank of Japan pro m ise to re p u rch ase ) and force d
banks to raise the share of b onds in th e ir p o rtfo lio s at a
tim e w hen a ttra ctive a lte rn a tiv e in ve s tm e n t o p p o rtu n i
ties were becom ing a vailable. T h e se d e ve lo p m e n ts led
to a nu m b e r of reform s lib e ra liz in g bond issue rates,
rem oving re s tric tio n s on the sale of bonds in s e c o n d a ry
m arkets, and exp a n d in g the b onds’ m a tu rity ra n g e .10 By
8Good discussions of Japanese financial reform through the
m id-1980s can be found in the OECD E conom ic S urvey— Japan
(Paris: OECD, 1984); Suzuki, The Japanese F inancial System: and
Thomas F. C argill, "Japanese M onetary Policy, Flow of Funds, and
Domestic Financial Lib e ra liza tio n ," Federal Reserve Bank of San
Francisco E conom ic Review, Summer 1986, pp. 21-32. For analysis
of the more recent liberalization process, see K. O sugi, "Japan's
Experience of Financial D eregulation since 1984 in an International
P erspective," BIS E conom ic Papers, no. 26, January 1990; Masaaki
Nakao and Akinari Horii, "C hanges in the M onetary Control
Techniques and Procedures by the Bank of Jap an," Bank of Japan
Research and S tatistics D epartm ent, S pecial Paper no. 195, 1991;
and Kumiharu Shigehara, "Japan's E xperience with the Use of
Monetary Policy and the Process of Lib e ra liza tio n ," Bank of Japan
M onetary and E conom ic Studies, vol. 9, no. 1 (M arch 1991).
»The high com m issions prom ised to the synd ica te upon resale of the
bonds after the holding period raised the effective interest rate
earned by subscribers.
10For a more detailed discussio n of these issues, see Suzuki, The
Japanese Financial System, or M ichael Dotsey, "Japanese M onetary
Table 1
Net Lending by Sector
(As a Percentage of Nom inal GNP)
C orporate business
Personal sector
Public sector
Rest of world
Memo
Real GNP growth
1965-74
Average
1975-84
Average
1985-90
Average
-7 .1
9.4
-2 .6
-0 .7
-2 .9
10.3
- 7 .1
-0 .8
-4 .3
9.0
-1 .3
-2 .9
8.1
4 0
4.8
Sources: "Flow of Funds in Japan in 1990," Bank of Japan
Research and S tatistics D epartm ent, S pecial Paper no. 204,
July 1991; "Flow of Funds in Japan in 1989," Bank of Japan
Research and S tatistics D epartm ent, S pecial Paper no. 191,
A ugust 1990.
the early 1980s, turnover in Japan’s secondary govern
ment bond market had become the second largest in
the world.11
The increased supply of government bonds also
encouraged the development of short-term money mar
kets. In the late 1960s, the gensaki market, involving
repurchase transactions largely using government
bonds, arose as a vehicle for nonbank short-term
financing. Liquidity in this market was significantly
boosted by the growth in government bond issuance,
and by the mid-1970s, the gensaki market had become
a major unregulated short-term money market for nonfi
nancial corporations.
The growth of the gensaki market made it difficult for
the Bank of Japan to maintain deposit rate ceilings.
Attracted by rising market interest rates in the late
1970s, corporations were shifting their bank deposits to
gensaki assets. Pressure by banks led authorities, in
May 1979, to permit banks to issue certificates of
deposits (CDs).12
The emergence of freer domestic capital markets
coincided with the loosening restrictions on interna
tional capital transactions. Capital outflows were gradu
ally liberalized to contain upward pressure on the yen
after 1973 while capital inflows remained highly
restricted throughout the decade. However, after the
second oil price shock placed downward pressure on
the yen, a more general relaxation of controls was
implemented under the Foreign Exchange and Foreign
Trade Control Law in December 1980.13
The rise in Japan’s global surpluses in the first half of
the 1980s, particularly its bilateral surplus with the
United States, placed increased international pressure
on Japan to accelerate financial liberalization. In 1984,
F o o tn o te 10 ( c o n tin u e d )
Policy, A C om parative A na ly s is ,” B a n k o f J a p a n M o n e ta ry E c o n o m ic
S tu d ie s , vol. 4, no. 2 (1986).
" A c c o rd in g to the O E C D E c o n o m ic S u rv e y -J a p a n , turnover in the
J ap a n e se bond m arket reached 2 0 0 trillion yen in 1981, about oneq u arte r the size of the turnover in U .S. bond m arkets and almost
three tim es the turnover in the U.K. bond m arket.
12S eco n d a ry m arket trading in C D s did not begin until May 1982.
13The lifting of these capital controls resulted in large increases in
both inward and outward c ap ital flows. In addition, the lifting of
controls on nonresident transactions in Jap a n e se money m arkets
led to c onsid erably closer integration of Jap a n e se money m arkets
with those in Europe and the U nited States. As a num ber of studies
have shown, interest rates in Euroyen m arkets and in the dom estic
gensaki m arket be c a m e virtually e q u a lized by 1982. For exam ple,
see B ruce Kasm an and C harles Pigott, “Interest R ate D ivergences
am ong the M ajor Industrial N ations," this Q u a rte rly R eview , Autumn
1988, pp. 2 8 -4 4 ; and Jeffrey Frankel, “International Financial
Integration: Relations am ong Interest Rates, Exchange Rates, and
M on etary In d ic a to rs ,” in In te rn a tio n a l F in a n c ia l In te g ra tio n a n d the
C o n d u c t o f M o n e ta ry P o lic y , Federal Reserve Bank of New York,
1990.
a package based on the findings of a special committee
set up by the U.S. Treasury and Japanese Ministry of
Finance was announced. Most notably, new measures
reduced restrictions on Euroyen activities, including
Japanese resident borrowing and bond issues by Jap
anese and foreigners. In addition, limits on forward
foreign exchange transactions and swap limit rules on
Japanese banks were abolished. Subsequently, limits
on the purchase of foreign securities by Japanese non
bank institutional investors were lifted.
The second half of the 1980s saw continued efforts to
deregulate domestic markets. The liberalization of inter
est rates on bank time deposits began in 1985 and is
expected to be completed in 1993. Money market certifi
cate deposits were introduced in 1985; restrictions on
the minimum denomination, length of maturity, and
amounts issued have been steadily relaxed on these
accounts as well as on CDs and time deposits.14
The changing financial m arket environm ent
Financial liberalization and the associated process of
financial innovation have had far-reaching effects on
Japan’s financial system. Many constraints on portfolio
and expenditure choices have been removed, altering
the tightly controlled flow of funds patterns that sup
ported the m onetary control m echanism of the
mid-1970s. Three changes have been particularly sig
nificant in the evolution of the Bank of Japan’s operating
strategy: First, the importance of bank loans as a
source of funds has greatly declined. Second, the range
of instruments used by banks to raise funds has
expanded dramatically. Third, assets with market-deter
mined prices now predominate in the portfolios of all
sectors of the economy.
We have seen that bank credit was employed as an
intermediate target of policy in the mid-1970s largely
because of its central role in channeling funds between
lenders and borrowers. Before 1974, bank lending
accounted for close to three-quarters of intermediated
funds in Japanese markets (Table 2). In the second half
of the 1970s, however, the importance of bank lending
declined sharply as public sector bond issues increased
and corporate sector capital spending growth slowed.
Recent years have seen a further decline in the size
of domestic loans in Japan’s flow of funds. The interna
tionalization of Japan’s financial activities has combined
with the corporate sector’s steady move towards
securitization to reduce the share of domestic loans to
less than half of all intermediated funds flowing through
14A num ber of actions have also been taken to prom ote d e e p e n in g of
short-term m oney m arkets. A yen -d e n o m in ate d bankers’ a c c e p ta n c e
m arket was launched in June 1985 and a com m ercial p a p e r m arket
op en ed in 1987. In addition, a variety of short-term governm ent
bond issues have been introduced, and m easures have b een taken
to e xp and the maturity structure in the in terbank m arket.
FRBNY Quarterly Review/Autumn 1991
31
J a p a n .15
At the sam e tim e that d o m e stic cre d it d e clin e d in
im p o rta n ce, the Bank of J a p a n ’s control over bank le n d
ing d e c is io n s w e a k e n e d . T h e g ra d u a l re m oval of
re strictio n s on bank beh a vio r enabled banks to expand
th e ir fu n d in g sources (both at hom e and abroad) and to
ad ju st p rices of th e ir s e rv ic e s more independently. As a
result, banks’ re lia nce on Bank of Japan credit d e clined
significantly, along w ith the B a n k ’s leverage in using
w in dow g u id an ce or o th e r a d m in istra tive co n tro ls to
a ffe ct bank behavior.
T he d e v e lo p m e n t of E uroyen and CD m a rke ts in
recent years has been p a rtic u la rly im p o rta n t in this
p rocess (Table 3). Both m a rkets, free from official con-
trols, have expanded d ra m a tic a lly : E uroyen lia b ilitie s
have grow n more th a n fo u rfo ld and o u ts ta n d in g CDs
more than do u b le d since 1985. C urrently, th e y re p resent
nearly half of J a p a n e se m oney m a rke ts and exceed the
size of d o m e stic in te rb a n k m a rkets.
The in cre a se d a v a ila b ility of m a rk e t-p ric e d a s s e ts
exte n d s beyond th e fin a n c ia l sector. In v e s tm e n ts in
in stru m e n ts w ith m a rk e t-d e te rm in e d in te re s t rates by
the private n o n fin a n cia l se c to r have risen sig n ifica n tly,
p a rtic u la rly since 1984, w hen bank d e p o s it rates began
to be lib e ra liz e d (Table 4).
The rising share of m a rk e t-p ric e d in s tru m e n ts in p o rt
folios has u n d o u b te d ly in creased the im p o rta n c e of
in te re s t ra te s in e x p e n d itu re d e c is io n s . M o re o ve r,
p o te n tia l d is in te rm e d ia tio n betw een a d m in is te re d and
m a r k e t-p r ic e d a s s e ts h a s w e a k e n e d th e B a n k o f
J a p a n ’s a b ility to tra n s m it p o licy by a lte rin g spreads
betw een in te rb a n k rates and (a d m in iste re d ) loan and
15C orporate issues of securities, which accounted for roughly 10
percent of the funds raised by the corporate sector before 1973,
rose close to 15 percent over 1975-79, and in recent years have
risen to more than a third of corporate fund raising.
Table 2
Funds Interm ediation in Japan
(Fiscal Year Average)
Total funds s up plied (trillions of yen)
C om position (p ercen tage of total)
Funds raised by do m estic sectors
Loans from do m estic banks
Securities
Governm ent bonds
Foreign funds
Funds s up plied to overseas m arket
1965-74
1975-84
1985-90
20.4
58.6
122.91
92.1
70.2
19.2
12.9
2.7
7.9
89.1
54.6
32.0
26.6
2.5
10.9
72 3
47.5
19.7
7.2
5.1
27.7
Sources: "Flow of Funds in Japan in 1990," Bank of Japan Research and S tatistics D epartm ent, S pecial Paper no. 204, July 1991; "Flow of
Funds in Japan in 1989," Bank of Japan Research and S tatistics D epartm ent, S pecial Paper no. 191, A ugust 1990.
Table 3
Major Japanese Money Markets
(Trillions of Yen, End of Period Data)
Interbank market
D om estic interbank yen lia b ilities
Euroyen inte rban k lia b ilitie s
Open m arkets
Bond gensaki
CDs
C om m ercial paper
Total
Memo
D om estic interbank m arket as a share
of total money m arkets (p ercen tage points)
32 FRBNY Q uarterly Review/Autumn 1991
1975
1980
1985
1989
6.7
_
9.8
2.5
19 8
9.9
45.3
41.8
1.8
—
—
4.5
2.4
—
4.6
9.7
—
6.3
21.1
13.1
8.5
19.2
44.0
127.6
78.9
51.0
45.0
35.5
d e p o sit rates. C om paring in te rb a n k interest rates w ith
tw o ra te s s u b je c t to a d m in is tra tiv e c o n tr o l— tim e
d e p o sit and loan ra te s — du ring three episodes of m o n e
ta ry tig h te n in g p ro vides evid e nce of the reduced im por
tance of this chan nel (C h a rt 2). In both 1973-74 and
1979-80, w ide d iffe re n tia ls opened betw een o vernight
call rates and adm in iste re d loan and d eposit rates w hen
p o licy tig h te n e d . However, in 1990, the m ost recent
e p isode of tig h te n in g , spreads betw een th e se rates
rem ained roug hly unchanged.
R ecent s tru c tu re o f the m on e ta ry
c o n tro l m ech an ism
In response to these deve lop m e n ts, the Bank of Japan
has g ra d u a lly m oved away from a control m echanism
aim ed at re gulating the q u a n tity of bank credit. Instead,
it has in cre a sin g ly sou gh t to a ffe ct expenditure d e c i
s ions thro ugh o p e ra tio n s de sig ne d to affect m arket
interest rates. The current policy control mechanism, outlined
in Chart 3, shows a dramatic change from the mid-1970s.
On the level of po licy instru m e n ts, the s h ift away from
bank cre d it is reflected in the e lim in a tio n of co n tro ls that
d ire ctly affe cte d banks’ a b ilitie s to extend credit. In
p a rticular, w indow guidan ce , in the form of Bank of
Japan in s tru ctio n s to in d ivid u a l banks regarding lending
plans, was ended in 1982, and at about the sam e tim e,
the active use of reserve requirem ents as a p o licy tool
was d ro p p e d .16 As show n earlier, the use of in te re st rate
i«A more lim ited form of w indow guidance, in which the Bank of
Japan clarifie d its po licy orientation and discussed aggregate
co n tro ls as a m eans of ra tio n in g cre d it has also slo w ly
d im in is h e d , p a r t ic u la r ly fo llo w in g th e m a jo r p u s h
to w a rd s d e re g u la tin g b a n k lo a n a nd d e p o s it ra te s
begun in 1985.
The Bank of Japan has replaced th e se in s tru m e n ts
w ith a c tiv itie s o u ts id e th e in te rb a n k m a rk e t. It has
u n d e rta ke n o p e ra tio n s in s h o rt-te rm g o ve rn m e n t bills
(1981), CDs (1986), g e n sa ki (1987), and c o m m e rcia l
p aper (1989). A lth o u g h o p e ra tio n s o u ts id e the in te rb a n k
m arket have increased in fre q u e n cy in re ce n t years, the
Bank of Japan co n tin u e s to rely la rg e ly on its lending
p o licie s and o p e ra tio n s in in te rb a n k m a rke ts to a lte r
reserves.
A long w ith the reserve p rogress ratio, in te rb a n k in te r
est rates rem ain the p rim a ry o p e ra tin g ta rg e t of the
Bank of Japan. S ig n ific a n t ste p s have been ta ke n , how
ever, to link in te rb a n k and o th e r m oney m a rke ts m ore
closely, a d e ve lop m e n t th a t reflects the g re a te r im p o r
ta n ce placed on fin a n c ia l p rice s in the m o n e ta ry co n tro l
m echanism . In 1979, the B ank acted to a llo w in te rb a n k
rates to adjust more ra p id ly to open m a rke t c o n d itio n s,
and in s u b se q u e n t years, it co n tin u e d to reform its
procedures for in te rv e n in g in in te rb a n k m a rke ts. W hen
the Bank becam e co n ce rn e d th a t a ctio n s ta ke n to low er
in te rb a n k in te re st rates d u rin g 1987-88 were not being
tra n s m itte d to m oney m a rke ts, it im p le m e n te d a m a jo r
Footnote 16 (continued)
lending plans with individ ual banks, con tinued after 1982 and was
finally abolished in 1991.
Table 4
Financial Investments of the Domestic Nonfinancial Sector
Total investm ents (trillio n s of yen)
C om position (p ercen tage of total)
A ssets w ith m arket-determ ined
interest rates
Bank deposits*
Trust and insurance deposits
D om estic securities
Foreign credits
Assets w ith regulated interest rates*
1975-79 Average
1984
1988
1990
43.9
62.5
106.0
113.8
30.0
—
15.8
12.6
1.6
70.0
50.7
4.2
23.2
13.4
9.9
49.3
86.0
50.7
32.8
- 5 .1
7.6
14.0
147.1
94.7
25.7
16.9
9.8
- 4 7 .1
Memo
March 1984
Bank lia b ilities with m arket-determ ined
interest rates (share of total liabilities)
13.5
S eptem ber 1989
50.3
Sources: “ Flow of Funds in Japan in 1990," Bank of Japan Research and S tatistics D epartm ent, S pecial Paper no. 204, July 1991; “ Flow of
Funds in Japan in 1989,” Bank of Japan Research and S tatistics Departm ent, S pecial Paper no. 191, A ugust 1990.
in c lu d e s unregulated tim e de posits, certific a te s of deposit, and money market certificates.
in c lu d e s currency, dem and deposits, regulated tim e de posits, postal savings deposits, and trust fund bureau de posits.
FRBNY Q uarterly Review/Autum n 1991
33
set of in te rb a n k m arke t reform s in N ovem ber 1988.17
T he se reform s involved s h iftin g in te rb a n k o p e ra tio n s to
17For de tails on the evolution of Bank of Japan operations in
interbank m arkets, see Toshihiko Fukui, "Recent Developments of
the Short-Term Money Market in Ja p a n ,” Bank of Japan Research
and S tatistics D epartm ent, S pecial Paper no. 130, January 1986;
and Nakao and Horii, "C hanges in the M onetary Control
Te chniques.”
Chart 2
Interest Rate Movements during Periods of
Monetary Tightening
Percent
14
1973-74
Call rate
Loan rate
Time deposit rate
Percent
14
Percent
12
Source: Bank of Japan, E co n om ic Statistics M onthly.
34 FRBNY Q uarterly Review/Autumn 1991
s h o rte r m a tu ritie s and re p la cin g th e q u o ta tio n s yste m in
the in te rb a n k m arket by an o ffe r-b id syste m to p rom ote
g reater a rb itra g e betw een m arkets.
The ch ange in J a p a n ’s m o n e ta ry co n tro l m e th ods
over the past fifte e n years is m ost e v id e n t in the use of
fin a n cial va ria b le s in the in te rm e d ia te sta g e of the p o l
icy process. As e a rly as 1975, the B ank of Japan began
its s h ift away from bank cre d it as an in te rm e d ia te ta rg e t
and increased its e m p h a sis on the role of broad m oney
in its p olicy o p e ra tio n s. In a sense, cre d it and m oney
ta rg e ts had been e q u iv a le n t up u n til th is tim e b e cause
of th e ir close re la tio n sh ip on bank b a la n ce sh e e ts. But
w ith the la rg e -sca le flo ta tio n of g o ve rn m e n t bonds, s u b
s ta n tia lly u n d e rw ritte n by banks, th e c h a n n e ls of m oney
creation were no lo n g e r lim ite d to in cre a se s in le nd ing.
M oney thus becam e view ed a b e tte r in d ic a to r of levels
of aggregate e xp e n d itu re and assum ed a le a d in g role in
the m o n e ta ry co n tro l m e ch a n ism .
The Bank p ro b a b ly never a c tive ly e m ployed broad
m oney (M2 + C Ds) as an in te rm e d ia te ta rg e t, however.
Instead, broad m oney becam e the p rim a ry in d ic a to r
am ong a group of fin a n c ia l v a ria b le s th a t p ro vid e d in fo r
m ation on a c tiv ity and the s ta n ce of p o licy.18 Indeed, the
Bank of Japan re fra in e d from se ttin g e x p lic it ta rg e ts for
broad m oney and in ste a d chose to p u b lish q u a rte rly
forecasts fo r M2 + C D s from 1978 onw ard.
By the m id-1980s, fin a n c ia l lib e ra liz a tio n had begun
to blur the b o u n d a rie s of s p e cific fin a n c ia l a sse ts in
Japan. The w e a lth -h o ld in g p ro p e rtie s of b ank lia b ilitie s
in the form of CDs or d e re g u la te d tim e d e p o s its were
enhanced, w hile th e liq u id ity c h a ra c te ris tic s of s e c u ri
ties packaged in the form of tru s t and in s u ra n c e fund
accounts increased. T he rem oval of c o n tro ls on in te rn a
tio n a l c a p ita l m o v e m e n ts fu r th e r e d th e s e tre n d s
because in ve sto rs were able to tre a t a sse ts issue d in
Japan or in foreign m a rke ts m ore in te rch a n g e a b ly.
In recent years, the B ank of Japan has re sp o n de d to
these d e ve lop m e n ts by g ra d u a lly reducing its e m ph asis
on broad m oney in im p le m e n tin g policy. The d im in ish e d
im p o rta n ce of broad m oney was h ig h lig h te d in 1987
w hen M2 + CDs grew above B ank of Japan fo re c a sts for
three co n se cu tive q u a rte rs w ith o u t p rovoking a p o licy
response.
A lthough several v a ria b le s, in c lu d in g e xch a n g e rates
and asset prices, have been em ployed along w ith broad
m oney as key in te rm e d ia te v a ria b le s o ve r th e pa st
decade, m arket in te re st rates have be co m e in c re a sin g ly
18For studies sup portin g this view, see M ichael M. H utchinson,
"Japan's ’Money Focused’ M onetary Policy," Federal Reserve Bank
of San Francisco Econom ic Review, Summ er 1986, pp. 33-46;
Koichi H am ada and Fumio Hayashi, “ M onetary Policy in Postwar
Jap an,” in A lb e rt Ando, Hidekazu Eguchi, Roger Farmer, and
Yoshio Suzuki, eds., M onetary P olicy in Our Times (C am brid ge; MIT
Press, 1985); and Shigehara, “ Ja p a n ’s E xperience w ith Use of
M onetary Policy.”
va rie ty of fa cto rs, a p o s s ib ility th a t can u n d e rcu t the
a b ility of m o n e ta ry a u th o ritie s to in flu e n c e in te re s t rates
in a pre d icta b le way.
For s im ila r re a so n s, th e u s e fu ln e s s of m o n e ta ry
aggregates in the m o n e ta ry c o n tro l m e ch a n ism may
now be lim ited. The role of M2 + C D s as an in d ic a to r of
a c tiv ity was la rg e ly tied to re s tric tio n s th a t m ade it the
p rin cip a l m eans of liq u id ity in the J a p a n e se econom y. In
this e n viro n m e n t, M2 + CDs te n d e d to re fle ct a c tiv ity
fa irly c lo s e ly . A s th e w e a lth -h o ld in g p ro p e r tie s of
M2 + CDs and liq u id ity c h a ra c te ris tic s of o th e r fin a n c ial
assets have increased, all fin a n c ia l a g g re g a te s may
h ave b e c o m e le s s a c c u ra te in d ic a to r s o f a c tiv ity
because th e ir s h o rt-ru n b e h a vio r has be co m e se n sitive
to m ovem ents in relative rates of return. Moreover, as
the lines betw een fin a n c ia l a sse ts have becom e blurred,
p olicy ch a n g e s a ffe ctin g g e n e ra l e co n o m ic c o n d itio n s
or interest rates may have a w e a ke r link to any sp e cific
a ggregate, p a rtic u la rly in the s h o rt term .
In this se ctio n , we assess the B ank of J a p a n ’s a b ility
to tra n s m it p o licie s to fin a n c ia l m a rke t v a ria b le s in the
current lib e ra liz e d fin a n c ia l m a rke t e n v iro n m e n t. S p e c if
ically, we exam ine the re la tio n s h ip betw een m ovem ents
in in te rb a n k in te re st rates, the B a n k ’s m ain o p e ra tin g
ta rg e t, and the in te rm e d ia te va ria b le s view ed as key in
the m o n e ta ry co n tro l m e ch a n ism : m a rke t in te re s t rates
and broad m oney M2 + CDs.
O u r e a r lie r d is c u s s io n s u g g e s te d th a t by th e
m id-1980s fin a n cial lib e ra liz a tio n had caused s ig n ific a n t
changes in the fu n c tio n in g of Ja p a n e se fin a n c ia l m ar
kets. Thus we fo cu s o u r a tte n tio n on the post-1984
e x p e rie n c e to a s s e s s how c lo s e ly B a n k o f J a p a n
a ctions are being tra n s m itte d to in te re s t rates and m o n
e ta ry aggregates in a lib e ra liz e d e n v iro n m e n t. In a d d i
tion, this period is c o n tra s te d w ith 1974-84 to p rovide
in sig h t into c h a n g e s in th e se re la tio n s h ip s over tim e.
ce n tra l fo r p o licy o p e ra tio n s. M arket interest rates are
an in d ic a to r of e co nom ic c o n d itio n s and help to tra n sm it
in te rb a n k rate m ovem ents to real activity. C oncerns that
a ction s in in te rb a n k m arkets were not stro n g ly c o n
n ected to o th e r open m arket rates prom pted reform s in
the in te rb a n k m arket in 1988. Moreover, in 1989 and
1990, the Bank of Japan c o n s is te n tly cited the rising
level of m arket inte re st rates as a m otivation fo r tig h te n
ing m o n e ta ry policy.19
M onetary control of interest rates and broad
money: empirical evidence
We have seen that fin a n cia l lib e ra liz a tio n has led to
s ig n ifica n t ch ange s in the Bank of Ja p a n ’s op e ra tin g
strategy. M a rke t-d e te rm in e d fin a n c ia l prices, interest
rates in p articula r, play a more im p o rta n t role as both a
ta rg e t and an in d ica tor of policy. In co ntrast, attem pts to
c o n tro l bank credit or oth e r fin a n cial aggregates have
g ra d u a lly d im in ished.
W hile fin a n cial m arket changes have increased the
a tte n tio n given to interest rates in policy fo rm u la tio n,
they may also have m ade in te rp re ta tio n and control of
in terest rates more com plex. In the 1970s, m arket s e g
m e n ta tio n an d r e s tr ic tio n s on p o r tfo lio a c tiv itie s
ensured th at ce ntral bank actions w ould result in a
p re d icta b le pattern of su b s titu tio n betw een the in te r
bank m arket and the ge nsaki m oney m arket. In the
cu rre nt en viro n m e n t, ag en ts have a greater choice of
m oney m a rket in stru m e n ts (both do m e stic and foreign)
and can m ore ea sily move betw een these in stru m e n ts
and long -te rm se cu ritie s. T hese linkages may produce
a c lo se r co n n e ctio n am ong in te re st rates. N evertheless,
they allow in te re st rates to be influenced by a w ider
19See Nakao and Horii, "C hanges in the M onetary Control
Techniques," for a more de tailed discussion of the factors
determ ining m onetary p o licy changes during the 1980s.
Chart 3
Japan’s Monetary Control Mechanism: Late 1980s
Instruments
Operating Targets
Bank of Japan lending
Interbank market operations
Open market operations
— ►
Bank
reserves
Interbank
interest
rates
Intermediate Objectives
— ►
Market interest
rates
Broad money
-------►
Policy Goals
GNP
Inflation
Discount rate
FRBNY Q uarterly Review/Autum n 1991
35
E c o n o m e tric analysis
O ur e co n o m e tric a na lysis, e xplained in gre a te r de tail in
the ap p e n d ix, attem pts to in te g ra te sh o rt-te rm re la tio n
ships am ong Japan e se in te re st rates and m o n e ta ry
a ggregates w ith m odels governing th e ir lo n g er term
behavior. T his e m p irica l s tra te g y is m otivated by the
te n d e n cy of all the va ria b le s analyzed to d rift over tim e
w ith o u t co nverging tow ard a u nique long-term level.
A lth o u g h interest rates and m o n e ta ry aggregates may
d rift, eco n o m ic th e o ry su g g e sts that com m on u n d e rly
ing fa cto rs may d e te rm in e th e ir m ovem ent. The e xp e c
ta tio n s th e o r y of th e te rm s tru c tu re , fo r e x a m p le ,
su g g e sts tha t long-te rm in te re st rates ap p ro xim a te ly
equal an average of cu rre n t and expected future s h o rt
term rates. Thus, if s h o rt-te rm interest rate changes are
p e rsiste n t (because of p e rm a n e n t changes in in fla tio n
a ry e xp e cta tio n s or real in te re st rates), th e se changes
should be reflected across the term structure. S im ilarly,
com m on fa cto rs m ay explain the evolution of broad
m oney and the m o n e ta ry base. Interest rates may also
be an im p o rta n t p a rt of th is re la tio n sh ip if the so u rce s of
p e rs is te n t in te re s t rate c h a n g e s have a s y s te m a tic
effe ct on broad m oney in d e pe n d e n t of changes in the
m o n e ta ry base.
T he firs t p a rt of o u r a n a ly s is se a rc h e s fo r links
betw een Ja pane se in te re st rates or m o n e ta ry a g g re
g ates and in te rb a n k in te re st rates. The resulting “ cointe g ra tin g ” regressio ns d e s c rib in g these links capture
the lo n g -te rm response of va ria b le s to th o se p e rsiste n t
ch a n g e s in m o n e ta ry p o lic y in d ic a te d by s u s ta in e d
ch anges in the in te rb a n k call rate.
U nfortunately, th e re g re ssio n s do not p ro vid e in fo rm a
tion on th e se lin ka g e s at a h o rizon relevant to the
w o rkin g s of m o n e ta ry policy. T hus, th e se co n d p a rt of
the a n a lysis de ve lop s a m odel of the d y n a m ic response
of in te re st rates and m o n e ta ry a g g re g a te s to ch a n g e s in
th e c a ll ra te c o n s is te n t w ith th e s e c o in te g r a tin g
re lationships.
B efore tu rn in g to o u r s ta tis tic a l a n a ly s is , we present
in Table 5 som e d e s c rip tiv e e vid e n ce on in te re s t rates,
broad money, and e co n o m ic a ctiv ity .20 S p e cifica lly, the
ta b le show s m ean levels of m o n th ly in te re s t ra te s and
tw e lve -m o n th ch a n g e s in M2 + CDs, co n s u m e r price s,
and in d u s tria l p ro d u ctio n , along w ith s ta n d a rd d e v ia
tions fo r the p e rio d s 1975-84 and 1985-90.
S u b sta n tia l d e c lin e s in both th e level and the v a ri
a b ility of m oney m a rke t in te re s t ra te s are a p p a re n t
since 1984. O ver 1985-90 m oney m a rke t in te re s t rates
averaged roughly 5 to 51/2 p e rce n t, 200 b asis p oints
below th e ir 1975-84 levels. T h e ir v a ria b ility, as m e a
sured by the sta n d a rd d e v ia tio n , d e c lin e d by ro u g h ly
o n e -th ird.
2°The data in this section are drawn from various issues of Bank of
Japan, Econom ic S tatistics M onthly, and includ e the unconditio nal
call rate (month end), the bond re purch ase— or g e n sa ki— rate
(month end), the benchm ark governm ent bond rate (m onth end),
the bank certificate of de posit rate (80-179 days, month average),
the average rate on short-term bank loans (m onth end), the rate on
one-year time deposits (month end), the m onetary base (month
end, seasonally adjusted by the authors), and M2 + CDs (seasonally
adjusted, month end).
Table 5
Descriptive Statistics for Economic Activity and Interest Rates
(Period Averages)
1975-84
Mean
1985-90
Standard
Deviation
Mean
Standard
Deviation
Money market rates
Interbank call rates
G ensaki rates
C ertifica te of deposit rates1
7.1
7.2
7.6
2.2
1.9
17
5.1
5.2
5.6
1.5
1.3
1.3
R egulated rates
S hort-term loans
S hort-term time deposits
6.8
4.1
1.2
09
5.1
2.5
1.1
0.8
Long-term governm ent bonds
7.9
11
5.4
1.2
10.5
26
10.0
1.6
3.7
5.6
6.6
3.5
4.5
1.4
Growth in broad money (M2 + C D s )T
Econom ic activity
Ind ustria l production growth*
C onsum er price inflation*
■''Because c e rtific a te s of d e posit were not introduced until 1979, the values in the first two colum ns are averages for 1979-84.
♦Twelve-month percentage changes.
36 FRBNY Q uarterly Review/Autumn 1991
3.7
1.2
v : '
•
T h e se d e c lin e s are c o n s is te n t w ith o v e ra ll m a c
ro e conom ic d e velopm ents. Ja p a n e se c o n su m e r price
in flation averaged less than V/2 perce n t over 1985-90, a
drop of more than 4 perce n ta g e points from its 1975-84
average. Moreover, a sharp fall in the v a ria b ility of
c o n s u m e r p rice in fla tio n , b road m oney g ro w th , and
in d u stria l pro d u ctio n grow th since 1985 su g g e sts that
e co no m ic a c tiv ity has becom e co n sid e ra b ly more stable
in recent years.
Interest rates on oth e r fin a n cia l assets have also
d e clined but, in co n tra st to m oney m arket rates, exhibit
no s ig n ifica n t change in th e ir variability. From 1975 to
1984, loan and d e p o sit rates as well as long-term bond
yie ld s were c o n s id e ra b ly less va ria b le than m oney m ar
ket rates. The lower va ria b ility in these rates probably
reflected re strictio n s lim itin g th e ir responsiveness to
m arket c o n d ition s. A lthou g h fin a n cia l lib e ra liz a tio n has
u n d o u b te d ly allow ed these rates to a djust to changing
m arket co n d ition s, overall e co n o m ic co n d itio n s appear
to have becom e more stab le . As a result, the effe cts of
lib e ra liza tio n are seen not in the increased v a ria b ility of
th e se interest rates but rather in a co n ve rg e n ce in
in te re st rate va ria b ility th ro u g h o u t the econom y.
L o n g -te rm re la tio n s h ip s
The results of three te sts fo r com m on trends, or cointe g ra tin g re la tion s, betw een the o vernight call m oney
in te rest rate and various oth e r interest rates are pre
sented in Table 6 along w ith p a ra m e te r e stim a te s for
sp e cific e quations. T hese te sts d e te ct w h e th e r a single
u n derlying fa cto r e xp lains the d rift in the regression
va riab les (see appendix).
O ve ra ll, th is e vid e n c e in d ic a te s th a t the lin ka g e s
b e tw ee n th e B ank of J a p a n ’s o p e ra tin g ta rg e t and
m oney m arket interest rates have been q uite strong
th ro u g h o u t the 1974-91 period. The call rate appears to
have been co in te g ra te d w ith both the g ensaki rate and
the CD rate d uring 1974-84 and 1985-91. D uring the
e a rlie r period, a 100 basis p oint increase in the call rate
was associate d w ith a ro ughly equal change in the
g ensaki rate. In the later period, the response of the
g e nsaki rate was som ew hat sm aller, e stim ated at 82
basis points.
In co n tra st, n e ith e r loan rates nor long-term g o ve rn
m ent bond yields were co in te g ra te d w ith the call rate
betw een 1974 and 1984. T his result is co n s is te n t w ith
our e a rlie r c o n te n tio n tha t a d m in istra tive co n tro ls on
loan rates and re strictio n s on the d evelopm ent of a
se c o n d a ry m arket in long-te rm governm ent bonds may
have p a rtia lly isolated th e se m arkets from in te rb a n k
and s h o rt-te rm m oney m arkets.
F in a n c ia l lib e ra liz a tio n d oes a p p e a r to have in te
grated lo ng-term bond m arkets and m oney m arkets.
B etw een 1974 and 1984 the call rate did not have a
c o n s is te n t lo n g -ru n c o n n e c tio n to g o v e rn m e n t bond
yields. A fte r 1984, however, stro n g e v id e n c e of a cointe g ra tin g relation betw een lo n g -te rm bond y ie ld s and
call rates em e rg e s: lo n g -te rm bonds in cre a se by 69
basis p oints in response to a 100 basis p o in t rise in the
call rate.
There is little e vid e nce , however, th a t fin a n c ia l reform
has tig h tly in te g ra te d m oney m a rke ts w ith loan m a rkets.
A lthough the loan rate reacted m ore s tro n g ly to call rate
changes a fte r 1984, bank loan rates were not c o in te
grated w ith the call rate b etw een 1985 and 1991, s u g
gesting th a t there has not been a c o n s is te n t lo n g -te rm
relation betw een the rates.
Table 6
Cointegration Relationships for Monthly
Interest Rates
Response to Call Rate of
Gensaki
R2
C ointegration tests
ADF
SW
PP
CD*
R2
C ointegration tests
ADF
SW
PP
Long-term bond
R2
C ointegration tests
ADF
SW
PP
Loan rate
R2
C ointegration tests
ADF
SW
PP
1974-84
1985-91*
1.00
.84
.82
.97
- 3.28*
- 5 1 .6 7 * “
-3 2 .3 7 ’ “
.90
.97
-2 .9 7
- 2 9 .7 5 * ”
-2 5 .2 2 * *
.34
.65
-1 .7 8
- 1 0 .8 2
- 1 2 .4 2
-2 .7 1
-3 3 .9 0 * * *
- 4 4 .2 0 * * *
.85
.96
- 2 .4 2
-3 5 .2 6 * * *
-4 3 .3 3 “ *
.69
.83
- 3 .3 3 *
-2 3 .7 7 “ *
-2 0 .7 9 “
.49
.75
.73
.88
- 2 .4 0
-7 .8 7
- 5 .9 8
- 1 .3 3
- 9 .6 2
- 1 1 .1 4
Notes: ADF is the augm ented D ickey-Fuller sta tistic (using
seven lags). PP is the Phillips-Perron Z„ sta tistic (using seven
autovariance lags). SW is the Stock-W atson sta tistic (using
seven lags). C ritical values for the D ickey-Fuller and P hillipsPerron statistics were o b tained from P.C.B. P hillips and Sam
O uiliaris, "A sym ptotic P roperties of R esidual-B ased Tests for
C ointe gra tion,” E co n o m e trica , vol. 58, no. 1 (January 1990),
pp. 165-91 C ritical values for the Stock and Watson statistic
are from James Stock and Mark Watson, “ Testing for Common
Trends," Journal of the A m erican S tatistical A ssociation, vol.
83 (D ecem ber 1988), pp. 1097-1107.
*Sample covers January 1985-May 1991,
^Earlier sam ple covers 1980-84.
‘ S ignificant at 10 pe rcent level.
‘ ‘ S ignificant at 5 pe rcent level.
“ ‘ S ignificant at 1 pe rcent level.
FRBNY Q uarterly Review/Autum n 1991
37
O ur m odel c o n n e c tin g m o n e ta ry p o licy actions to
broad m oney is based on a standard view of the money
su p p ly p ro c e s s .21 Bank of Japan o p e ra tio n s in in te rb a n k
m arke ts are a ccom p an ie d by changes in both reserves
available to the banking system and in te rb a n k interest
rates. G iven un changed asset a llo ca tio n s by banks and
d e p o sito rs, ch ange s in reserves can be expected to
a lte r broad m oney (M2 + C Ds) in a p redictable manner.
Interest rate m ovem ents can a lte r p o rtfo lio choices,
th u s in flu e ncin g the m oney supply in d e p e n d e n tly of
ch ange s in the m o n e ta ry base. H igher m arket interest
rates, all else equal, raise the cost of holding bank
reserves and co n se q u e n tly may increase the m oney
m u ltip lie r (the ratio of broad m oney to the m o n eta ry
base). At the sam e tim e, an increase in ce n tra l bank
lending rates or in te rb a n k rates relative to m arket rates
could increase dem and fo r reserves and thus low er the
21Money dem and is also im portant in money stock determ ination.
Because we do not e x p licitly m odel money demand, our analysis
should not be viewed as a full behavioral model for the
determ ination of interest rates and the money stock.
Table 7
Cointegrating Money Models
Response of M2 + CDs to:
In (base)
Trend
R2
C ointegration tests
ADF
SW
PP
1974-84
1985-91f
.554
.005
.993
-2 .0 9 1
-8 .1 7 4
-6 .2 1 2
.412
.005
.998
-3 .1 3 8
-3 6 .4 6 9 ** *
-4 5 .2 2 7 ***
Response of M2 + CDs to
In (base)
Gensaki rate
Call rate
Trend
R2
Cointegration tests
ADF
SW
PP
1.000*
.009
-.0 1 9
.002
.997
1.000*
.007
-.0 0 4
.000
.993
-2 .3 4 2
-3 9 .7 0 4 ***
-4 9 .0 8 1 ***
-2 .3 2 4
-4 5 .7 3 8 * ’ *
-5 6 .6 5 0 ***
Notes: R2 is the square of the correlation coefficient between
actual and predicted In (M2) for the two regressions with base
coe fficients equal to one. ADF is the augm ented Dickey-Fuller
sta tis tic s (using seven lags). PP is the Phiilips-Perron Z„
statistic (using seven autovariance lags). SW is the StockWatson s ta tistic (using seven lags). C ritical values for the
D ickey-Fuller and P hiilips-Perron statistics were ob tained from
P hillips and O uiliaris, “A sym ptotic Properties of ResidualBased Tests." C ritical values for the Stock and Watson statistic
are from Stock and Watson, “ Testing for Common Trends.”
tS am ple covers January 1985-May 1991.
♦C onstrained to equal 1.
’ S ignificant at 10 pe rcent level.
’ •S ign ificant at 5 pe rcent level.
“ ‘ S ignifican t at 1 pe rcent level.
38 FRBNY Q uarterly Review/Autumn 1991
m oney m ultiplier. We su g g e ste d e a rlie r th a t B ank of
Japan a ctio n s a sso cia te d w ith h ig h e r in te rb a n k in te re st
ra te s , in c lu d in g w in d o w g u id a n c e a n d c h a n g e s in
reserve req u ire m e n ts, m ay have re in fo rce d a d e c lin e in
the m oney m u ltip lie r in the p a s t .22
N on e th e le ss, in te re st rate e ffe cts on the m oney m u lti
p lie r m ight be tra n sito ry, p a rtic u la rly b e ca u se re g u la to ry
re stra in t on bank b e h a vio r was a p p lie d o n ly te m p o ra rily.
T h u s , we fir s t m o d e l th e lo n g e r te rm b e h a v io r of
M2 + CDs as a fu n c tio n of the m o n e ta ry base alon e,
in clu d in g a tim e trend term to a llow fo r te c h n o lo g ic a l
fa cto rs th a t may have a ltered the m oney m u ltip lie r over
tim e.
The co in te g ra tin g re g re ssio n s fo r th is m odel, p re
sented in the u pper ha lf of Table 7, p ro vid e no e vid e nce
of a c o in te g ra tin g re la tio n b e tw e e n th e b a s e a nd
M2 + CDs before 1985. T h is result su g g e s ts th a t fa cto rs
le ading to p e rs iste n t m ovem ents in th e m oney m u ltip lie r
were an im p o rta n t d e te rm in a n t of th e lo n g -te rm b e h a v
ior of b road m o n e y d u rin g th is p e rio d . In c o n tra s t,
betw een 1985 and 1991, e vid e n ce of a c o in te g ra tin g
regression is present, and th u s m oney base ch a nges,
th rough a sta b le m u ltip lie r, a d e q u a te ly e xp la in th e lo n g
term evolution of M2 + CDs.
To assess w h e th e r the p e rs is te n t m ove m e n t in the
m oney m u ltip lie r before 1985 was a sso cia te d w ith in te r
est rate m ovem ents, we add the call and g e n sa ki rate to
our regression m odel. In th is c o n te xt, the g e n sa ki rate
captures a lte rn a tive bank in ve stm e n t o p p o rtu n itie s th at
becam e available b e g inn in g in the se co n d h a lf of the
1970s. C a ll ra te m o v e m e n ts m e a s u re th e c o s t of
re se rve s h o rtfa lls to b a n ks and a lso p ro x y fo r the
effe cts of policy a ctio n s not related to ch a n g e s in the
m o n e ta ry base. The co e ffic ie n t on the m o n e ta ry base is
re stricte d to equal one in th is fra m e w o rk b e ca u se, by
d e fin ition , base c h a n g e s are fu lly reflected in th e m oney
su p p ly w hen the m u ltip lie r is u n ch a n g e d .
Including in te re st rates in th e m odel y ie ld s strong
evidence of c o in te g ra tio n fo r the 1974-84 p e rio d . M o re
over, the p a ra m e te r e stim a te s are of th e c o rre c t sign
and su g g e st a large e ffe c t of call rate ch a n g e s on broad
money. S pecifically, w hen m a rke t ra te s and th e m o n e
ta ry base are held c o n sta n t, a 100 basis p o in t increase
in the call rate is a sso cia te d w ith a p e rm a n e n t d e clin e
of nearly 2 p e rce n ta g e p o in ts in M2 + C Ds.
N evertheless, th e s e call rate e ffe c ts d e c lin e s u b s ta n
tia lly after 1984. Indeed, th e sm all size of the in te re st
22ln particular, Japanese banks were forced to constrain lending
activities when po licy tigh tene d because loan rates were regulated
and the Bank of Japan im posed qu antita tive restrictions on lending.
These restrictions, tog ethe r with the B ank’s active use of reserve
requirement changes to im plem ent m onetary policy, forced banks to
increase their reserve-deposit ratios as po licy tigh tene d, an o u t
com e that lowered the money m u ltip lier and broad money for a
given m onetary base.
rate c o e ffic ie n ts in the m oney supply e q u a tio n s su g
g e sts th a t in te re st rate m ovem ents, in d e pe n d e n t of the
m o n e ta ry base, may no lo n g er have any lasting effect
on broad money.
D yn a m ic re la tio n s
The evid e nce of long-term lin ka g e s betw een call rates
and o th e r fin a n cial va ria b le s does not, by itself, cla rify
how m o n e ta ry p olicy cha n g e s are tra n sm itte d over a
h o rizo n re le va n t fo r p o lic y m a ke rs. To a d d re ss th is
q u e stio n , we e stim ate a d yn a m ic erro r-co rre ctio n m odel
fo r Jap a n e se fin a n cial m a rke t variables. The m odel,
p resented in detail in the a p p e n d ix, captures in a fa irly
u n re stricte d way the o bse rve d tim e series re la tio n sh ip s
am ong th e se va ria b le s by im posing the co n d ition that
the d yn a m ic b eh avior co n ve rg e to the lo n g -te rm cointe g ra tin g regressio ns estim a te d above.
A ssum ing th a t the Bank of Japan has c o n sid era b le
control over the in te rb a n k call rate, we use th e m odel to
assess the m o n e ta ry co n tro l m e ch a n ism by co m p a rin g
the responses of in te re s t rates and broad m oney to an
in itia l 100 basis p o in t increase in the call ra te .23 T hese
responses, presented in C h a rts 4 and 5, can be in te r
preted as the average re sp o n se of in te re s t ra te s and
broad m oney to p o licy ch a n g e s over the sa m p le . The
s im u la tio n s also tra ck s u b s e q u e n t ca ll rate m ovem ents
g e n erated by the m odel; th e se m o ve m e n ts ca p tu re the
te n d e n cy of Bank of Japan p o lic y s h ifts to o c c u r g ra d u
ally as well as the ty p ic a l re sp o n se of call rates to
c hanges in o th e r fin a n c ia l v a ria b le s.
The evidence in C h a rt 4 su g g e s ts s o m e w h a t s tro n g e r
tra n sm issio n of call rate sh o cks to in te re s t ra te s a fter
^A lth o u g h the results of this analysis are presented sep arately for
interest rates and m onetary aggregates, they are derived from a
single model that accounts for the interrelationship am ong these
variables.
Chart 4
Interest Rate Responses to a Call Rate Increase
Percentage point change
2.0 ---------------------------------------------------------------------------------------1974-84
Percentage point change
2.0-----------------------------
___________
1985-91
1 .5 -
1974-84
1 . 0 0 \ --------------Ratio of Gensaki to
\
call rate changes
0 .7 5 -
Months after shock
Note: Chart shows the predicted response to an increase of 100 basis points in the call rate.
FRBNY Q uarterly Review/Autum n 1991
39
1984, p rim a rily because of the increased re sp o n sive
ness of lon g -te rm bond yie ld s. From 1975 to 1984 only
the gen sa ki rate responded stro n g ly and im m e d ia te ly: a
100 basis point increase in the call rate prom pted an
im m e d ia te and equal rise in the gensaki rate. In c o n
tra st, th e im m ed iate responses of long-term bond yields
and loan rates were q u ite sm all. Three m onths fo llow ing
th e s h o c k o n ly a b o u t o n e -th ird of th e c u m u la tiv e
increase in call rates had been passed th rough to loan
rates, and less than o n e -fo u rth of this increase was
tra n sm itte d to bond yie ld s. Over tim e, loan rates c o n
tin u e d to rise, in p a rt re fle ctin g a d m in istra tive de cisio n s
by the Bank of Japan. But even after a year only about
o n e -th ird of the call rate increase was reflected in lo n g
term bond yield s.
D uring 1985-91, the g e nsaki rate responded more
slo w ly to the call rate shock, rising a bout 40 basis
p o in ts at the tim e of the shock. N onetheless, three
m onths a fte r call rates increased, th e re sp o n se of the
gensaki rate was clo se , in p ro p o rtio n a l te rm s, to both its
e stim ated lo n g -te rm response and th a t o b se rve d in the
e a rlie r p e rio d . L o n g -te rm bond y ie ld s , in c o n tra s t,
reacted much more s tro n g ly to call rate s h o cks a fte r
1984. Three m onths a fte r the in itia l sh o ck, n e a rly 60
pe rce n t of the call rate sh o ck was passed th ro u g h to
bond yie ld s, a response n e a rly three tim e s as great as
d u rin g th e e a rlie r p e rio d . T h e lo a n ra te re s p o n s e
show ed little change across th e tw o pe rio d s.
S im u la tio n results fo r broad m oney and th e m oney
m u ltip lie r in C h a rt 5 s u p p o rt the vie w th a t th e m oney
m u ltip lie r was an im p o rta n t p a rt of the p o lic y tra n s m is
sion m echanism before 1985. In the 1974-84 p eriod,
increases in the call rate caused a sh a rp fall in the
m oney m ultiplier. In the firs t th re e m onths fo llo w in g a
call rate increase, the m u ltip lie r d e c lin e d by n e a rly 2
percent, in d ica tin g la rg e p o rtfo lio s h ifts by banks.
Chart 5
Response of Broad Money and Money Multiplier to a Call Rate Increase
Percentage point change
3 ------------------------------------1985-91
Percentage point change
3
1974-84
Call rate
Call rate
-•----- #-
Broad money
Money multiplier
.1 _ _ _ _ _
Broad money
Money multiplier
-2
-2 ------
J_ _ _ L
J_ _ _ L
J_ _ _ L
-3
Proportional change
0.5
1974-84
-3 I------Proportional change
0.5 ---------------------------
Ratio of broad money to
call rate changes
■
Ratio of multiplier to
call rate changes
----- *----- •-1.5
J_ _ _ I_ _ _ I_ _ _ I_ _ _ L
2
3
4
5
J_ _ _ I
6
7
8
9
10
11
12
-
J _ _ _ I_ _ _ L
2.0
1
2
Months after shock
Note: Chart shows the predicted response to an increase of 100 basis points in the call rate.
40 FRBNY Q uarterly Review/Autumn 1991
3
4
5
6
7
8
Months after shock
9
10
11
12
The im m e d ia te e ffe ct of a call rate increase on broad
m oney was quite sm a ll, but over tim e, broad money
s te a d ily de clin e d . A t three m onths, broad m oney fell by
about .4 p erce nt; a year a fte r the shock, broad money
d e clin ed by more than 1 percent.
D uring 1985-91, the response of broad m oney to a
call rate increase has been roughly s im ila r to th a t seen
in the e a rlie r period. N o ne th e le ss, the channel of tra n s
m ission a p p ears to have ch anged dram atically. C all rate
m ovem ents no lo n g e r a lte r the m oney m ultiplier, w hich
rem ains roughly unchanged over the fo re ca st horizon.
Thus, m o n e ta ry p o licy influ e nce s broad m oney through
its e ffe ct on the supply of reserves rather than through
its influence on bank asset allo ca tio n .
P re d ic ta b ility o f d yn a m ic responses
To assess m o n e ta ry control, we m ust co n s id e r not only
the size of the response of fin a n cia l m arket variables to
p o licy but also the p re d ic ta b ility of these responses.
E vidence on p re d ic ta b ility of responses can be obtained
by com puting the fo recast standard errors of our va ri
ables at d iffe rent horizon s. T hese standard errors, pre
sented in Table 8 , ind ica te the degree to w hich the
actual responses of fin ancial variables to in te rb a n k rate
m o v e m e n ts are lik e ly to fa ll n e a r th e e s tim a te d
responses presented in C h a rts 4 and 5 .24 In co m p u tin g
th e se sta n d a rd e rro rs, we have exclu d ed th e u n ce r
ta in ty a ttrib u ta b le to flu c tu a tio n s in call rates. In our
fram ew ork, call rate m ovem ents re p re se nt m o n e ta ry
p o licy actions, and th e se e stim a te s sh o u ld ca p tu re the
u n c e rta in ty in fin a n c ia l m a rke t v a ria b le s a risin g from
fa cto rs oth e r than m o n e ta ry p o lic y .25
The estim a te d fo re ca st va ria n ce s fo r in te re s t rates
present a m ixed picture. S ta n d a rd e rro rs fo r the g e nsaki
rates d e clin e d s u b s ta n tia lly a fte r 1984, p o s s ib ly re fle ct
ing the e m e rgence of m ore sta b le e co n o m ic c o n d itio n s.
In c o n tra s t, the s ta n d a rd e rro rs fo r lo n g -te rm bond
y ie ld s and lo a n in te re s t ra te s s h o w e d little o r no
24Note that these forecast standard errors only represent the
uncertainty arising from u n pred ictable shocks affecting the system.
In com puting the standard errors, we ignore un ce rta in ty due to
im precision in our coe fficient estim ates. Thus, the forecast standard
errors in Table 8 probab ly underestim ate the total un certa in ty
surrounding these responses
25Despite this adjustm ent, a com parison of forecast variances for
1974-84 and 1985-91 is likely to be influenced by cha nging policy
objectives as well as changes in the general degree of econom ic
stability.
Table 8
Predictability of Response to Cali Rate Increases
Forecast S tandard Error*
(P ercentage Points)
Months after Shock
1974-84
Gensaki rate
3
6
12
.68
.92
1.39
.38
.55
.81
Long-term bond rate
3
6
12
.57
.72
.98
.63
.72
.92
Loan rate
3
6
12
.21
.37
.65
.20
.41
.80
M 2 + CDs
3
6
12
1.45
2.56
4.97
1.61
2.79
5.53
M onetary base
3
6
12
2.11
2.98
5.18
2.16
3.06
5.55
Money m u ltiplier
3
6
12
1.74
1.91
2.18
1.65
1.67
1.71
1985-91
Notes: The forecast standard errors are derived for three-, six-, and tw elve-period-ahead forecasts con ditiona l on a predeterm ined call rate
path. Our ca lcu la tio n uses the un conditional forecast standard error and sub tracts the portion due to call rate shocks.
tE xclu d e s fra ction of variance a ttributa ble to call rate changes.
FRBNY Q uarterly Review/Autum n 1991
41
s y s te m a tic c h a n g e , w ith th e p o s s ib le e xc e p tio n of
increased u n c e rta in ty a tte n d in g loan rate responses at
lo n g er h orizon s.
The sm all pre-1985 s ta n d a rd errors fo r bond yields
and loan rates reflect the re strictio n s on rate m ove
m ents in e ffe ct at the tim e. As lib e ra liz a tio n has p ro
ce eded , g e nsa ki and o th e r rate forecast e rrors have
converge d. Together w ith the d ecline in g ensaki fo re
cast e rro rs, th is evide nce su g g e sts th a t the Bank of
Japan has been able to in flu e nce the broad spectrum of
m a rke t-d e te rm in e d in te re st rates w ith som ew hat greater
ce rta in ty.
T he fo recast standa rd erro rs fo r m o n e ta ry aggregates
un a m b ig u o u sly point to gre a te r u n c e rta in ty fo r the e s ti
m ated responses a fte r 1984. Both broad m oney and the
m o n e ta ry base responded less p re d icta b ly to call rate
chang es, d e sp ite evid e nce of a more sta b le e conom ic
e n v iro n m e n t. T h is in c re a s e d u n c e rta in ty in b ro a d
m oney re spo nse s m ay in p a rt reflect the d e clin in g use
of m o n e ta ry a gg re gate s in p o licy de te rm in atio n .
Conclusions
We have argued th a t the su b sta n tia l lib e ra liz a tio n of
J a p a n ’s fin a n cial system has profoundly affected both
the fu n c tio n in g of the Ja p a n e se econom y and the c o n
du ct of m o n e ta ry p o licy by the Bank of Japan. The
tig h tly re stricte d fin a n c ia l system in the m id-1970s p ro
m oted a m o n e ta ry po licy stra te g y that used bank credit
to in flu ence activity. F in a n cia l lib e ra liz a tio n has, how
ever, reduced the im p o rta n c e of banks in J a p a n ’s flow of
fu n d s and e lim in a te d a n u m b e r of p o lic y to o ls th a t
re s tric te d b a n k b ehavior. In re s p o n s e , th e B a n k of
Japan has g ra d u a lly becom e more d e p e n d e n t on its
a b ility to in flu e nce m a rke t in te re s t rates to tra n s m it its
policies.
O ur a n a lysis of the m o n e ta ry co n tro l m e ch a n ism s u g
gests th a t in the c u rre n t lib e ra liz e d e n v iro n m e n t, th e
Bank of Japan has been able to tra n s m it its p o lic ie s
e ffe ctive ly th rough m a rke t in te re s t rates. In p a rticu la r,
in te rb a n k in te re st rate m ovem ents, the key o p e ra tin g
ta rg e ts of the Bank of Japan, have pro d u ce d stro n g and
c o n s is te n t in te re st rate resp o n se s a cro ss th e Ja p a n e se
term stru ctu re since 1984. T he in cre a se d re s p o n s iv e
ness of lo n g -te rm bond yie ld s in re ce n t years is p a rtic u
la rly notable. Moreover, som e e v id e n c e s u g g e s ts th a t
the lin k a g e b e tw e e n p o lic y and in te re s t ra te s has
becom e more p re d icta b le .
In co n tra st, the lin ka g e b etw een m o n e ta ry p o lic y and
broad m oney has p ro b a b ly w e a ke ne d . In th e past, p o l
icy actions were la rg e ly tra n s m itte d to broad m oney
through the m oney m u ltip lie r as the B ank of Japa n
d ire ctly in flu e nce d banks’ p o rtfo lio d e c is io n s . Now, ho w
ever, fin a n cial lib e ra liz a tio n has reduced th e B ank of
J a p a n ’s leverage over bank behavior, and p o lic y ’s in flu
e nce on b road m o n e y w o rks m ore c lo s e ly th ro u g h
ch a n g e s in the m o n e ta ry base. A lth o u g h th e average
response of broad m oney to p o licy ch a n g e s rem ains
a bout the sam e, a g re a te r d egree of u n c e rta in ty a c co m
panies the tra n s m is s io n of p o licy to broad money.
Appendix: Cointegrating Money and Interest Rate Models
This appendix expands on the arguments in the text for
cointegration between interest rates and for cointegra
tion in the money supply relation. It also presents the
dynamic error correction models used to predict the
responses of interest rates and monetary aggregates to
monetary shocks.
Cointegration
To investigate the relation between Japanese interest
rates and monetary aggregates, we utilize the cointegra
tion methodology made popular by Engle and Granger.+
This methodology presupposes that a time series contSee R obert Engle and C.W.J. Granger, "C o-integration and
Error C orrection: R epresentation, Estimation, and Testing,"
E conom etrica, vol. 35 (1987), pp. 251-76.
42 FRBNY Q uarterly Review/Autumn 1991
taining a unit root can only be explained over long peri
ods by other series with a unit root. Series with a unit
root (also called integrated series) are predicted well by
their own lagged values. Typically, regression models for
this type of series have substantial residual autocorrela
tion because other variables cannot explain the unit root
component in the outcome variable.
Testing for a unit root in a time series, r„ is commonly
carried out by testing whether the coefficient in a regres
sion on r, i equals one. The literature uses tests based
on the coefficient a in the regression A r, = a r, The
coefficient will be zero if the series has a unit root and
negative otherwise (unless the series is explosive). The
augmented Dickey-Fuller test uses the t-statistic for a in
the regression, adding lagged changes in r, to account
Appendix: Cointegrating Money and Interest Models (continued)
for possible stationary autocorrelation in rt. The StockWatson statistic, as used in this article, is based on
filtering r, to eliminate stationary autocorrelation before
computing the t-statistic. The Phillips-Perron statistic, Z,„
corrects for stationary autocorrelation by applying a nonparametric correction to 7a, where T is the sample
size, in each case, the distribution of the test statistic is
nonstandard and requires specially calculated critical
values.
Several series with unit roots may be linked through a
cointegrating model— that is, a model whose residual
does not contain a unit root. The regressors in this type
of model explain the permanent or unit root component
of the dependent variable. Although the variables in the
regression may deviate from the regression line in the
short run, they return to the regression relationship over
time in the absence of additional shocks. In fact, the
cointegrating model does not specify the dynamic adjust
ment to the model in the long run. This adjustment is
specified by auxiliary equations in the error correction
model that give the dynamic behavior of the variables.
The error correction models are a series of dynamic
equations relating the current change in each variable in
a cointegrated system to the lagged residual from the
cointegrating model and to lagged changes in the vari
ables. The coefficient of the lagged residual in our mod
els measures the speed of adjustment to long-run
equilibrium. Lagged changes appear as explanatory vari
ables in the models to allow variables to have differing
short-term effects.
Thus, we carry out a two-step procedure in our empiri
cal work: First, we test the time series for the presence of
a unit root. If a unit root is found, we proceed to test for
cointegration between those sets of series that could, in
theory, be linked. The unit root tests are shown in Table
A1 and the cointegration tests are discussed in the text of
the article. Second, we use the cointegrating equations
to formulate an error correction model for the dynamics
in the model. This error correction model, examined in
more detail beiow, is used to compute the impulse
response functions reported in the text.
Cointegration in interest rate and money models
The expectations hypothesis of the term structure pro
vides one model where short and long rates will be
cointegrated.* We assume the existence of the following
(approximate) relationship connecting short- and longer
*See Thomas S argent, “A Note on Maximum Likelihood
Estim ation of the Rational E xpectations Model of the Term
S tructure," Journal of M onetary Econom ics, vol. 5 (1979),
pp. 133-43; and John C am pbell and Robert Shiller,
"C oin tegration and Tests of Present Value M o dels,” Journal
o f P olitica l Economy, vol. 95 (1987), pp. 1062-88
term asset returns:
(A.1) r'd.D)
= (1/D) [r*(1,2)
+
f(2,3) + ... + f(D ~1,D )l
where r'ft.D j is the yield to maturity, in D periods, of the
longer term asset; r*(1,2) is the return from period one to
two on the shorter term asset; and f( j, j+ l) is today’s
forward rate between periods j and j+ 1 . The expecta
tions hypothesis connects the forward rate from j io j+ 1
to the expected future spot rate from j to j + 1 as follows:
(A.2)
f(j,j+ 1)
=
E, f Q j + l ) + a(l),
where a is a risk premium and E1 represents expecta
tions at period one. Finally, we suppose that the short
rate follows a simple random walk:
(A.3)
^ 0 + 1,1+2)
=
r*(j,J+1) + e(j+1),
where e is the unexpected component of the short rate.
Since expected future short rates will be directly related
to the current spot rate, E ^Q ) =
the current long
rate will follow:
(A.4)
i*(1,D)
=
r*(1,2)
+
(1/D) Ia (j).
Table A1: Unit Root Tests
(Monthly, January 1974 to May 1991)
Series
ADF
Call rate
Gensaki rate
Long-term bond rate
CD ratef
Loan rate
- 3 36
- 3 .7 3 *
- 1 .6 5
- 2 .3 4
-2 .5 5
In (M2)
In (Base)
- 2 .2 6
- .2 5
In (M2)*
In (Base)*
-2 .8 1
-2 .2 1
'
SW
PP
- 1 4 01*
- 18.38**
- 4 45
- 8 .2 6
-9 .0 1
- 8 .7 2
- 13.09
- 4 .5 4
- 8 .1 0
-6 .6 2
-.6 2
-.1 4
- 9 .6 9
-1 8 .3 7 *
-.4 9
-.4 4
- 5 .0 0
-1 5 .5 4
Notes: ADF is the augm ented D ickey-Fuller s ta tistic (using
seven lags). PP is the P hillips-Perron Zn sta tistic (using seven
autovariance lags). SW is the Stock-W atson sta tistic (using
seven lags). C ritical values for the D ickey-Fuller and P hillipsPerron statistics were ob tained from P hillips and O uiliaris,
“Asym ptotic Properties of R esidual-B ased Tests.” C ritical
values for the Stock and Watson sta tistic are from S tock and
Watson, "Testing for Comm on Trends."
t First sam ple covers 1980-84.
♦Includes time trend.
‘ S ignificant at 10 pe rce n t level.
“ S ignificant at 5 pe rcent level.
“ ‘ S ignificant at 1 pe rcent level.
FRBNY Q uarterly Review/Autum n 1991
Appendix: Cointegrating Money and interest Models (continued)
The short rate may have a unit root if it is influenced by
either inflation or the real interest rate and if changes in
these variables tend to be permanent. Under these con
ditions, equation A.4 implies that the long rate should be
cointegrated with the short rate. Of course, if the risk
premium has a unit root, then the long rate will be
cointegrated with the (measurable) short rate and the
(not directly measurable) risk premium term, and we
would not expect the short rate-long rate pair to be
cointegrated by themselves.®
^Because we do not restrict the term spread to be stationary,
our interest rate m odels are more general than those strictly
im plie d by the expectations hypothesis. A lthough some of
our coe fficient estim ates in the cointegrating interest rate
models seem far enough from one to cast doubt on the
expectations hypothesis, our assets have different issuers
and potentially quite different risk cha racteristics,
co m p lic a tin g the risk term s in our earlier form ulation.
In our money model, we assume that the broad money
aggregate is connected to base money through a money
multiplier, M2 = m Base, where M2 is the broad aggre
gate, Base is base money, and m is the money multiplier.
Broad money and base money would be cointegrated if
the influences on bank and depositor asset allocation
typically only have transitory effects on the money multi
plier. To obtain a cointegrating relationship between the
base and broad money during 1974-84, we find it neces
sary to allow for both trend and interest rate effects.
These modifications suggest that the multiplier has a unit
root arising from trends in asset choice and permanent
interest rate effects.
Dynamic models
The cointegrating models in the text are incorporated
into dynamic models for the call, gensaki, and ten-year
Table A2: Error Correction Models
D ependent Variables
Ind epe nde nt Variables
Call C hanges
Gensaki C hanges
Bond C hanges
January 1974 to D ecem ber 1984
Sum of
56(.28)
lagg ed call changes
lagg ed gensaki changes
- ,06(4)9)
.3 1 (1 5 )
lagg ed bond changes
lag g e d loan changes
.28( 55)
4.73(4.69)
la g g e d M2 changes
-1 .5 8 (3 .3 6 )
lagg ed base changes
.3 5 (2 1 )
—1.11 (.43)
.57( 22)
1.54(.73)
2.37(0.93)
3.86(4.75)
08(.08)
.01 (.06)
.02(.21)
- .07(.24)
.67(3.17)
- 2 36(2.09)
24(.09)
1.12(2.63)
.0 1 (1 4 )
1.43(3.87)
- .1 3 (0 6 )
Residual (gensaki)
R esidual (money)
R2
January 1985 to May 1991
Sum of
lagg ed call changes
lagg ed gensaki changes
lagg ed bond changes
lagg ed loan changes
lagg ed M2 changes
lagg ed base changes
Residual (gensaki)
Residual (long-term bond)
Residual (money)
R2
.22
.30
.12
- .1 8 ( 1 0 )
.39(.26)
,06(,08)
.1 4 (3 6 )
2.32(2.39)
-.9 3 (1 .4 3 )
02 (.17)
- ,23(.30)
. 1 7 (3 8 )
,53(.46)
3.14(4.28)
-2 .8 0 (2 .3 4 )
.1 9 (1 8 )
-2 5 (1 1 )
.1 3 (0 6 )
-1 .8 3 (1 .5 8 )
-r.2 4 (2 1 )
- ,1 8 (.1 5 )
.45
.44
1 0 (0 4 )
,05(,03)
11 (.03)
58(.07)
— 1.06(.69)
.3 2 (4 6 )
—
~ .72 (.31 )
.89(.27)
.1 0 (1 1 )
.7 6 (4 0 )
4.84(3.77)
-2 .8 7 (2 .2 6 )
-2 .6 9 (2 .4 9 )
Loan C hanges
.87
.03( 05)
.0 2 (0 7 )
,09(.03)
.8 2 (0 6 )
22( 49)
.. 25( 27)
M2 C hanges
.002(.002)
.000(.001)
-.0 0 1 (.002)
- .008(.006)
,90(.08)
—,04(.05)
.01( 004)
.001 (.003)
.001 (.005)
- ,02 (.01)
,33(.18)
,16(.24)
-.0 6 (.0 4 )
.20(.08)
.01
- .0 0 0 ( 0 0 3 )
.004(.006)
.0 0 5 (0 0 3 )
- .02(.01)
1.08(.14)
- .21(.06)
.26
-,0 0 2 (,0 0 7 )
006(.013)
006(.006)
.0 0 9 (0 1 8 )
18(.25)
,48(.32)
—
- .1 0 (0 7 )
.17
Base C hanges
.91
.26
.6 0 (1 5 )
.40
Notes: S tandard errors are shown in parentheses. The residual for the gensaki rate is obtained from the c o in te g ra tin g m odel con n e ctin g the
gensaki and call rates; the residual for the long-term bond rate is obtained from the co in te g ra tin g m odel co n necting the long-term
governm ent bond and gensaki rates; the residual for money is the residual from the cointegra ting m odel for money supply.
44 FRBNY Q uarterly Review/Autumn 1991
Appendix: Cointegrating Money and Interest Rate Models (continued)
government bond rates as well as the monetary base and
the M2 + CDs aggregate. Since we do not detect a long
term relationship between the loan rate and other inter
est rates, our cointegrating equations do not include the
loan models presented in Table 6 of the text. The coin
tegrating, or long-run, interest rate models are:
(A.5) G = a0 + a,C +
LJ — a 2 + a3G
+
/== 1
€lJ,
ln(M2) = k0+ In(Base) + k,G
+ k2C + k3t + eM2.
Our estimates of these models are given in text Tables
6 and 7.
The dynamic equations have a general error correction
form that relates the current change in each variable to
lagged changes in all of the variables and to the
residuals from the cointegrating coregressions.+t Esti
mates of these dynamic models are shown in Table A2.
We typically include four lags of the dependent variable
and one lag of the other variables.§§ln most cases, we
If The main text presents a c o integra ting model connecting the
governm ent bond yield to the call rate. Our m odeling
stra tegy uses the equation A .6 connecting the bond yield
and the gensaki rate as the long-term relation for the bond
yield. The correspo ndin g equations are:
1974-84 LJ = 275 + 28G. R2 = .51
1985-91 LJ * 1.01 + 84G, R2 = .86.
There is strong evidence of cointegra tion in the second
period and essentially none in the first.
ttF or details, see Engle and Granger, "C o-integ ration and
Testing," and Jam es Stock, “A sym ptotic Properties of Least
Squares Estim ators of C ointe gra ting Vectors," Econom etrica,
1987, pp. 1035-56
SfOur cho ice of lag structure is m otivated by the
autocorrelations of first differences of the data. These
generally seem consistent with fou rth-ord er autoregressive
m odels. We in clu d e more lags of other interest rates in the
loan equation to allow for possible effects over several
periods.
AC(f) = —5ieG(f —1) —82e ^ (f-1 )-8 3eM2(t-1 )
4
+ X P,AC(f - i ) + y-[ AG(t —1) + ^ ALJ(t -1 )
+ 01ALoan(?-1) + (,1AlnM2(t~'\) + <i>,AlnBase(t-l).
where C represents the call rate, G is the gensaki rate,
and LJ is the Japanese benchmark long-term govern
ment bond yield.!l Our cointegrating equation for broad
money supply relates broad money, M2, to the base,
Base, the gensaki rate, G, the call rate, C, and a time
trend t, restricting the coefficient on the base so that the
interest rates and time trend affect the money multiplier:
(A.7)
(A.8)
eG
and
(A.6)
only include residuals from cointegrating equations that
contain the dependent variable. The general form of the
model is illustrated by the call rate equation below:
The lagged residuals in this equation ensure that the
short-run behavior in the dynamic model will converge to
the long-run behavior embodied in equations A.5, A.6,
and A.7. The lagged changes of the variables allow the
short-run impacts of interest rate shocks to differ from
the long-run behavior in the cointegrating equations.
Analogous to the call rate equation are the equations for
the gensaki rate, the Japanese long-term government
bond yield, and bank loan rate given below:
(A.9)
AG(f) = -S 1eG( f- 1 ) - 5 2€ ^ (f-1 )-S 3eM2(t-1 )
4
+ M C (f~ 1 ) 2 "y, AG(f - /) + ALJ{t -1 )
/'= 1
+ 0, ALoan(t -1 ) + £AlnM2(t -1 ) + <J>,AlnBase(t -1),
(A.10)
- S1c<3(f—1) —82€ ^ (f-1)
4
+ p1AC (f-1) + 7 lA G (f-1)+ £ 4/,ALJ( t- i)
i- 1
+ 0, ALoan{t-1 ) + £, AlnM2(t -1 ) + <}>,AlnBase{t -1),
ALJ(t) =
and
(A.1!)
ALoan{t)=
3
X
/'=0
3
P,AC(f-/)+ X
3
3
•+ £ %ALJ( t - i) + 2
i= 1
+ £1 AlnM2(t -1 )
i=
y A G (t-i)
1
®ALoan{t-i)
/'=1
+ 4), MnBase(t -1).
These equations allow us to estimate the dynamic
response of Japanese interest rates to shocks in the call
rate (our proxy for Japanese monetary policy actions).
The dynamic error correction equations for the mone
tary base, Base, and broad money, M2, have the follow
FRBNY Q uarterly Review/Autum n 1991
45
J|§
f ji
| §*
A n n o n H iv P
n in to n rn tin n Money
M
Appendix:
Cointegrating
and Interest Rate Models (continued)
ing forms, in which the lagged residual is obtained from
equation A.7:
(A.12)
AlnM2(t)= -5 e M2(f-1 ) +
0, AC(t -1 ) + , AG(f -1 ) + 4>,ALJ(t -1 ) + d>,ALoan(t -1 )
7
4
+ I
1
-/) + <{>!MnBase(t -1 )
I—I
and
(A.13)
AlnBase{t) = - BeM2(f -1 )
+ 01A C (f-1) + 7 tA G (t-1) + \\>,ALJ( t - 1) + (M Z-oanff- 1 )
siai®
4
+ [,,& lnM 2(t-1)+ ' i <$>,UnBase(t-i).
/= 1
46 FRBNY Q uarterly Review/Autumn 1991
To simulate the models and to compute forecast stan
dard errors, we have to impose a structure on the current
disturbances in the error correction equations. We use
an ordering of the disturbances (call, base, gensaki,
long-term bond, M2 + CDs, loan) in which current shocks
to each variable in the list affect contemporaneous
shocks in variables listed later. Our ordering allows the
current call rate shocks to affect shocks in all the other
variables in the system. When we analyze predictability
of policy responses, we remove the component attribut
able to call rate shocks from the forecast standard errors
to exclude uncertainty related to policy changes.
Although these standard errors correctly measure uncer
tainty when monetary policy is designed to minimize the
variance of a single variable, the interpretation of the
errors may be more difficult when monetary policy has
multiple objectives.
............ ,
‘ ..............r .ll..-
Expected Inflation and Real
Interest Rates Based on
Index-linked Bond Prices:
The U.K. Experience
by Gabriel de Kock
Recently some analysts have suggested that the Trea
sury finance part of the federal deficit by floating
indexed bonds.1 One of the claims made for this strat
egy is that it would yield significant monetary policy
benefits. In particular, the prices of indexed bonds could
offer timely and accurate market measures of expected
inflation and ex ante real interest rates. As such they
could provide the Federal Reserve System with valu
able information about market perceptions of, and reac
tion to, its policies. The argument has also been made
that a real-time market measure of expected inflation
might provide the Federal Reserve System with a valu
able indicator of the future course of inflation and offer
the public a ready means of monitoring the Fed, thereby
encouraging public interest in better policies.
A market measure of expected inflation may be useful
in the formulation of policy even if is not a good gauge
of the actual future inflation performance of the econ
omy. It may be a poor indicator of future inflation
because private sector inflation expectations are, in
fact, not realized. Even in this case, however, the asso
ciated real interest rate should nevertheless be a good
measure of ex ante real interest rates faced by the
private sector. That is, at the macroeconomic level,
indexed bond prices should contain useful information
about real economic activity.
The potential value of a real-time market measure of
expected inflation to policy makers can only be
assessed indirectly because no “true” alternative mea
sure of private sector inflation expectations exists. This
article evaluates the usefulness of a market measure of
expected inflation by applying two closely related tests:
’ See, for e xam p le, R obert H etzel, "A Better Way to Fight Inflation,”
The W all S tre e t J o u rn a l, April 25, 1991.
first, whether the market measure of expected inflation
is a good indicator of inflation developments; and sec
ond, how well the market measure captures private
sector inflation expectations even if the expectations do
not reflect actual inflation performance.
The second of these tests is based on two proposi
tions. The first proposition is the familiar Fisher hypoth
esis that nominal interest rates should equal expected
inflation plus the ex ante real interest rate; the second,
a prediction shared by all standard dynamic mac
roeconomic models, is that true real interest rates
should provide information about future economic activ
ity.2 If the market-expected real interest rate, measured
as the difference between a long-term nominal rate and
the market measure of expected inflation, has no signifi
cant effect on future real economic activity, it seems
likely that it is also a poor measure of the true real interest
rate. Under these circumstances, the market measure
of expected inflation would then seem likely to be a poor
measure of “true” private sector inflation expectations.
If a real-time measure of inflation expectations neither
anticipates future inflation developments nor conveys
useful information about real economic activity, it is
probably of only limited use to policy makers.
Drawing on this framework, this article examines the
U.K. experience with indexed gilts (IGs) to assess
whether indexed bond prices convey information useful
in form ulating and m onitoring m o n etary policy.
2ln theory, higher real interest rates do not necessarily lead to lower
real econom ic activity (G N P ); in fact, the two variab les m ay be
positively a sso ciated with e ac h other. In the context of s tan d a rd
m acroeconom etric m odels, however, real interest rates and
econom ic activity are, ceteris paribus, negatively related to e ac h
other.
FRBNY Quarterly Review/Autumn 1991
47
More specifically, the article evaluates whether the
expected inflation rate derived from the prices of
indexed and nominal bonds predicts future inflation and
whether the corresponding expected real interest rate
provides information about real economic activity not
obtainable from more traditional information variables
and measures of policy stance. Although several coun
tries have experimented with indexation since World
War II, the experience of the United Kingdom, where
marketable index-linked gilts have been issued since
March 1981, is likely to be the most relevant to the
United States.
While the market for index-linked gilts does provide a
real time measure— accurate or not— of expected infla
tion and ex ante real interest rates, this information
does not appear to be of much practical value in for
mulating and evaluating monetary policy. This general
interpretation of the data derives from two specific con
clusions: First, the expected inflation rate embodied in
nominal bond yields is no better than simple measures
of inflation expectations based on past inflation alone. It
is a biased predictor of the future level of inflation,
although it does provide minimal information about
acceleration and deceleration of inflation. Second,
indexed gilt prices do not seem to provide information
about future movements in real economic activity, sug
gesting that the real interest rate on indexed gilts is
unlikely to be a good measure of ex ante real interest
rates faced by the private sector in the markets for
goods and services. The expected inflation rate embod
ied in U.K. bond yields, therefore, appears to be a poor
measure of true inflation expectations, given our empiri
cal results. By contrast, lagged inflation and nominal
interest rates often used to derive real interest rates
that may measure monetary policy do have predictive
content for U.K. real GNP growth.
The first section of the article summarizes the devel
opment of the U.K. indexed gilt market. More specifi
cally, it focuses on the particular circumstances
surrounding the introduction of IGs in the United King
dom and the main features of the IG market at present.
There follows a brief discussion of the potential mone
tary policy role of indexed bonds and a review of the
decomposition of nominal yields into expected real
interest rate and expected inflation components. The
article then examines the information about future infla
tion provided by indexed bonds and the ability of
indexed gilt prices to predict developments in real eco
nomic activity.
The evolution of the U.K. market
for index-linked gilts
The Conservative government introduced IGs in 1981 as
part of its anti-inflation program. Three reasons were
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48 FRBNY Quarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
given for the move: (1) the introduction of IGs would
improve the Bank of England’s control over monetary
aggregates, (2) indexation would result in substantial
savings to the Treasury if, as anticipated, inflation
declined significantly as a result of the government’s
policies, and (3) indexing government debt would signal
the government’s determination to reduce inflation.
These reasons reflected policy concerns specific to the
United Kingdom in the early eighties, although the third
is sometimes viewed as relevant to current U.S. policy,
if only on a theoretical level.3
Issuing IGs was expected to bring about closer con
trol over the monetary aggregates by ameliorating con
straints on monetary policy imposed by the distinctive
structure of the U.K. gilts market. Because market
makers, considered essential to the smooth functioning
of the market, were weakly capitalized, the authorities
felt obliged to minimize fluctuations in gilt prices that
would threaten the market makers’ survival. Thus, the
Bank of England was constrained to stabilize nominal
interest rates, a policy that entailed loss of control over
monetary aggregates.4 Most notably, in the late seven
ties and early eighties, market expectations of rising
inflation forced the Bank of England to follow a
destabilizing expansionary policy to prevent gilts prices
from falling too steeply. Index-linked gilts, which could
be sold in times of market expectations of rising infla
tion, enabled the authorities to reestablish control over
monetary aggregates. In this way, inflation expectations
could be kept in check, thereby mitigating fluctuations
in conventional gilts prices.
The second argument for issuing indexed bonds, that
the real cost to the government of issuing index-linked
debt would be lower than that of borrowing on conven
tional terms, is valid if the government expects an infla
tion rate lower than the market expectation of inflation
embedded in nominal yields (assuming that the tax
system is neutral with respect to inflation). These condi
tions were clearly fulfilled in the United Kingdom in
early 1981: long-term bond rates were close to 14 per
cent and retail prices were still rising rapidly, but the
government expected its firm anti-inflation policies to
pay off in the near future. However, under the indexation
scheme envisaged, the tax system would not be neutral
with respect to inflation, because the inflation compo
nent of nominal rates would be fully taxable while only
3See C harles A. E. G o odh art, M o ne y, In fo rm a tio n a n d U n c e rta in ty
(C am brid ge: M .I.T. Press, 1989), for a discussion of the policy
d e b a te surrounding the introduction of in dexed gilts.
«The Bank of England could not use op en m arket o p eration s at the
short end of the m arket b e c a u s e the stock of Treasury bills
outstanding was very small by U .S. stan dard s. At the e nd of M arch
1980, for exam ple, Treasury bills a c c o u n te d for only 2 .9 p e rce n t of
m arket holdings of governm ent deb t.
part of the inflation compensation on index-linked gilts
would be taxed. Thus, although government interest
outlays would be lower with indexed gilts, conventional
gilts could be expected to produce much more tax
revenue. Initial calculations suggested that inflation
would have to decline very rapidly before the reduction
in outlays brought about by indexation IGs would
exceed the loss in revenue entailed.
The government’s third reason for introducing indexed
bonds— to enhance the credibility of its anti-inflation
program— derives from the fact that index-linking
reduces the benefits of unanticipated inflation. Investors
will be justly skeptical of announced anti-inflationary
policies if the government at the same time issues
nominally denominated debt, because the government
may always be tempted to resort to unanticipated infla
tion to reduce the real value of its debt. By contrast,
unanticipated inflation does not promise any capital
gains to the government if its debt is indexed. Conse
quently, by issuing indexed debt, the government could
enhance its credibility.5
The advantages of IGs were partly offset by a number
of possible disadvantages. First, there was concern that
issuing an attractive long-term asset could have a nega
tive impact on equity prices and corporate financing
opportunities. Second, it was feared that foreign
demand for IGs might put upward pressure on the
pound, which was already overvalued as a result of tight
monetary policies and the discovery of North Sea oil.
Third, since capital gains were not yet indexed for tax
purposes, issuing indexed debt would entail the taxa
tion of purely inflationary capital gains on IGs— a step
that could in turn stimulate political pressure for the
indexation of taxes. Finally, the United Kingdom faced
strong political pressure from other OECD governments
concerned that any form of indexation would fuel OPEC
pressure to index-link oil prices.
Indexed gilts were issued consistently throughout the
1980s at coupon rates mostly between 3 and 4 percent
(compared with 2 percent for the first two issues). As
early as end-March 1985, IGs made up 6.5 percent of
total market holdings of U.K. public debt; by March
1990, the total amount of IGs outstanding was about
5Note, however, that indexation could also have an adverse im pact
on exp ectatio ns (and thereby m ake it more difficult to reduce
inflation) if it was interpreted as an effort by the m onetary
authorities to dec re a s e the political cost of inflation before giving
up the battle against inflation altogether. But such an adverse
im pact would be more likely if indexation covered a w ide range of
co n tra cts — som ething the C onservative governm ent had taken great
pains to avoid. Issuing index-linked gilts probably also enhanced
the go vernm ent’s credibility over tim e bec a u s e it effected im m ediate
cosm etic im provem ent in the Public Sector Borrowing Requirement:
In d e x a tio n was im plem ented in a way that pushed com pensation for
inflationary d e p reciation of principal into the future, w hereas the
governm ent would have had to pay higher nominal interest rates
im m ed iately had it issued conventional bonds.
£17.5 billion, or 10.9 percent of market holdings of
British government debt and 19.3 percent of the value of
gilts outstanding (Table 1).6 The stock of IGs outstand
ing is made up of thirteen issues with maturities varying
from two to thirty-three years. Long-dated issues make
up the bulk of the value of IGs outstanding; only 7.3
percent of the amount outstanding are of maturities
shorter than five years, and 16.5 percent of maturities
shorter than ten years (Table 2).
Anecdotal evidence suggests that holdings of IGs
remain concentrated and that the number of customers
remains small.7 The most important holders of IGs are
pension funds, followed by insurance companies, while
individual investors are largely confined to the short end
of the market. These features match those of the gilt
market as a whole. The IG market is thin in comparison
with the market for conventional gilts. Although turnover
varies, it only amounted to 2.9 percent of total gilt
turnover in 1990 and 3.1 percent for the first four months
of 1991 (Table 2). The demand for new issues of IGs has
been disappointing when the ex post real yields on
conventional gilts exceed those on IGs, as they did
during the rapid decline of inflation in 1982-83 and
1985-86. (The issuance of IGs also resulted in substan
tial savings to the Treasury during this period for the
same reason.)
The potential role of indexed bonds
in monetary policy
Proponents of issuing indexed bonds in the United
States have emphasized monetary policy benefits that
depend critically on the informational role of indexed
bond prices. This consideration figures importantly in
the academic literature on indexed bonds although it did
not arise in the U.K. policy debate.8 Advocates argue
6By the end of 1990 this p e rc e n ta g e had risen to 2 0 .5 p ercen t
b e cause under the s ch em e of indexation used in the U nited
K ingdom , the value of the IG s ou tstand ing rises in line with
inflation. N onm a rk e ta b le national savings c ertific ate s or “granny
bonds," which have been issued since 1975, have had a lim ited
im pact, m aking up less than 2 p e rce n t of m arket holdings of British
governm ent d e b t as of e n d -M a rc h 1990.
7The Bank of England does not com pile s e p a ra te statistics on
holdings of IGs and conventional gilts by type of institution. At the
end of M arch 1990, pension funds and in surance co m p a n ie s held
5 7 .5 percen t of U.K. governm ent d e b t ou tstand ing , while individuals
and private trusts held about 3 8 p e rce n t (“The N et D eb t of the
Public Sector: E n d -M arc h 1 9 9 0 ,” B a n k o f E n g la n d Q u a rte rly
B u lle tin , N ovem ber 1990, pp. 519 -2 6 ).
8A theoretical literature on the relative e fficiency of op en m arket
operations in indexed and nom inal bonds d ates b ack to Jam es
Tobin, "An Essay on the P rinciples of Public D eb t M a n a g e m e n t,”
reprinted in M a c ro e c o n o m ic s , vol. 1 of E s s a y s in E c o n o m ic s
(M arkham Publishing C o., 1971). Tobin a rg u e d that op en m arket
operations in in dexed bonds will affect real activity more strongly
and with greater certa in ty than will op en m arket op eration s in
FRBNY Quarterly Review/Autumn 1991
49
Table 1
Com position of U.K. National Debt
March 1985
Billions
of Pounds
M arket ho ld ings
S terling m arketable debt
G overnm ent stock
Index-linked
Other
Treasury bills
S terling nonm arketable de bt
National savings'^
Index-linked
Other
Other
Foreign currency de bt
O fficial holdings
M arch 1990
Percentage
B illions
of Pounds
P ercentage
146.7
100.0
160.0
100.0
114.4
78.0
117.0
73.2
9.5
103.7
1.2
6.5
71.5
0.8
17.5
90.5
9.0
10.9
56.6
5.7
29.3
20.0
36.5
22.8
3.6
18.8
6.9
2.4
12.8
6.8
3.0
26.1
7.4
1.9
16.3
4.6
2.9
2.0
6.5
4.0
11.6
32.5
Source: Bank of England.
t N ational savings includ e a variety of no n-negotiable savings instrum ents issued by the governm ent.
th a t indexed bond p rice s w ould provide p olicy m akers
and the pub lic w ith info rm a tio n on inflation e xpectations
and ex ante real in te re st rates on a re al-tim e basis. In
this view, the Federal R eserve System w ould gain valu
able info rm a tio n about m arket p e rce p tio n s of, and reac
tio n to, its po licies. F u rtherm ore, this in form ation could
provide p o licy m akers w ith a good pre d icto r of inflation
and a m easure of the im pact of m o n e ta ry p o licy on both
in fla tio n and real eco n o m ic activity. H etzel has argued
th a t the ready a va ila b ility to policy m akers and the
pub lic of an in d ica to r of the in fla tio n a ry co n se q u e n ce s
of m o n e ta ry p o licie s w ould have three benefits. F irst, it
w ould increase pu b lic u n d e rsta n d in g of, and su p p o rt for,
a n ti-in fla tio n a ry policie s. S econd, it w ould serve as a
b a ro m e te r of Fed cre d ib ility and co n se q u e n tly increase
in ce n tive s for the Fed to c o m m it itse lf to a n ti-in fla tio n
a ry p o licie s. Finally, by exposing the true co n se q u e n ce s
of p o licie s th a t trade o ff in flation for s h o rt-te rm outp u t
g ains, it w ould stren g th e n the Fed’s e ffo rt to focus
a tte n tio n on its lo n g -te rm p rice s ta b ility o b je c tiv e s .
Table 2
Features of U.K. Market for Index-linked Gilts
M aturity C om position of IGs O utstanding
M illions of Pounds1
Less than one year
One to five years
Five to ten years
Over ten years
Total
P ercentage
865
638
1,920
17,365
20,788
4.2
3.1
9.2
83,5
100.0
C ontributio n of IGs to M onthly G ilt Turnover
Total Turnover_____IG Turnovef____
(B illions
(B illions (Percentage
of Pounds) of Pounds) of Total)
1990 average
January-April 1991 average
75,445.5
85,702.9
2,196.6
2,630.6
2.9
3.1
Source: Bank of England.
tln c lu d e s indexation of p rin cip a l up to the b e ginnin g of 1991.
Footnote 8 (continued)
nominal bonds because indexed bonds are likely to be a closer
substitute for equity than nominal bonds. Tobin has been criticized
by Stanley Fischer, “The Demand for Indexed Bonds," Journal of
P olitical Econom y, vol. 83, no. 3 (June 1975), pp. 509-34, and by
Paul Beckerman, "Index-linked Government Bonds and the
Efficiency of Monetary Policy," Journal o f M acroeconom ics, vol. 2,
no. 4 (Fall 1980), pp. 307-31. Fischer suggested that open market
operations in nominal bonds that are complements for equity in
private portfolios may have a more pronounced impact on Tobin’s q,
and thus on real activity, than open market operations in indexed
bonds that serve as substitutes for equity. Beckerman has pointed
out that Tobin's conclusion requires a set of potentially inconsistent
assumptions.
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50 FRBNY Q uarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
However, the b e n e fits cite d by H etzel are o n ly lik e ly to
m a te ria lize if the m a rke t m easure of e xp e cte d in fla tio n
is a re liable in d ic a to r of the e ffe cts of m o n e ta ry p o lic ie s
and m a cro e co no m ic d is tu rb a n c e s on in fla tio n .9
9Earlier advocates of indexation— for example, Alicia H. Munnell and
Joseph B. Grolnic ("Should the U.S. Government Issue Index
Bonds," New E ngland E conom ic Review, September-October 1986,
pp. 3-22)— have emphasized the provision of index-linked
government liabilities that could be used to back indexed pension
contracts.
To be sure, a market-based decomposition of nominal
interest rates into their expected inflation and expected
real interest rate components may be useful to policy
makers, even if the market measure of expected infla
tion is not a good predictor of inflation. This would be
especially true if the market measure of expected infla
tion were a good gauge of private sector inflation expec
tations. If so, it would provide a reliable measure of ex
ante real interest rates and thus convey information on
private decisions and future economic developments.
More generally, indexed bond prices could offer policy
makers and private agents up-to-date information about
the sources of macroeconomic disturbances. In fact,
Boschen, using a simple model, has shown that a mar
ket for indexed bonds could reduce the magnitude of
business cycle fluctuations by allowing private agents to
distinguish real and nominal disturbances more
accurately.10
The empirical analysis in this section evaluates
whether the U.K. market for index-linked gilts actually
conveys the policy-relevant information about future
inflation and real economic activity that advocates of
indexed bonds have attributed to it. Data on IG and
conventional gilt prices are used to construct a monthly
series of expected inflation rates and expected real
interest rates spanning the period from March 1982 to
March 1991. As detailed below, these data indicate that
the derived measure of expected inflation, termed the
IG measure, is a poor predictor of inflation and that the
IG market does not provide information about future real
economic activity.
Inflation expectations and expected real interest rates
derived from IG prices
The data on expected real interest rates and expected
inflation are constructed by using an indexed bond's
price to decompose the yield on a nominal bond into
expected real interest rate and expected inflation com
ponents. The calculation assumes that the expected
real yields of indexed and conventional bonds of the
same maturity must be equal and consequently that the
IG measure of the inflation rate expected by investors to
prevail over the rem aining lifetim e o f the bonds can be
estimated as the difference between the redemption
yields on the nominal and indexed bonds. In practice,
the calculation and interpretation of the expected infla
tion rate and expected real rate are somewhat more
involved, because IGs are not fully indexed, investors
may be risk averse, and the indexed and conventional
bonds may differ in liquidity and tax treatment. More
detailed information on the nature of indexed gilts and
10S ee John F. B oschen, “The Inform ation C ontent of In dexed B onds,"
J o u r n a l o f M o ne y, C re d it, a n d B a n k in g , vol. 18, no. 1 (February
1986), pp. 7 6 -8 7 .
the calculation and interpretation of expected inflation
and real interest rates is given in the box.
Data on the IG measure of expected inflation and the
corresponding long-term real interest rate are derived
by decomposing the nominal yield on a conventional
bond maturing in 1996. This calculation yields the long
est data series because the first IGs issued mature in
1996. The top panel of Chart 1 illustrates the decom
position of this long-term yield into its expected real
yield and expected inflation components. Note that
there is a break in the series in 1986.11 For reference,
the lower panel of Chart 1 shows the nominal yield on
the 1996 bond along with the yield on ten-year gilts. The
yield on the 1996 bond moves closely with the yield on
ten-year gilts, confirming that the 1996 bond (as well as
the decomposition of its yield) is representative of the
long-term government bond market in the United
Kingdom.
Chart 1 suggests that changes in expected inflation
account for the bulk of nominal interest rate move
ments. This result is the counterpart of the striking
stability of the real interest rate measure, which varies
between 2.36 percent and 4.37 percent per annum. The
underlying stability of the real yield on IGs is confirmed
by the Bank of England’s calculations: although based
on the assumption of a fixed 5 percent inflation rate
over the remaining lifetime of the IG, the Bank’s esti
mate of the real yield varies over a similar range.12
Chart 1 also illustrates the response of the IG mea
sure of expected inflation and the corresponding real
11The data for the period from O c to b er 1986 to M arch 1991 p ertain to
a 10 percent coupon bond m aturing in N ovem ber 1996 and those
for the period from M arch 1982 to S e p te m b e r 1986 to a 14 p ercen t
coupon bond m aturing in July 1996. The decom p o s itio n of the
nominal yield is bas e d on price d a ta for a 2 p e rce n t in d e xe d gilt
m aturing in S e p te m b e r 1996. In both cas e s the m aturity m atch is
probably close enough not to affect the results. The decom position
is som ew hat sensitive to the p a rtic u la r m atc h e d -m a tu rity
conventional bond used, presum ably b e c a u s e the different bonds
are not equally liquid. The yields on the 14 p e rce n t bond and the
10 percent bond differ by 2 9 basis points, on ave ra g e , in the
months for which overlapping ob servations are available. Our
results nevertheless in dic a te that only the le v e ls of c a lc u la te d
e x p e cte d real yields and inflation rates are a ffec te d , not their
m ovem ents over tim e. For e xam p le, the correlation coefficient of the
two exp e cte d inflation rates c a lc u la te d for the overlapping
observations is 0 .9 9 . For the pu rp oses of C h a rt 1 and the em p irical
analysis discussed below, the e arlie r observations w ere a d ju ste d by
the m ean difference c a lc u la te d from the overlappin g observations.
Specifically, the real rate in creased abou t 6 basis points and the
e xp e cte d inflation rate d e c lin e d by abou t 3 5 basis points.
12The Bank of E n g la n d ’s real interest rate m easure is som ew hat
higher, on average, than the m easure rep orted here (3 .6 4 p ercen t
com pared with 3 .5 4 p e rce n t) and som ew hat less volatile; its
s tandard deviation is abou t 4 0 .5 basis points, com p a re d with 4 2 .6
basis points for the m easure used here. The correlation coefficient
of the two m easures is 0 .9 3 . In “Sources of Fluctuation," G aske
docum ents that in the early part of the sam ple the co-m ovem ents
betw een the Bank of England's series and m acroecono m ic variab les
are quite different from those of an ex ante real rate m easure like
the one used in this article.
FRBNY Quarterly Review/Autumn 1991
51
Box: Extracting Ex Ante Real Yields and Expected Inflation from Indexed Gilts Prices
Features of U.S. index-linked gilts
The form of index-linking adopted in the United Kingdom
may be called principal value indexation. The value of the
principal of an IG is linked to the retail price index (RPI),
and coupon payments, payable every six months, are
calculated as a fixed percentage of this inflation-adjusted
principal. Holders of IGs are not fully protected against
inflation, and hence the real return on an IG is uncertain,
because the principal— and consequently interest pay
ments— are indexed to the RPI with an eight-month lag.
That is, the value of the bond for the purpose of calculat
ing the coupon payment for a given six-month interest
period exceeds its initial face value by the increase in the
RPI over the period starting eight months before the
issue date and ending eight months before the date on
which the coupon payment is made. For example, the
principal in period t of a £100 bond issued on date I and
paying a 2 percent coupon is £100x(RPIt _8/RPI,_8), and
the coupon payment on date t will be £ 1 x(R P It_a/
RPIi „ 8), where time is measured in months. Note also
that, because of the lag in indexation, an IG is a pure
nominal bond during the last eight months of its lifetime.
The eight-month lag is needed to ensure that the rate of
interest accrual in money terms for any six-month period
is known before the start of that period so that pur
chasers can compensate sellers for interest accrued
since the last coupon payment preceding the transaction.
The period for which interest is due accounts for six of
the eight months; the normal lag in availability of the RPI
data for the seventh month; and the need to avoid prob
lems that could arise if the publication lag exceeded one
month, for the eighth months
Eligibility to take up the initial offerings of IGs in 1981
and early 1982 was restricted to domestic tax-exempt
institutions (pension funds, life insurance companies tak
ing pension business, and charitable societies) in order
to forestall potential tax problems and to avoid repercus
sions for the exchange rate. These restrictions on the
ownership of IGs became redundant and were removed
when the government introduced indexation of capital
gains for tax purposes in March 1982. Since that time,
IGs have enjoyed a significant tax advantage relative to
conventional gilts. While holders of conventional gilts are
taxed at the income tax rate on nominal interest earn
ings— a rate that consists in part of compensation for
tFor a discussio n of institutional features of the indexed gilt
market, see Patrick P hillips, Inside the New G ilt-e dged
M arket, 2d ed. (C am bridge, England: W oodhead-Faulkner,
1987).
FRBNY Q uarterly Review/Autumn 1991
depreciation of principal— holders of IGs pay no taxes at
all on inflationary increases in the nominal value of the
IGs. Consequently, an anticipated increase in inflation
will tend to depress conventional gilts prices relative to
IG prices by more than necessary to equate pretax
nominal yields on conventionals and IGs.
Calculation method
The imperfect indexation of IGs makes it impossible to
calculate an expected real redemption yield on the IG
without making an assumption about inflation over the
bond’s remaining lifetime. Nevertheless, as long as
investors are risk neutral, the prices of an IG and a
nominal gilt of matched maturity can still be used to
decompose the nominal yield on the conventional gilt
into an expected real rate and an expected inflation rate.
The procedure used to calculate the expected real yield
and the expected inflation rate derives from work by Arak
and Kreicher, Woodward, and Gaske.* It can be
explained by a simple example that captures the salient
features of the U.K. gilt market. Consider a maturitymatched pair of nominal and indexed bonds with face
values Fn and F (at issue) maturing in period T. Let P"
and P't denote the (nominal) prices of the nominal and
indexed bonds, respectively, at the beginning of period t,
with time measured in months. Coupon payments are
made every six months. The first payment after period t
occurs in period t+ j (j«6), and the last payment coin
cides with redemption in period T. The coupon payment
on the nominal bond is denoted by Cn. If the IG was
issued in period I, its redemption value will be FV= F® x
(RPIT_8/RPI, „ 8), and the nominal coupon paid in period
t + j will be CUj = O x (RPI,+j_8/RPIi-e)> where C® is the
face coupon (C1 = F® x c = face value x coupon rate)
on the index-linked bond. Finally, let it denote the periodt annual yield to maturity on the nominal bond, -nf the
annual inflation rate expected to prevail from period t to
period T, and rt the annual expected real interest rate
from period t to period T. Then i, is the solution to
*See M arcelle Arak and Lawrence Kreicher, "The Real Rate of
Interest: Inferences from the New U.K. Indexed G ilts ,”
International Econom ic Review, vol. 26, no. 2 (June),
pp. 399-407; G. Thomas W oodward, "C om m ent: ‘The Real
Rate of Interest: Inferences from the New U.K Indexed
G ilts,'" International E conom ic Review, vol. 29, no. 3
(August), pp. 565-68, and Mary Ellen Gaske, "S ources of
Fluctuations in E xpected Long-term Real Rates: E vidence
Extracted from U.K. Indexed Bond R ates,” U npu blished
Ph.D. D issertation, U niversity of M aryland.
Box: Extracting Ex Ante Real Yields and Expected Inflation from Indexed Gilts Prices (continued)
K
(A.1) Ptn =: C" J
(1+ it/1 2 )-« *+i> + {1 + it/12)-(T-«)Fn,
k= 0
where K * [ T - ( t + j)]/6. The decomposition of i, into irf
and r, must satisfy
K
(A.2) Pj= J
[(1 + r,/12)(1 +irf/12)]-U+6k)C[+t+6k ]
k= 0
+ [(1 + r,/12)(1 +irf/12)]-(T-'>FV
and
(A.3)(1 + r,/12)(1 + Trf/12) = (1 + it/12).
Since the expected nominal coupon on the indexed bond in
period t+ j + 6k is
Ci +j +6k = (1 + 'jrf/12)6kCj+j • (1 + irf/12)6k
X
(RPIt+j. 8/R PI,.8) x O
and the expected nominal redemption value of the indexed
bond is
FV= (1+'rrf/12)T- t' 8 x (RPI,/RPI,_8) X P,
equation A.2 can be simplified to
(A.4) P; = (R PIt+j. ft/R P tl . 8)[(1 +n f/1 2 )(1 + rt/1 2 )]-iO
K
+ (RPIt/R P I,.a)(1 + irf/1 2 )-8{ J (1 +r,) <6k+i>Ci
+ (1 +r,)~<T-,)F'j.
k= 0
The gilt prices provided by the Bank of England are
clean prices: that is, they do not include accrued interest.
For these calculations, accrued interest was added to the
clean prices in proportion to the number of months that
had elapsed since the previous interest payment.
Limitations of the IG measures of the expected
inflation and real interest rates
The decomposition of the nominal long-term bond rate
into its expected inflation and real interest rate compo
nents will be strictly correct if two requirements are met:
(1) investors are risk neutral, and (2) nominal and
indexed-linked bonds are identical in all respects other
than indexation (specifically: risk, maturity, liquidity, and
tax treatment). If investors are not risk neutral, interpret
ing the IG measure of expected inflation and the corre
sponding expected real yield is complicated by the
existence of risk premia. Suppose, for expository pur
poses, that the indexed bond is fully indexed so that the
ex ante real interest rate on the IG can be determined
from its price alone. In this case, the expected real yield
on the nominal bond will differ from that on the indexed
bond by an inflation risk premium. Consequently, the
estimate of expected inflation will be contaminated by an
inflation risk premium that may vary systematically with
expected inflation. It is difficult to predict the sign of this
correlation on a priori grounds or to separate the con
taminated measure of expected inflation into its two com
ponents because standard models of asset pricing under
uncertainty fit real-world data very poorly. A further prob
lem arises because holders of IGs are not fully compen
sated for inflation; the expected real return on the IG will
also contain an inflation risk premium if investors are not
risk neutral; This premium is likely to be negligible
unless the IG is close to maturity, because the investor’s
exposure to inflation depends on the change in the RPI
during the last eight months of the IG’s lifetime. For a
long-dated IG, this change is likely to have only a small
impact on the average ex post real return over its life
time. Thus, it would be safe to assume that the measure
of expected inflation is much more likely to be contami
nated by a risk premium than is the expected real interest
rate.
If the two bonds used for the calculations differ in their
liquidity— say the indexed bond is less liquid— the mea
sure of expected inflation will be further contaminated by
a liquidity premium. As noted in the text, IG turnover is
small relative to the turnover of conventional gilts, and
thus IGs are presumably less liquid overall than conven
tional gilts. Nevertheless, because this general state
ment may not necessarily apply to a particular maturitymatched pair of index-linked and conventional gilts, it is
difficult to determine whether a specific IG pays a liqui
dity premium. Finally, although index-linked gilts enjoy
tax advantages compared with conventional gilts, the
procedure for decomposing nominal yields ignores tax
effects for two reasons. First, the difference in tax treat
ment may not be of much practical import because of the
dominant role that tax-free institutions play in the gilt
market. Second, Gaske found that taking tax effects into
account did not change results qualitatively.
FRBNY Q uarterly Review/Autum n 1991
in te re st rate to ch an ge s in m o n e ta ry policy. B ritish m on
e ta ry po licy was tig h te n e d very sharply in the m iddle of
1988 and rem ained re strictive until S e p te m b e r 1990.
T his tig h te n in g is reflected in the thre e -m o n th in te rb a n k
rate, w h ich rose by m ore than 700 basis points from
May 1988 to the end of 1989 and rem ained in the
n e ig h b o rh o o d o f 15 p e rc e n t u n til S e p te m b e r 1990
(C h a rt 1, low er panel). The term structure, inverted
since m id-1988, also show s the effects of m o n e ta ry
tig h tn e s s . T he sharp slo w in g of the U.K. econom y in
1990 te n d s to confirm th a t high nom inal rates did in fact
reflect m o n e ta ry tig h tn e s s ra th e r than a mere run-up of
54 FRBNY Q uarterly Review/Autumn 1991
interest rates in a n tic ip a tio n of a c c e le ra tin g in fla tio n ; it
also su g g e sts th a t real rates p ro b a b ly rose from 1988 to
1990. However, th e d e c o m p o s itio n of th e lo n g -te rm
bond yie ld in d ica tes th a t th is m o n e ta ry tig h te n in g did
not raise expected real lo n g -te rm in te re s t rates by m uch
or, more im plausibly, did not low er e xp e cte d in fla tio n at
a ll .13 In fact, the d e c o m p o s itio n la rg e ly a ttrib u te s the
13The real rate did rise som ewhat from m id-1988 to m id-1989 and
again from February 1990 to S eptem ber 1990. However, the average
real rate during the tw enty-nine months of po licy tigh tening was
actually 30 basis points lower than the average real rate over the
preceding tw enty-nine months.
s te a dy rise in lon g-te rm g o vernm ent bond yie ld s from
e a rly 1988 to m id-1990 to an increase in expected
in fla tio n. It is d iffic u lt not only to reconcile this pattern of
real and nom inal interest rate m ovem ents w ith the c o n
ve n tio n a l view that m o n e ta ry p o licy affects real e c o
nom ic a c tiv ity th rough lo n g -te rm real in te re st rates, but
also to believe that private se c to r long-term inflation
e x p e cta tio n s were not a d ju ste d dow nw ards in the face
of a ve ry resolute tig h te n in g of p o licy .14 These stylize d
facts s u g g e st that, at least fo r the 1988-90 period, the
IG m easure was a poor m easure of expected inflation
and tha t the co rre sp o nd in g real interest rate did not
a c c u ra te ly reflect ex ante real in te re st rates faced by the
private sector.
1996, w h ile the a ctual in fla tio n rate is the p e rce n ta g e
change in the RPI over the past tw elve m onths. C h a rt 2
g e n e ra lly c o n tra d ic ts a s s e rtio n s th a t the IG m easure of
expected in flation sim p ly m im ics the b e h a vio r of actual
in fla tio n .15 The IG m easure of e xp e cte d in fla tio n has
rem ained above the actual rate fo r m ost of th e sam ple
period and is also ra th e r less v o la tile than a ctual in fla
tion. It has, however, been fa irly c lo se to th e a ctual rate
from late 1989 onw ards.
A lthough the IG m easure of exp e cte d in fla tio n a p p lie s
to an interval th a t is lo n g e r th a n the s h o rt- to m edium term horizon of p rim a ry co n ce rn to p o lic y m akers, it
may n e ve rth e le ss p rovide a good fo re ca st of in flation
over a h orizon of im m e d ia te p o lic y in te re st. To assess
th is p o ssib ility, the IG m easure of e xp e cte d fu tu re in fla
tion was com pared w ith th re e sim p le m e a su re s based
on past in fla tio n: the average rates of RPI in fla tio n over
the past tw elve, tw enty-four, and th irty -s ix m o n th s .16
Each m easure of exp e cte d in fla tio n was e valuated as a
The IG m e asure o f e x p e c te d in fla tio n as a p re d ic to r
o f in fla tio n
C h a rt 2 show s the IG m easure of expected inflation
along w ith actual inflation m easured by the tw e lve
m onth perce n ta g e cha ng e s in the retail price index
(RPI). N ote that the expected inflation rate on a p a rtic u
lar date is the average annual rate from that date to late
1sSee, for exam ple, Anthony Harris, “ Lessons from the Indexed
D ecad e,” Financial Times, A pril 29, 1991.
14E xpectations of rising inflation could c o incid e with a m onetary
p o licy tigh tening if the tigh tening occurred at the same time as an
exogenous increase in dem and.
16A measure based on a sho rter period of past inflation was not used
because the RPI is not available on a seasonally ad ju sted basis.
Chart 2
Actual versus Expected Inflation
Percent
\
\
w * .
\
\
\
\
A
V
\ \
\
i\
l\
i
\ 1 1 \ •
A
,
\
\
i
r.
*
V
7 *
v\
^
» ft1
’
«
* j
K 7 " V - \
I
A
V
\
^
i
\
\
Expected inflation
i
I
V VV .//
*
\
11
y
\
/
V* f
rr
\
\
I
I
VJ
Actual inflsition
Ll l I
i i
I
i i, l i i
1982
illllllllll
Ll l u l l 1 111
1983
1984
m
I
i i
I
i i
1985
I
h
I I 1 I I I I I I LL I I l - l l l l i l I L -11 I n 1 I I I l l
1986
1987
1988
II I I I In
1989
ll 1 I l l l l l l l l l l
1990
1, 11
1991
Sources: Bank of England data and author’s calculations.
Notes: Actual inflation is measured by the percentage change in the retail price index over the preceding twelve months. Expected inflation is the
measure derived from prices of index-linked gilts.
FRBNY Q uarterly R eview/Autum n 1991
55
fo re ca st of inflation over three horizons: tw elve, tw entyfour, and th irty -s ix m onths. The results of th is exercise,
repo rted in Table 3, rely on tw o yardsticks of forecast
a ccu ra cy: the root m ean squared forecast error, w hich
m easures average p redictive accuracy over the forecast
pe riod, and the regression co e ffic ie n t of actual inflation
on the m easure of expected inflation, w hich m easures
b ia s— th a t is, the te n d e n cy to over- or u n d e rp re d ict
persisten tly. The average rate of inflation over the past
tw elve m onths is the m ost a ccurate predictor, on aver
age, of the level of in flation over all three horizons
co nsid ered (it has the sm a lle s t root mean squared fo re
cast error). A lth ough by no m eans unbiased, it has the
sm a lle st bias; the regression co e fficie n t of actual in fla
tio n on this m easure is clo s e r to unity at the one- and
tw o -ye a r h orizon s than are th o se on the o th e r m ea
sures. The p erfo rm ance of the IG m easure of expected
in fla tio n is s ig n ifica n tly b e tte r (over all three horizons)
than th a t of the average in flation rate over the past
th irty -s ix m onths and s im ila r to that of the average
in fla tio n rate over the past tw e n ty-fo u r m onths. Note,
however, th a t the IG m easure is o nly m o derately in fe rio r
to the best of the auto re g re ssive m easures (the tw e lv e
m onth m easure). For exam ple, the root m ean squared
fo re cast errors ind ica te th a t a bout tw o -th ird s of actual
inflatio n rates over a o n e -ye a r horizon w ill lie no fu rth e r
than 2.42 p erce ntag e p o in ts from the rate predicted on
the basis of the past tw elve m onths’ inflation, and no
fu rth e r than 2.79 pe rce n ta g e points from the rate pre
d icte d by the IG m easure.
It should com e as no su rp rise that the IG m easure
fares no b e tte r than recent in fla tio n in fo re ca stin g fu ture
inflation. As C h a rt 2 show s, the IG m easure is a biased
p re d icto r of in fla tio n, e xce e d in g the a ctual in fla tio n rate
over m ost of the sam ple p e rio d .17 A p la u s ib le in te rp re ta
tion of the upw ard bias in the IG m easure of in flation
e xp e cta tio n s in the e a rly p a rt of the sa m p le is th a t the
U.K. m o n e ta ry a u th o ritie s were not c re d ib le in the early
eig h ties. A fte r a long p e riod of fa irly high in fla tio n,
several years of low in fla tio n m ight have been n e c e s
sa ry to c o n vin ce m a rke t p a rtic ip a n ts th a t th e m o n e ta ry
a u th o ritie s w o u ld m a in ta in n o n in fla tio n a r y p o lic ie s .
A lternatively, the bias may sim p ly reveal th a t the IG
m easure is flawed.
One m ight ask w h e th e r the IG m easure, if p u rged of
bias, w ould p re d ict in fla tio n over a p o lic y -re le v a n t h o ri
zon more a ccu ra te ly than w ould naive m e asures based
on past inflation. U sing the m o n th ly ch a n g e in the IG
m easure, rather than its level, to fo re ca st fu tu re in flation
17The poor forecasting perform ance of the IG measure relative to
autoregressive measures is also to be exp ected from a purely
statistical view point. The RPI inflation rate is nonstationary; that is.
changes in the inflation rate tend to be perm anent and
consequently the inflation rate does not tend to return to its longrun average after a change. U nder these con ditions, inflation over
the recent past will generally be the best sim ple p re d icto r of future
inflation. Note, however, that using past inflation as a p re d icto r of
(the level of) inflation over longer horizons results in large
expectations errors. That is, the variance of the forecast error,
conditional on inform ation available at the tim e the forecast is
form ed, is proportional to the length of the forecast horizon.
Similarly, if the IG measure were an accu rate p re d icto r of future
inflation, it would tend to move very closely with current inflation.
Thus, from a purely statistica l view point, it m ight be considered
surprising that the IG measure does not respond one-to-one to
changes in actual inflation.
Table 3
Comparison of Indexed Gilt Measure and Naive Autoregressive Measures of Expected Inflation
Root Mean Squared Forecast Errors
(Percentage per Year)
Forecast Horizon
Measure of E xpected Inflation
IG measure
Inflation over past twelve months
Inflation over past tw enty-four months
Inflation over past thirty-six months
One Year
Two Years
Three Years
2.79
2.42
2.81
3.58
2.73
2.41
2.77
3.57
2.78
2.52
2.77
3.57
Regression C oefficient of Actual on Expected Inflation
Forecast Horizon
Measure of E xpected Inflation
IG measure
Inflation over past twelve months
Inflation over past tw enty-four months
Inflation over past thirty-six months
One Year
- 0 .0 7
0.26
-0 .0 2
- 0 .1 8
Two Years
- 0 .3 7
0.00
- 0 .2 4
- 0 .2 0
Three Years
- 0 .4 4
-0 .3 2
- 0 .2 8
- 0 .1 6
Note: Sam ple pe riods are as follows: M arch 1982 to July 1990 for one-year forecasts, M arch 1982 to July 1989 for tw o-year forecasts, and
M arch 1982 to July 1988 for three-year forecasts.
http://fraser.stlouisfed.org/
56 FRBNY Q uarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
w ill e lim in a te bias th a t rem ains co n sta n t over tim e. But
if the bias is in fact due to c re d ib ility problem s, it has
pro bab ly d ecreased over tim e, and co n se q u e n tly som e
bias is like ly to rem ain in the data. N e vertheless, unless
one know s the process w hereby m arket p a rtic ip a n ts
change th e ir view s on the c re d ib ility of the m onetary
a u th o ritie s , any m e th o d of e lim in a tin g bias w ill be
im perfect. The resu lts in Table 4 suggest th a t w hen
purged of bias in th is manner, the IG m easure does
m a rg in a lly b e tte r than sim p le autoregressive m easures
in p re dicting futu re a cce le ra tio n and d e ce lera tio n of
inflation. The ta ble co m pares the perfo rm a n ce of the
m o nth ly ch ange in the IG m easure in forecasting the
change in the RPI infla tion rate over the fo llo w in g tw elve
m onths w ith th a t of fo u r naive autoregressive m easures:
the ch a nge s in the RPI in fla tio n rate over the preceding
on e -, th re e -, s ix-, and tw e lv e -m o n th p e rio d s .18 The
m onthly cha nge in the IG m easure is not only the m ost
a ccurate predictor, on average, of the RPI inflation rate
(it has the sm a lle st root m ean squared forecast error),
but it is also a p p re cia b ly clo s e r than the naive m ea
sures to offe ring an unbiased fo re ca st of changes in the
RPI in flation rate (the regression co e ffic ie n t of actual on
p redicted ch ange s is the c lo s e s t to u n ity ).19 It should be
18The tw elve-m onth forecast horizon was chosen somewhat arbitrarily
for illustrative purposes. A lthough clearly of interest to policy
makers, this horizon is not necessarily the most relevant.
19C om paring the a b ility of alternative measures of inflation
expectations to predict changes in inflation is also advisable on
purely statistica l grounds. Because the inflation rate is
nonstationary, the change in the inflation rate contains all the new
inform ation pe rtainin g to the future course of inflation.
Table 4
Forecast Performance for Changes in
Retail Price Index Inflation: Comparison of
Indexed Gilt and Naive Measures
Criterion of Forecast Accuracy
Measure
Root Mean
Squared Error
(Percentage
per Year)
Regression Coefficient
of Actual on
Predicted Inflation
2.34
0.69
2.37
0.32
2.57
0.03
2.95
-0 .0 1
3.53
0.03
One-month change
in IG Measure
One-month change
in RPI inflation
Three-month change
in RPI inflation
Six-month change
in RPI inflation
Twelve-month change
in RPI inflation
Notes: Changes in RPI inflation are measured as twelve-month
changes in the twelve-month percentage change in the RPI.
Sample period is April 1982 to July 1990.
em phasized, however, th a t the d iffe re n ce in the fo re
casting p e rfo rm a n ce of th e IG m easure and th e ch ange
in the RPI in flation rate over th e p re ce d in g m onth is so
sm all as to be of no p ra c tic a l s ig n ific a n c e . O ver lo n g er
data sam ples or d iffe re n t tim e p e rio d s, th e ra n kin g of
the tw o m easures could e a sily be reversed.
The preceding c o m p a ris o n s of the IG m easure and
naive a u to re g re s s iv e m e a s u re s of e x p e c te d in fla tio n
were de sig ne d to show w h e th e r th e IG m easure is a
be tte r p re d icto r of in fla tio n than are s im p le a lte rn a tive s.
The p redictive value of the IG m easure of expected
inflation can also be asse sse d by d e te rm in in g w h e th e r
it p rovides in fo rm a tio n th a t im proves th e fo re ca stin g
a b ility of an a u to re g re ssive m odel based on past in fla
tion. The te st results re p o rte d in Table 5 p ro vid e e v i
dence th a t the a d d itio n a l in fo rm a tio n c o n trib u te d by the
IG m easure, w h ile s ta tis tic a lly s ig n ific a n t, is too m ar
ginal to be of p ra ctica l use. The te s t e va lu ate s w h e th e r
th e IG m e a s u re o f e x p e c te d in fla tio n ca n p re d ic t
changes in the RPI in fla tio n rate, m easured as the
m onthly p e rce n ta g e ch a n g e in the RPI, once th irte e n
la g g ed c h a n g e s in th e in fla tio n rate (a n d s e a s o n a l
dum m ies) have been ta ke n into a cco u n t in fo rm in g the
pre d ictio n s. The firs t fo u r lagged va lu e s of m o n th ly
changes in the IG m easure of e xp e cte d in fla tio n are
jo in tly s ta tis tic a lly s ig n ific a n t at the 2.5 p e rc e n t level,
and e ig h t lagged values are jo in tly s ig n ific a n t at the 10
p e rce n t level. However, the a d d itio n of e ig h t lagged
changes in the IG m easure o nly ra ise s th e a d ju ste d R2
of the fore ca stin g e q u a tio n from 0.8 to 0.82, to o sm all
an im provem ent to be of p ra ctica l im p o rt. T h e se results
are not se n sitive to the n u m b e r of lagged c h a n g e s in
the inflation rate in the re g re ssio n , a lth o u g h the nu m b er
included (th irte e n ) is so m e w h a t a rb itra ry. In a d d ition ,
th e lim ite d p r e d ic tiv e v a lu e o f th e IG m e a s u re
Table 5
Contribution of Indexed Gilt Measure in
Predicting Monthly Changes in Inflation
Lags of IG Measure
of Expected Inflation
None
One to four
One to eight
One to thirteen
R2
M arginal S ignificance*
(F-Test)
0.80
0.82
0.82
0.81
0.024
0.094
0.146
—
Notes: In the baseline regression, the m onthly cha nge in the
RPI is regressed on thirteen own lags and a set of seasonal
dum m ies. Sample period is May 1983 to A pril 1991.
TMeasures the highest sig nifica nce level at w hich one can
reject the null hypothesis that the num ber of lagg ed changes
in the IG measure of exp e cte d inflation does not co n trib u te to
forecasting the change in RPI inflation over the next m onth.
The confidence level in the null hypothesis is given by 1m arginal significance.
FRBNY Q uarterly R eview/Autum n 1991
57
may sim p ly reflect the degree of p re d ic ta b ility of the RPI
th a t derive s from purely m e ch a n ica l aspects of its c a l
cu la tio n . For exam ple, ch a n g e s in the banks’ base in te r
est rates have a fore se e a b le e ffe ct on m ortg a g e interest
rates b ecause variable rate m ortg a g e s are the rule in
th e U n ite d K in g d o m ; c o n s e q u e n tly , th e s e c h a n g e s
a ffe ct the RPI p re d icta b ly th ro u g h the e ffe ct of m o rt
gage paym ents on the co st of housing.
E x p e c te d in fla tio n a n d e x p e c te d real in te re s t rates as
p re d ic to rs o f re a l e c o n o m ic a c tiv ity
T he IG m a rket may convey in form ation useful in the
fo rm u la tio n of m o n e ta ry p o licy even though the IG m e a
sure of expected infla tio n is no more su cce ssfu l in
fo re ca stin g in flation than are sim ple m easures based on
past in flation. The IG m easure may be an im p e rfe ct
p re d icto r of future inflation sim p ly because it fa ith fu lly
reflects the private s e c to r’s unrealized e xp e cta tio n s of
in fla tio n. In this case, the IG m arket may provide policy
m akers w ith a reliable m easure of ex ante real interest
rates. If private se cto r sp e n d in g is in te re st-se n sitive ,
the m arket may also yield in form ation about private
se cto r spe nding plans and near-term e conom ic d e ve l
opm e nts. In sum , the IG m easure of expected real lo n g
term in te re st rates may be a p o te n tia lly valuable in d ic a
to r of m acroe co nom ic d evelopm ents.
O ne way to te st the accu ra cy of the IG m easure in
gaug ing private se cto r in fla tio n e xp e cta tio n s is to d e te r
m ine w h e th e r the IG m easure of ex a nte real lo n g -te rm
in te re st rates p rovides in fo rm a tio n a b o u t fu tu re real
e c o n o m ic a ctiv ity . C o m p a re d w ith m ore d ire c t te s ts
such as co rre la tin g the IG m easure w ith su rv e y m e a
sures of in flation e x p e cta tio n s, th is te s t has a d is a d v a n
tag e : it can only p rovide in fo rm a tio n a b o u t th e a c cu ra cy
of the IG m easure as a ya rd s tic k of private s e c to r
in flation e xp e cta tio n s to the exte n t th a t private se c to r
d e cisio n s are in te re s t-s e n s itiv e . T h is s h o rtc o m in g is
also an im p o rta n t a d vantage, however, b e ca u se th e te st
m easures the a ccu ra cy of the IG m easure in te rm s of a
goal variable of d ire ct in te re st to p o licy m a ke rs. F u rth e r
more, the te st ca p tu re s any p o te n tia l le a d in g in d ic a to r
role of indexed bond p rice s and as such is of in te re st to
p olicy m akers.
To d e te rm in e w h e th e r th e indexed g ilt m a rke t p ro
vides in fo rm a tio n a b o u t fu tu re real e co n o m ic a ctivity,
real GNP grow th is regressed on la g g ed G N P grow th
and various nom inal in te re st rates and in fla tio n rate
m easures, in clu d in g th o se o b ta in e d from in d e x -lin ke d
g ilts. The results from th e se re g re ssio n s are p ro b a b ly
best a ppreciated in the co n te x t of the s ty liz e d facts
e sta b lishe d by s im ila r re g re ssio n s on U.S. data. In
particular, a lth o u g h U.S. real GNP grow th is ty p ic a lly
very d iffic u lt to p redict, E strella and H a rd o u ve lis and
S tock and W atson have found th a t s h o rt-te rm nom inal
in te re st rates and m easures of the slo p e of the term
stru ctu re convey in fo rm a tio n a b o u t fu tu re m o ve m e nts in
Table 6
Predictive Value of Nominal interest Rates for Real GNP Growth in the United States
and the United Kingdom
United States
Four Lags of
S tatistic
F
LR
Real GNP
Growth
Ten-Year Governm ent
Bond Yield
Six-M onth
C om m ercial Paper Rate
0.99
4.27
0.27
3.61***
15.04***
6.48***
26.00***
United Kingdom
Four Lags of
S tatistic
F
LR
R*
Real GNP
Growth
3.45**
14.38***
0.12
Ten-Year Governm ent
Bond Yield
1.20
5.35
Three-M onth
Interbank Rate
2.44*
10.65**
Notes: Sam ple period for the United States is 1954-11 to 1991-1: for the United K ingdom , 1965-1 to 1991-11. F and LR are the F sta tistic and
likelihood ratio s ta tistic for testing the null hypothesis that a pa rticu la r variable has no explanatory power for future real GNP grow th in a
regression inc lu d in g the listed variables as regressors. One asterisk denotes significa nce at the 10 pe rcent level; two, sig n ifica n ce at the 5
pe rcent level; and three, sign ific a n c e af the 1 percent level.
Interest rates are measured as qu a rte rly averages of m onth-end observations. GNP growth is measured as q u a rte rly pe rce n ta g e cha nges
seasonally ad ju sted at an annual rate.
58 FRBNY Q uarterly Review/Autumn 1991
real o u tp u t .20 Table 6 provides ben ch m a rk regression
results illu s tra tin g the p red ictive value of nom inal sh o rtand lo ng-term in te re st rates fo r U.S. real GNP grow th. It
also show s that nom inal in te re st rates are less in fo r
m ative in the U nited K ingdom than in the U nited S tates.
R esults fo r the U nited K ingdom spa n n in g the period
s ince the in ce p tion of the in d e x-lin ke d m arket are pre
se n ted in Table 7. The ta b le illu stra te s how adding
d iffe re n t va ria b le s in an eq u a tio n to forecast real GNP
g row th a ffects the a d justed R2 of the equation. Note th a t
the co m p o n e n ts of the real in te re st rate are added
separately, in p a rt because the hyp o th e sis that the
nom inal rate and the inflation rate have co e fficie n ts
equal but o p p o site in sign is g e n e ra lly rejected by the
data. The regression results s u p p o rt three sp e cific co n
clu sio n s: First, U.K. real G NP grow th, like its U.S.
c o u n te rp a rt, is hard to predict, but in c o n tra st to the
U.S. e xpe rien ce , long- and s h o rt-te rm nom inal interest
rates (and by im p lica tio n m easures of the slope of the
term structure) do not fore ca st future m ovem ents in real
GNP (co lu m ns 1 and 2). S econd, the d e co m p o sitio n of
lo n g -term nom inal rates based on indexed g ilt prices
provides no s ig n ifica n t in fo rm a tio n about future real
GNP grow th in the U nited K ingdom . (The ad ju ste d R2 of
the fore ca stin g eq ua tio n a ctu a lly falls w hen the IG m ea
sure of expected inflation is added to the forecasting
e q u a tio n .) Finally, backw a rd -lo o kin g m easures of sh o rta n d lo n g -te r m re a l in te r e s t ra te s , o fte n u se d as
20See A rturo Estrella and Gikas Hardouvelis, “ Possible Roles of the
Yield Curve in M onetary Policy,” Interm ediate Targets and
Indicators for M onetary Policy, Federal Reserve Bank of New York,
New York, 1990; and James H. Stock and Mark W. Watson, “ The
Business C ycle Properties of Selected U.S. Economic Time Series,
1959-1988,” National Bureau of E conom ic Research, Working Paper
no. 3376.
Table 7
Predictive Value for Future Real GNP Growth:
Comparison of Indexed Gilt and Other
Variables
Regression Number
Four lags of
Real GNP growth
Long-term government
bond yield
Three-month
interbank rate
IG measure of
expected inflation
Twelve-month change
in RPI
R2
Standard error
(1)
X
(2)
X
(3)
X
(4)
X
X
X
X
x
X
X
X
0.16
2.69
0.17
2.67
Note: Sample period is 1983-11 to 1990-IV.
0.14
2.72
X
0.55
1.97
in d ica tors of the s ta n ce of m o n e ta ry policy, do yield
s ig n ific a n t in fo rm a tio n a b o u t fu tu re real e co n o m ic a c tiv
ity. Four lags of the th re e -m o n th in te rb a n k rate to g e th e r
w ith fo u r lags of the RPI in fla tio n rate (the p e rce n ta g e
change in the RPI over the p re ce d in g tw elve m onths)
raise the ad ju ste d R2 of a re g re ssio n of real GNP grow th
on four lagged values of real G NP grow th from 0.16 to
0.55 (colum n 4).
The results re p o rte d here show th a t the p rice s of
in d e x-lin ke d g ilts do not convey p o lic y -re le v a n t in fo rm a
tio n a b o u t fu tu re tre n d s in e c o n o m ic a c tiv ity . If we
a ccept th a t private s e c to r d e c is io n s are s e n s itiv e to real
in te re st rate m ovem ents, the re su lts im p ly th a t the IG
m easure of ex ante real lo n g -te rm in te re s t rates does
not a c cu ra te ly reveal real in te re s t ra te s fa ce d by the
private sector. T he lim ita tio n s of the IG m easure of
expected in flation as a gauge of private s e c to r in fla tio n
e xp e cta tio n s co u ld be due to any of th re e ca u se s. F irst,
lim ite d p a rtic ip a tio n in the U.K. indexed g ilt m a rke t may
have m ade the IG real in te re s t rate relevant to o n ly a
v e ry sm a ll p a rt of th e p riv a te se cto r. S e c o n d , th e
expected rate of in fla tio n and the c o rre s p o n d in g real
in te re st rate in the bond m a rke t m ay not be relevant to
the m a jo rity of p a rtic ip a n ts in th e g o o d s and fa c to r
m arkets. Finally, the p oor p e rfo rm a n c e of the IG m e a
sure of expected in fla tio n m ay d e rive from ta x d is to r
tio n s or the fa ct th a t m a rke t p a rtic ip a n ts are risk averse.
Two caveats to th e se c o n c lu s io n s d e s e rv e m en tio n,
however. First, the p re d ictive pow er of IG p rice s w ill
depend on the m o n e ta ry p o lic y rule fo llo w e d by the
au th o ritie s. If the B ank of E ngland were s ta b iliz in g real
interest rates, one w ould not e xp e ct the IG m easure of
th e real in te re s t ra te to have p re d ic te d re a l G N P
c h a n g e s . S e c o n d , th e c o n c lu s io n s a re te n ta tiv e ,
because the U.K e xp e rie n ce w ith indexed b onds is
co m p a ra tively sh o rt. The a d d itio n of o n ly a few ye a rs’
data may ve ry w ell lead to c o n c lu s io n s m ore favor
able to the p o sitio n held by p ro p o n e n ts of indexed
bonds.
Conclusion
T his a rtic le has used data from th e U.K. m a rke t for
in d e x-lin ke d g ilts to assess the a lle g e d p o lic y b e n e fits
of indexed bonds. It has been su g g e ste d th a t a re a l
tim e m arket m easure of exp e cte d in fla tio n (and the
co rre sp o nd in g ex ante real in te re st rate) d e rive d from
indexed bond p rice s co u ld p rovide the F ederal R eserve
S ystem w ith va lu a ble in fo rm a tio n a bout m a rke t p e rc e p
tio n s of, and reaction to, its p o licie s, and convey in fo r
m ation about fu tu re in fla tio n and real e c o n o m ic d e v e l
o p m e n ts . T h e e v id e n c e p re s e n te d in th is a rtic le ,
however, su g g e sts th a t th e p rice s of in d e x -lin k e d g ilts
may not convey m uch in fo rm a tio n a b o u t fu tu re in fla tio n
and real e co n o m ic activity. For th is reason, a u th o ritie s
FRBNY Q uarterly R eview/Autum n 1991
59
may question whether a real-time market measure of
expected inflation can shed light on private sector reac
tions to monetary policy.
The ability of the IG measure of expected inflation to
anticipate future inflation developments appears to be,
at best, mixed. It is a biased predictor of future inflation,
fares no better than simple inflation expectations mea
sures based on past inflation, and does not add appre
ciably to the predictive power of a more sophisticated
backward-looking model of inflation expectations.
U.K. indexed bond prices also do not seem to convey
policy-relevant information about future real economic
60 FRBNY Quarterly Review/Autumn 1991
activity. This finding is consistent with the IG measure’s
being an imperfect gauge of private sector inflation
expectations and with the failure of the corresponding
real interest rate to reflect accurately ex ante real inter
est rates faced by the private sector. The behavior of
the IG measures of expected inflation and the real long
term interest rate during the period of restrictive mone
tary policy from mid-1988 to late 1990 further supports
this judgment. In sum, these results suggest that the
U.K. IG measure of inflation expectations seems to offer
only limited, if any, information for the conduct of mone
tary policy.
Tracking the Economy with the
Purchasing Managers’ Index
by Ethan S. Harris
In the last several years the purchasing managers’
index has emerged as a key indicator of manufacturing
activity. This “index” consists of five separate indexes
measuring monthly changes in manufacturing output,
employment, new orders, inventories, and vendor deliv
eries, together with a composite index that gives a
weighted average of the other five. Financial markets
are now quite sensitive to the index, and news reports
on the economy regularly feature it. The index receives
such close attention for several reasons: it is the first
broad indicator released each month, it covers the
cyclically sensitive manufacturing sector (Chart 1), and
the data are easy to interpret and are virtually never
revised.
Despite the index’s popular appeal and market-moving power, some skepticism about the utility of this
indicator is warranted. It is not constructed with the
scientific sampling and statistical methods that underlie
most official macroeconomic series (see appendix). A
qualitative measure of activity, it reports whether busi
ness has increased or decreased but makes no assess
ment of the strength of the change. Most important, the
index has not been rigorously tested: although there is
ample evidence that the index tracks the general ups
and downs of the economy, analysts have not demon
strated that the purchasing managers’ data yield infor
mation on the economy beyond that already provided by
other indicators.
This article analyzes the strengths and weaknesses
of the index as a forecasting tool. It begins by explain
ing how the index is constructed. The next section
presents the basic correlations between the five compo
nent indexes and the economic aggregates they are
supposed to track. The remainder of the article investi
gates the predictive power of the purchasing managers’
data: Do the indexes lag or lead economic activity? Do
they foreshadow turning points in the business cycle?
Can the indexes improve on the forecasts of simple
economic models or on consensus forecasts?
Our results give mixed support for the purchasing
managers’ index. One shortcoming is the index’s ten
dency to pick up activity in the weeks preceding the
month it is supposed to measure. Another limitation is
that none of the components explains more than half of
the monthly variation in the corresponding official statis
tics. Furthermore, the index is not a reliable leading
indicator: it sends too many false signals and its lead
time is too erratic to be of use in anticipating cyclical
swings. Nevertheless, the index does add significantly
to the explanatory power of simple econometric models
and consensus forecasts. And it could be even more
useful to forecasters if the sampling and statistical
methodology were improved. Thus, although the index
has some important limitations, with careful application
it can be useful in forecasting economic activity.
Description
About the middle of each month the National Associa
tion of Purchasing Managers (NAPM) surveys roughly
300 association members representing twenty-one
manufacturing industries in all fifty states. The survey
asks each purchasing manager how the current level of
five key economic indicators— production, new orders,
employment, inventories, and vendor delivery time—
FRBNY Quarterly Review/Autumn 1991
61
c o m p a re s w ith th e p re v io u s m o n th ’s le v e l .1 T h e
respon se s are sim p ly “ higher,” “ lower,” or “ the sa m e .”
T he unw eighted p erce n ta g e of firm s in each ca te g o ry is
then ta b u la te d and a d iffu s io n index is c o n stru cte d by
su m m in g the p e rc e n ta g e of p o sitive re sp o n se s and
o n e -h a lf of those responding “ the s a m e .”2 A reading
above 50 percent in a d iffu s io n index m eans that more
firm s are expan ding a c tiv ity than co n tra ctin g activity.
Finally, th e se data are se a so n a lly adjusted and co m
bined into a sin gle w e ig h ted com p o site index.
A lth o u g h the surve y has been published since 1931
(w ith an in te rru p tio n fo r W orld War II), several of its
m ore so p h istica te d features were only introduced in
recent years. The data were o rig in a lly p u blished in raw,
se a so n a lly u na djuste d form ; in the early 1980s, w ith
help from the C om m erce D epartm ent, the asso cia tion
began p ub lish in g se a so n a lly adjusted d iffu sio n indexes.
T he sam p le size has also been increased to a lm ost 300
1The survey also includes questions on com m odity prices and
buying policy. In the last several years new export orders and
im ports have been added.
2The NAPM survey treats vendor de live ry time somewhat differently.
The responses for this in dica tor are “ slower,” “ faster,” and “ no
cha n g e ." The diffusion index for vendor deliveries is the sum of the
percentage reporting slower delivery time and half the percentage
reporting no change.
62 FRBNY Q uarterly Review/Autumn 1991
from about 225. S ince th e su m m e r of 1989, w hen fin a n
cia l m a rke ts b e ca m e in c re a s in g ly in te re s te d in th e
index, the survey has been released e a rlie r and at the
sam e tim e each m o n th . It now u s u a lly “ b e a ts ” th e
em p lo ym e n t re p o rt by several days and th u s c a p tu re s
m axim um a tte n tio n in the m a rk e t .3
The index as a measure of economic activity
The NAPM co m p o n e n t indexes have c o u n te rp a rts in
official data p u b lish e d by the fe d e ra l g o ve rn m en t. S ince
the indexes are m easures of the d iffu s io n of the e c o
nom ic activity, they sh ould have ro u g h ly a lin e a r re la
tio n s h ip to the g ro w th in co rre s p o n d in g g o ve rn m e n t
d a ta .4 In oth e r w ords, if a h ig h e r p ro p o rtio n of firm s are
re p o rtin g e xp a n d e d a c tiv ity , th e n we w o u ld e xp e ct
hig h er grow th in a g g re g a te activity.
Table 1 presents evid e nce of how c lo s e ly the NAPM
data track the econom y. The ta b le show s the re sults of
regressing the p e rce n t ch a n g e in the o fficia l d a ta on the
co rre sp o nd in g c o m p o n e n t of the index. T he firs t three
colum ns present e stim a te s using m o n th ly d a ta fo r the
1959-91 p e rio d . As th e t-s ta tis tic s (in p a re n th e s e s )
show, the NAPM c o e ffic ie n t is s ig n ific a n t in all of the
equations. The overall fit (R -square), however, is g e n e r
ally m odest; the indexes e xplain less than h a lf of the
m onthly variation in grow th fo r all v a ria b le s . T he w e a k
est results are for new o rd e rs and p rice s, tw o h ig h ly
vo la tile se rie s; the best re su lts are fo r e m p lo ym e n t.
The fo u rth and fifth co lu m n s of Table 1 show the
overall fit w hen q u a rte rly and ann u a l d a ta are used.
A lthough this tim e a g g re g a tio n g e n e ra lly im proves the
fit, the NAPM data still leave a good p o rtio n of the
variation in grow th u n e xp la in e d . T he last tw o c o lu m n s of
Table 1 show th e im p lie d b re a k -e v e n p o in t fo r the
indexes. T h e o re tica lly, w hen a g g re g a te grow th is zero,
a d iffu sio n index sh ould average out to a b o u t 50 per
cent, w ith equal nu m b e rs of firm s re p o rtin g h ig h e r and
lower activity. The regression e stim a te s s u g g e st th at
using 50 p e rce n t as the break-even p o in t can be m is
leading. For exam ple, in the regression of in d u stria l
production on the c o m p o s ite index, e s tim a te s for the
1980-91 period show a break-even rate of ju s t 46.9
p e rc e n t .5
in te re s t in the index has also drawn attention to som e of the
regional purchasing m anagers’ surveys. The C hicag o index is
closely watched, in part because it is released before the national
index.
♦Specifically, if (1) firm s have id en tical but no nsynchronous cycles
and (2) growth is evenly d istrib u te d am ong large and sm all firms
along a rectangular d istrib ution , then there will be an exact linear
relationship between the proportion of firm s exp andin g and the rate
of growth of aggregate activity. Regression tests found no evidence
of significant nonlinearities.
5Recent experience illustrates the d a nger of using 50 pe rcent as the
break-even point. From May 1989 to A pril 1990 the com posite
NAPM index dip p e d below 50 percent, averaging 47.6 percent, If
The index as a leading indicator
pe rce n t as the e co n o m y slip s in to re ce ssio n . E m p irical
w ork by C ox and Torda show s th a t the c o m p o s ite index
"reached its c y c lic a l p eak a b o u t 1 1 1/2 m onths before the
on se t of the seven po stw a r re ce ssio n s” and th a t “ the
lead tim e of the c o m p o s ite index of le a d in g e co n o m ic
in d ica tors is sim ilar, a b o u t 12 m o n th s .”6 C ox and Torda
also fin d th a t th e c o m p o s ite in d e x g e n e ra lly le a ds
cyclica l recoveries.
U nfortunately, average lead tim e is a p o o r c rite rio n for
ju d g in g a leading indicator. To be u se fu l, a le a d in g
in d ica to r m ust p re d ict tu rn in g p o in ts w ith a re la tive ly
regular lead of at least a few m onths. It m u st a lso give a
re la tive ly sm all n u m b e r of fa lse s ig n a ls . Here we te st
the predictive pow er of tw o ty p e s of m o ve m e n t in the
c o m p o site index: tu rn in g p o in ts in the index and p e rio d s
w h e n th e in d e x c ro s s e s v a rio u s “ b r e a k - e v e n ” o r
“ th re s h o ld ” points.
N e ith e r NAPM sig n a l re lia b ly p re d icts b u s in e s s cycle
tu rn in g points. As C h a rt 2 show s, the index ofte n tu rn s
dow n long before a b u sin ess cycle peak, re fle ctin g the
slow ing of grow th fo llo w in g the in itia l c y c lic a l recovery.
Even if we ignore th is in itia l peak, the index has m u ltip le
peaks in the co u rse of each e x p a n sio n , and th e peak
Tracing the gen eral m ovem ents in e conom ic a c tiv ity is
not a ve ry rig o ro u s te st of an indicator. M uch of the
in te re st in the purchasing m an a g e rs’ index am ong b u s i
ness e co n o m ists stem s from its a lleged a b ility to signal
c h a nges in eco n o m ic trends. The tre m e n d o u s attention
the index now receives sta rte d in the sum m er of 1989
w hen the index, fa llin g below 50 percent, appeared to
presage a recession. C le a rly the in d e x’s e arly release
m akes it a “ tim e ly in d ic a to r” ; the more d iffic u lt question
is w h e th e r it in fact a n tic ip a te s a c tivity in the m onths
ahead. D oes it lead a ctiv ity or m easure c o n te m p o
ra n eou s a c tiv ity ? And are b u sin ess eco n o m ists correct
in a s s u m in g th a t it g iv e s a re lia b le w a rn in g of
re cession?
The purch asin g m anag e rs’ index, like all d iffu sio n
indexes, has leading in d ic a to r q u a litie s. C h a rt 2 show s
the re la tio n sh ip betw een the co m p o site index and the
grow th in m anu fa ctu rin g o u tp u t over the bu sin ess cycle.
The index peaks w hen grow th is highest, d e clin e s to 50
pe rcent as grow th levels off, and then fa lls below 50
Footnote 5 (continued)
50 percent is the break-even point, this drop in the index implies
about a 2 pe rcent de clin e in m anufacturing output. In fact, as the
regression estim ates predict, output showed no change over this
period.
6W illiam A. Cox and Theodore S. Torda, "Survey By Purchasing
M anagers Can Provide Signal On End Of R ecession,” Business
A m erica , July 14, 1980, p. 21.
Table 1
“ B reak-even” Regressions for Manufacturing Growth
Predictive Power (R2)
Break-even Point
Constant
Slope
Monthly
Q uarterly
Annual
1959-91
1980-91
- 3 .6 2
(11.7)
- 2 .4 4
(17 9)
0.070
(12,9)
.300
.666
.582
51.5
49.3
0.050
(18.4)
465
.772
.681
49.1
47.4
New orders*
- 3 .2 2
(4 5 )
.074
.558
.588
51.4
49.3
M aterials inventories
- 2 .2 0
(10.2)
0.063
(4.8)
0.049
(1 1 2 )
.246
.512
.239
44.7
47.0
C apacity utilization*
69.24
(89.2)
.426
.439
.435
—
—
Crude produce r prices
- 1 .8 9
(3.4)
0.236
(16.9)
0.034
(4.1)
.041
.169
.506
56.3
56.1
Industrial production
- 3 .3 0
(10.3)
0.067
(11.4)
.250
.556
.669
49.0
46.9
Real GNP
- 3 .6 0
(7.4)
0.081
(9 1 )
.393
.713
44.4
44.5
Series E xplained
Industrial production
Payroll em ploym ent
With the C om posite Index
Notes: Regression c oe fficients are based on the January 1959-M ay 1991 sam ple. Except in the ca p a city utilizatio n equation, the d e pend ent
variable enters as a sim ple percentage change. A bsolute t-values are in parentheses.
f Sample starts in 1967 and the de p e n d e n t variable is deflated using the im plicit deflator for shipm ents.
*The inde pen den t variable is vendor deliveries, lagg ed three months.
FRBNY Q uarterly Review/Autum n 1991
63
th a t fin a lly “ c o rre c tly ” sig n a ls recession can o ccu r a n y
w here from zero to tw e n ty m onths before the onset of
re ce ssion s. The index is ju s t as e rratic in predicting
cyclica l trou ghs, botto m in g out anyw here from zero to
tw elve m onths before the e co n o m y -w id e tro u g h .
If the 50 perce n t th re sh o ld is used ra th e r than the
in d e x’s peak, e q u a lly vexing p ro b le m s e m e rg e (Table 2).
The index u su a lly drops below 50 before c y c lic a l peaks,
Chart 2
The Purchasing Managers’ Index and the Growth in Manufacturing Output
Percent
80 -------
Percent
30
r
NAPM composite index
^
r
!
Manufacturing output 11
Scale---►
20
ii i l i i i l i
1959 60
m
61
1 1 1 1 ! 111 li i ii 111 m l i i l l
62
63
64
65
66
67
11
iliiliiiiiiilllllilit
68
69
70
71
72
73
74
ll i 11h ; iII ll 11111111111 i I ■H , 11
-LL
75
76
77
78
79
80
81
82
83
84
85
i l l 1.11 .1,
86
87
88
-30
89
90 91
Notes: Chart shows quarterly averages of monthly data. Manufacturing output is from the industrial production index. Shaded areas denote recessions.
Table 2
Does the Composite Index Signal Business Cycle Turning Points?
Lead ( + ) or Lag ( - ) Time in Months
NAPM Threshold
Peak
50.0
49.0
NAPM Threshold
44.5
O ctober 1949
50.0
N ovem ber 1948
+8
+8
July 1953
+2
+2
-1
May 1954
A ugust 1957
+5
+5
-2
A pril 1958
-2
A pril 1960
+ 1
+ 1
- 1
February 1961
D ecem ber 1969
-1
- 1
-9
N ovem ber 1973
-1 0
-1 0
-1 1
+5
+2
July 1981
0
July 1990
+ 14
January 1980
Average
False alarm s
0
Through
49.0
44.5
+ 1
+ 1
+2
0
0
+2
-1
-2
-2
-1
November 1970
-3
-3
-1
March 1975
-5
-5
-3
-2
July 1980
-2
-2
-1
0
-2
November 1982
-3
-3
-2
0
-3
+ 2.7
+ 0.8
-3 .4
Average
-2 .0
-1 .9
-0 .5
4
4
1
0
0
1
False alarms
-1
Notes: In keeping w ith the leading indica tor literature, the com posite index is assum ed to signal a turning point when it crosses the
threshold value for three or more consecutive months. The signal is dated from the first m onth the threshold is crossed. S ignals reversed for
at least three m onths before a c y c lic a l turning point occu rs are considered “ false alarm s."
64 FRBNY Q uarterly Review/Autumn 1991
but the lead time is quite variable. In the 1973-75
recession, the index did not signal recession until
almost a year after the onset of the downturn; by con
trast, in the most recent recession, the index stumbled
along at just below 50 percent for more than a year
before the economy turned down. Even worse, it falsely
predicted four business cycle peaks, with several sig
nals lasting as much as a year. Its record for cyclical
troughs is equally dismal: the index usually surpassed
the 50 percent break-even point two to five months after
the economy had moved out of recession.
Similar problems arise when threshold values below
50 percent are used. The findings in Table 1 suggest a
49 percent break-even value for industrial production
and a 44.5 value for real GNP. Using 49 percent rather
than 50 percent as the break-even value has virtually no
impact on the timing of the signal. Using 44.5 percent
changes the results, but not for the better. At cyclical
peaks, the index falls below 44.5 percent after the
turning point in all but one downturn, with an average
lag of three months. At cyclical troughs, the signal is a
little more timely, but again it usually fails to anticipate
the recovery. The only advantage of the 44.5 percent
threshold is that it produces only two false signals in the
postwar period.
Thus the composite index has two problems as a
leading indicator. First, because business cycles do not
follow a smooth growth pattern, the index often peaks
during the initial recovery and then reaches several
mini-peaks in the course of an expansion. Second,
because growth usually does not flatten out gradually at
the peak of the business cycle, the composite index
may not dip below 50 percent until after the recession
starts. Furthermore, as noted in the appendix, the
NAPM data may lag economic activity by about half a
month because respondents have incomplete data on
the current month when they fill out the survey. There
fore, as a cyclical indicator, the index is better used to
confirm recent turning points than to anticipate them.
Three horse races
Clearly the composite index and its components have
important limitations as stand-alone indicators of the
strength of the economy. This section tests how the
indexes perform in comparison with alternative forecast
ing tools. In particular, the analysis explores how the
indexes stack up against economic models and consen
sus forecasts in explaining the growth in nonfarm
payroll employment, industrial production, and real
GNP. The results confirm that the indexes are poor
stand-alone predictors, but they also demonstrate that
the indexes provide helpful incremental information to
forecasters. In other words, the indexes represent an
imperfect but useful addition to our knowledge of cur
rent economic conditions.
Forecasting nonfarm em ploym ent growth
In the last several years the employment report has
become the most important economic indicator for datawatchers.7 Recognizing this, the purchasing managers’
survey committee has pushed up the release date for
the index so that it now usually precedes the employ
ment report. Not surprisingly, the composite index and
its employment component are viewed as vital informa
tion in the payroll employment guessing game.
The explanatory power of the index is tested against
two standards. First, the predictions of a simple eco
nomic model of employment growth are compared with
those of the NAPM data. Second, the performance of
the NAPM data is measured against that of the consen
sus forecast reported by Money Market News Service.
The informal economic model used here is con
structed from variables available to forecasters before
the purchasing managers’ data are released. These
include several interest rate spread variables identified
in work by Bernanke, Estrella and Hardouvelis, and
others as reliable predictors of economic activity.8 Spe
cifically, the model includes the six-month commercial
paper rate, the spread between the commercial paper
and Treasury bill rates, the spread between corporate
BAA bonds and ten-year Treasuries, and the difference
between ten-year and three-month Treasury rates. Also
included are several “real” variables watched by payroll
forecasters: domestic auto sales, initial claims for
unemployment insurance, and the index of leading eco
nomic indicators. All told, this ad hoc economic model
has eight explanatory variables. The four interest rate
variables are entered contemporaneously and with six
lags, autos and claims enter currently and with a lag,
and both the index of leading indicators and the depen
dent variable enter with six lags. Adding the NAPM
employment index to this model yields a rigorous test of
its incremental explanatory power.9
Table 3 compares the explanatory power of the eco
nomic model, the NAPM employment index, and the full
7The m arkets a p p e a r to have a “flavor of the m onth" a p p ro a c h to
econom ic indicators, with m erc h a n d is e trad e, c onsum er prices,
producer prices, m oney growth, and the em p loym ent report eac h
getting top billing at various tim es. O verall, however, em ploym ent
seem s to be the most consistent leader.
8Ben S. B ernanke, “On the Predictive Power of Interest R ates and
Interest Rate S preads," Federal R eserve B ank of Boston N e w
E n g la n d E c o n o m ic R evie w , N o v em b e r-D e c e m b er 1990, pp. 51-68;
Arturo Estrella and Gikas A. H ardouvelis, “The Term Structure as a
Predictor of Real Econom ic Activity," Federal R eserve B ank of New
York, R esearch P aper no. 8 9 0 7 , M ay 1989.
®Each m odel was also tested using the N APM com posite index and
using m anufacturing em ploym ent as the d e p e n d e n t variab le, and
the results were very similar.
FRBNY Quarterly Review/Autumn 1991
65
m odel (th at is, the eco n o m ic m odel com bined w ith the
NAPM index) over tw o sam ple periods, one of extended
d u ra tio n (1959-91) and the o th e r lim ited to recent years
(1980-91). For each m odel the table show s the c o e ffi
cie n t on the e m plo ym e n t index w ith its t-s ta tis tic and
the overall fit of the m odel as m easured by the adjusted
R -s q u a re . S e v e ra l re s u lts are n o te w o rth y . F irs t,
alth o u g h both the eco n o m ic m odel and the purchasing
m an a g e rs’ index are h ig h ly s ig n ifica n t, the econom ic
m od el e x p la in s so m e w h a t m ore of th e va ria tio n in
e m p lo ym e n t grow th. T h is finding should not be s u rp ris
ing, however, because the NAPM data m easure only
grow th in the m anu factu rin g sector, w hile the econom ic
m odel has a rich array of e xp la na to ry va riables. S e c
ond, and m ore im p o rta n t, w hen the NAPM variable is
added to the econ om ic m odel, th is variable co n tin u e s to
be h ig h ly sig n ifica n t. In fact, the adjusted R -squares
su g g e st th a t the best m odel com bines the NAPM data
and the econom ic m o d e l .10
Even stro n g e r su p p o rt fo r the NAPM index com es
from c o m p a rin g it w ith th e co n s e n s u s fo re c a s t fo r
payroll em plo ym e n t grow th issued by M oney M arket
News S e rvice . T his in fo rm a l su rve y of data w atchers is
taken ju s t before the NAPM and e m ploym ent data are
released. The sam ple is lim ite d to the period since 1985
because of the difficulty in obtaining earlier data. Table 4
10ldeally, it w ould make sense to m odify the estim ation in two ways:
(1) sim plify the model by d roppin g the less significa nt lags on each
variable and (2) use unrevised data for the independent variables
(to d u p lic a te what is available to forecasters). Our purpose here,
however, is to stack the od ds against the NAPM index as much as
possible rather than to devise an optim al model. Furthermore,
prelim inary tests show that the results are not sensitive to either of
these changes.
Table 3
Explaining the Percent Change in Nonfarm
Payroll Employment
Sample:
1959-91
Model
NAPM*
NAPM index
0.022 .423
(16.9)
Economic model*
Full model§
E2
—
.429
0.021
(7.3)
.506
show s the results of th is c o m p a ris o n . A g a in , both the
NAPM index and the c o n s e n s u s e xplain a large p o rtio n
of the variation in e m p lo ym e n t, but th e best re su lts are
ob tained w hen the c o n se n su s and N APM are c o m b in e d
in the sam e equ a tio n . T h is fin d in g su g g e s ts th a t payroll
fo re ca ste rs sh o u ld m o d ify th e ir fo re ca st in lig h t of the
NAPM release. For exam ple, all else e q u a l, a 1 p e rc e n t
age p oint increase in the NAPM index sh o u ld in d u ce a
10,000 upward revision in exp e cte d p a yro ll e m p lo ym e n t
grow th.
So
fa r we have fo cu se d on the in -s a m p le fit of the
various e m p lo ym e n t m odels. T he u ltim a te te s t of th e se
equ a tio n s, however, is how th e y perform o u t of sam p le.
For each m odel a se rie s of o n e -m o n th -a h e a d fo re ca sts
is ca lcu la te d by using data from 1959 to 1984 and then
extending the sam ple forw ard one m onth at a tim e.
Table 5 show s the relative size of the p re d ictio n erro rs
fo r each of the m odels. As w ith the in -s a m p le tests,
adding e ith e r the c o m p o s ite or e m p lo y m e n t index to the
oth e r m odels reduces the average p re d ictio n errors.
The b e s f re s u lt c o m b in e s a s im p le a u to re g re s s iv e
Table 4
Explaining Employment Growth with the
NAPM Employment Index and the Consensus
Model
Model
NAPM index
Consensus model
NAPM com bined with
consensus m odel
Constant
NAPM
- 0 .7 7 5
(6.0)
-0 .0 3 5
(1-6)
0.020
(7.3)
- 0 .4 2 9
(4.0)
0.009
(3-7)
C onsensus
w
_
.408
—
1.173
(10.7)
.600
0.917
(7.5)
.658
Notes: Sample period is January 1985 to May 1991. The
de pend ent variable is the pe rcentage cha nge in total nonfarm
em ploym ent. C onsensus data are converted from cha nge to
pe rcentage change. A bsolute t-values are in parentheses.
Sample:
1980-91
NAPM*
R2
0.024 .522
(12.2)
—
.600
0.019 .646
(3.6)
*Values are coefficients on the NAPM index, with absolute tvalues in parentheses.
♦Includes the com m ercial paper rate, three interest rate spread
variables, auto sales, initial claims, the index of leading indica
tors, and lags of the dependent variable,
in c lu d e s both the NAPM employment index and all of the
econom ic variables.
66 FRBNY Q uarterly Review/Autumn 1991
Table 5
Out-of-Sam ple Prediction Errors for Payroll
Employment Growth
Model
W ithout
NAPM
_______ With NAPM
C om posite
Em ploym ent
—
.157
.145
Autoregressive m odel
.143
.141
.136
Econom ic m odel
.181
.158
.146
NAPM index only
Notes: Table shows the root mean square error for the January
19 85-M ay 1991 period. The “ autoregressive m o del" sim ply
uses six lags on the d e p e n d e n t variable.
m odel w ith the co m p o site in d e x .11
11Note that the econom ic model alone does the worst job of
predicting out of sam ple for this period. This result is consistent
with Bernanke's argum ent that a structural shift in the relationship
between the spread variables and econom ic activity occurred in the
1980s.
Table 6
Explaining Industrial Production Growth
Model
NAPM index
Consensus model
Hours
Economic model
NAPM index plus
consensus model
NAPM index
plus hours
NAPM index plus
econom ic model
Constant NAPM Consensus
-2 .7 2 6
(8 4 )
R2
0.055
(9 0 )
—
—
.371
—
0.926
(12.8)
—
.544
—
—
0,507
(11.4)
488
—
—
—
.618
—
.544
0.059
(1.3)
0.027
(4.6)
1.478
(1.6)
-0 .4 0 1
(0.9)
Hours
0.829
(7.2)
0.178
(4.4)
0.009
(1.1)
0.004
(5.1)
-0 .0 7 3
(0.1)
0.030
(2.1)
—
—
0.401
(8.8)
—
.568
.631
Notes: Sample period is January 1980 to May 1991. The depen
dent variable is the percentage change in the industrial
production index. The consensus is from Money Market News
Service. Absolute t-values are in parentheses.
Table 7
The NAPM Composite Index and
Real GNP Growth
Model
Constant
NAPM
NAPM index
-1 4 .4 6 0
(5.5)
0.546
(0.8)
0.326
(6 6 )
Consensus model
Economic model
NAPM index plus
consensus model
NAPM index plus
econom ic model
Consensus
-
R2
.361
—
0.832
(4.6)
.217
—
—
.658
-1 2 .0 0 0
(4-2)
0.261
(4.5)
0.363
(1.9)
.378
0.506
(0.1)
0.152
(1.7)
11.346
(4.6)
—
.669
Notes: Sample period is 1970-1 to 1989-11. The dependent vari
able is annualized one-quarter growth in real GNP. Absolute
t-values are in parentheses.
F o re ca stin g in d u s tria l p ro d u c tio n a n d re a l GNP
The NAPM data are also useful in fo re ca stin g in d u stria l
production and real GNP. Table 6 co m p a re s the e x p la n
ato ry pow er of fo u r m o d e ls of in d u s tria l p ro d u c tio n : the
NAPM p ro d u ctio n index, the grow th in em p lo ye e hours,
the M oney M arket c o n s e n s u s fo re ca st, and an e c o
nom ic m odel using the sam e va ria b le s d is c u s s e d in the
previous se ctio n . The t-s ta tis tic s on th e N APM c o e ffi
cie n ts su g g e st th a t the index adds s ig n ific a n tly to the
eco n o m ic m odel and the s im p le em p lo ye e hours m odel,
but th a t it is not a useful a d d itio n to the c o n s e n s u s
forecast. T h is finding sh o u ld not be a su rp ris e , however,
since the NAPM d ata are a va ilable to fo re c a s te rs before
the co n se n su s su rve y is ta ke n and th e re fo re sh o u ld
already be in co rp o ra te d into the co n s e n s u s fo re ca st.
Table 7 show s the re su lts of the final h o rse race. It
com pares the pow er of the co m p o s ite N APM index, the
eco n o m ic m odel, and a c o n s e n s u s fo re c a s t to p re d ict
grow th in real G N P For the e co n o m ic m odel the v a ri
ables used are the sam e as th o se in th e e m p lo y m e n t
and in d u stria l p ro d u ctio n e q u a tio n s, but each v a ria b le
enters co n te m p o ra n e o u s ly and w ith tw o lags. The c o n
se n su s data , c o m p ile d by th e A m e ric a n S ta tis tic a l
A s s o c ia tio n an d th e N a tio n a l B u re a u of E c o n o m ic
R esearch, are o n e -q u a rte r-a h e a d fo re ca sts, ta ke n in
the m iddle of the p receding quarter. A g a in , the re su lts
of th e c o m p a ris o n are g e n e ra lly s u p p o rtiv e of th e
NAPM data. The NAPM index p re d icts real G N P grow th
b e tte r than the c o n se n su s fo re ca st, a lth o u g h w orse
than the e co n o m ic m odel. The re la tive ly w eak p e rfo rm
ance of the co n se n su s is easy to e xp la in : the NAPM
and e co n o m ic m odels use u p -to -d a te in fo rm a tio n , w h ile
the co n se n su s is based o nly on in fo rm a tio n a vailable
before each quarter. A m ore im p o rta n t re su lt is th a t the
NAPM index c o n tin u e s to be s ig n ific a n t w hen added to
the oth e r m odels (a lth o u g h it is o n ly m a rg in a lly s ig n ifi
cant w hen co m b in e d w ith the e c o n o m ic m odel).
Conclusion
D espite its grow ing p o p u la rity, the NAPM index has
undergone ve ry little c ritic a l scrutiny. O u r re su lts s u g
gest th a t the index is a flaw ed but still useful indicator. It
is a poor leading in d ic a to r and, on its ow n, can be a
m is le a d in g m e a su re of s h o rt-ru n m o v e m e n ts in th e
econom y. In c o m b in a tio n w ith o th e r data, however, it is
very helpful in p re d ictin g co n te m p o ra n e o u s m a n u fa ctu r
ing activity. In sum , the index d e se rve s at le a st p a rt of
its reputation as a key e co n o m ic indicator.
FRBNY Q uarterly Review/Autum n 1991
67
Appendix: The Design of the NAPM Data Set
W ith one notable exception, the NAPM data have
received high praise in the literature.* Hoagland and
Taylor, for example, argue that the survey data “are
available sooner, are more reliable, and are much more
cost effective than government information.”* Klein and
Moore cite the early release of the data as an important
advantage; they recommend that the inventory index be
substituted for the official inventory data to improve the
timing of the index of leading indicators.5
Despite this strong support, the NAPM data need
improvement in at least three important areas.
Sampling bias
Unlike the surveys underlying official statistics, the pur
chasing managers’ survey does not use a scientific sam
ple. The NAPM data are drawn from hand-picked
members of larger, older firms rather than from a proba
bility sample. No attempt is made to account for industry
growth through the increase in the number of firms.
Furthermore, newer, fast-growing firms are added to the
sample only after they have become established in the
business, while declining firms remain in the sample until
they go out of business. In official statistics, both of these
downward biases are eliminated through adjustments
and rebenchmarking.
The sampling design has additional problems. The
sample is small, comprising less than 1 percent of the
association’s membership. Because of nonresponses
and the entry and exit of members, firms answering the
survey questionnaire can vary from sample to sample.
No attempt is made to correct for this variation by linking
companies that respond in both the current and previous
month— a procedure followed in the official statistics.
Finally, the data are never revised, implying that errors
are never corrected and late responses are never incor
porated into the data.
These sampling problems may explain the apparent
downward bias in the indexes. Theoretically, when aggretThe exce ption is Feliks Tamm, “An A genda for Inventories
Input to the Leading C om posite Index," in Kajal Lahiri and
Geoffrey H. Moore, eds., L eading Econom ic Indicators,
(C am bridge: C am b ridg e University Press, 1991), pp. 429-60.
Tamm points out a variety of flaws in the NAPM inventory
data. Some of his con cerns are discussed here.
*John H. H oagland and B arbara E. Taylor, “ Purchasing
B usiness Surveys: Uses and Im provements," Freedom o f
C hoice: Presentations from the 72nd Annual International
Purchasing C onference (O radell, N.J.: National A ssociation of
P urchasing M anagem ent, 1987), p. 1.
§Philip A Klein and Geoffrey H. Moore, "N .A .P M . Business
Survey Data: Their Value as Leading Ind icators," Journal of
P urchasing a n d M aterials M anagem ent, W inter 1988,
pp. 32-40.
FRBNY Q uarterly Review/Autumn 1991
gate activity is unchanged, the indexes should read
about 50 percent, with equal numbers of firms reporting
higher and lower activity. In fact, as Table 1 in the text
shows, the break-even values tend to be well below 50
percent. The results for the inventory index are particu
larly troubling. Not only is the break-even point well
below 50 percent, but the index also averages only 47.8
percent over the entire postwar period. This finding
implies that the level of inventories held by manufactur
ing firms has had a downward trend. Government statis
tics, on the other hand, show inflation-adjusted materials
and supplies for manufacturers roughly doubling over
this period.11
Backward-looking data
An important attribute of the NAPM data— its tim e
liness— is also one of its biggest shortcomings. Since
the results are released just after the end of each month,
the questionnaire must be answered in the middle of the
month. As a result, when respondents try to compare the
“current” month with the “previous” month, they may in
fact be comparing their impression of the last few weeks
(including part of the previous month) with their recollec
tions of the weeks before that interval. As the table below
shows, the timing of the responses means that in some
cases the NAPM data are more closely correlated with
lagged activity than with current activity.
Subjective responses
Survey respondents may not accurately assess whether
conditions are “better” or “worse ” Their answers may
reflect what should be or what is projected rather than
what is. The low average reading for inventories, for
!i C om paring the NAPM indexes for em ploym ent and output
with the official diffusion indexes for m anufacturing
em ploym ent and industrial productio n confirm s this bias.
Regression estim ates for the 1980-91 period show that the
break-even values for both officia l diffu sion series are closer
to 50 percent.
Correlation of NAPM Indexes and
M anufacturing Data
Official Series
Lead
C ontem porary
Lag
Industrial production
.410
.547
.614*
Employment
.569
.682
New orders
.211
.272
.720*
.426*
M aterials inventories
.481
.496-
.476
Notes: The sam ple period is January 1959 to May 1991.
The asterisk in dica tes peak correlation.
Appendix: The Design of the NAPM Data Set (continued)
example, may reflect the constant concern about exces
sive stocks rather than actual inventory management.
Wishful responses are particularly likely since the sam
ple is taken before the full results for the month are
known, and many of the questions refer to areas of the
firm not under the direct purview of the purchasing
manager.
The response that economic activity is “the same” is
equally problematic. Over time an average of more than
half the responses is “the same.” For example, from
January 1990 to June 1991 the percentage of “same”
responses was: new orders (46.5), production (53.9),
inventories (53.1), vendor deliveries (82.3), and employ
ment (64.4). Such stability at the firm level seems quite
unlikely in an unstable period for the economy as a
whole. Apparently, “the same” is a catch-all assessment
meaning “don’t know" and “no response” as well as “no
change."
Improving the data
In a real sense the NAPM data set is an uncut gem. By
using modern sampling and statistical techniques, the
association could greatly improve the accuracy of the
data. A probability sample should replace the handpicked sample; respondents should be linked from one
sample to the next; efforts should be made to reduce the
number of “sam e” responses and to ensure that
responses reflect actual activity; and respondents should
be encouraged to report on the current month’s activity
only. In addition, correctly accounting for inventories and
adjusting for lags and leads in the components would
improve the composite index.++ Of course, the NAPM
data neither could nor should mimic the official statistics:
this would require delays in its release and would put an
impossible burden on the respondents. The purchasing
managers’ association has made some efforts to refine
the data. Nevertheless, with the index increasingly in the
spotlight, further modernization is warranted.
ttThe inventory index should enter the com posite index as a
first difference rather than a level since it m easures a stock
rather than a flow. In a forthcom ing paper, Mark Flaherty and
the author present an alternative com posite index that has an
im proved track record in pre d ictin g industrial productio n, real
GNP, and the index of c o in cid e n t indica tors.
FRBNY Q uarterly Review/Autum n 1991
69
Treasury and Federal Reserve
Foreign Exchange Operations
May-July 1991
The dollar rose significantly in June and early July, only
to ease back during the next few weeks and end the
May-July reporting period with little net change. Over
the three months as a whole, the dollar rose about 2
percent against the mark, about 1 percent against the
yen, and just under 1 percent on a trade-weighted
basis.1
Shifting assessments of the strength of economic
recovery in the United States were important in stim
ulating movements of the dollar exchange rate during
the period. In addition, political turbulence in Eastern
Europe helped support the dollar against the mark
through most of the period, while intervention and evi
dence of international cooperation around the time of
the Group of Seven (G-7) summit meeting in July was
seen in the market as limiting the prospect of a continu
ing dollar rise.
The U.S. monetary authorities intervened on two
occasions to signal an interest in resisting the rise of
the dollar, selling a total of $150 million against marks
as part of their cooperation with other central banks.
The U.S. monetary authorities also engaged in offmarket transactions with foreign monetary authorities,
selling $8,548.5 million equivalent of their foreign cur
rency reserves for dollars.
A report p resen ted by Sam Y. Cross, Executive Vice President in
ch a rg e of the Foreign G roup at the Federal R eserve Bank of New
York and M a n a g e r of Foreign O p eration s for the System O pen
M arket A ccount. Vivek M oorthy was prim arily responsible for
preparation of the report.
’ The trad e -w e ig h te d basis is as m easured by the Federal Reserve
Board index.
70 FRBNY Quarterly Review/Autumn 1991
The dollar fluctuates without clear direction in May
As the period opened, sentiment toward the dollar was
favorable but market participants appeared uncertain
whether the dollar could extend the sharp rise that it
had experienced during the preceding two months. The
U.S. discount rate cut of 50 basis points to 5.5 percent
on April 30 had been unexpected, and that move gener
ated some downward pressure on dollar rates on May 1.
The U.S. employment data for April, released on May 3,
were stronger than expected, but on inspection, other
details of the report revealed areas of continuing weak
ness. In that environment, the dollar traded in a narrow
range for the first half of May.
Then, late in European trading on Friday, May 17,
Sweden’s Riksbank announced that it would link the
Swedish krona to the ECU, replacing its trade-weighted
basket of currencies, in which the dollar carried the
largest weight, with a basket composed entirely of Euro
pean Community currencies. Within a few. hours of the
announcement the dollar moved up by about four pfen
nigs against the mark as Swedish and other Scandina
vian entities rushed to adjust the currency composition
of their liabilities to that of the ECU by purchasing
dollars to repay dollar-denominated liabilities. With
Swedish interest rates relatively high, Swedish entities
had borrowed extensively abroad, partly to finance
domestic operations, confident that they were largely
shielded from exchange rate risk because the Swedish
authorities would limit the movement of the krona’s
exchange rate relative to the trade-weighted basket to
only a couple of percentage points. With the change in
the krona’s peg, the exchange risk these companies
would henceforth face on their dollar liabilities was
p e rceived to be m uch h ig h er than before, and they had
an in ce ntive to replace d o lla r-d e n o m in a te d lia b ilitie s
w ith tho se of E uropean cu rre n cie s more heavily repre
se n ted in the ECU. W ith U.S. m arkets still open a fte r
the Sw edish a n n o uncem e n t, and w ith the m ark rela
tive ly w id e ly trade d in the U.S. m arket, the pressures
re sulting from the May 17 e xch a n g e -m a rke t o p e ra tio n s
were co n ce n tra te d in d o lla r/m a rk tra n s a c tio n s , re su lting
in the sharp rise of the d o lla r a g a in s t the m ark. U nder
th e se c ircu m sta n ce s, there was som e in te rv e n tio n ; the
U.S. a u th o ritie s sold $50 m illio n on th a t Friday. A fte r the
w eekend, w ith p ressures c o n tin u in g , there was in te r
vention by a nu m b e r of fo re ig n ce n tra l banks. Soon
th e re a fte r th e m a rk e ts s e ttle d dow n and th e d o lla r
traded in a narrow range fo r the rest of th e m onth.
The dollar advances during June and early July
Chart 1
D uring ea rly June, a slew of U.S. e c o n o m ic in d ica to rs
were released th a t were g e n e ra lly m uch m ore favo ra b le
than expected, and m a rke t o b s e rv e rs began to ta lk
a bout the p o s s ib ility th a t the U.S. re co ve ry m ig h t be
m ore vig o ro u s than p re vio u sly a n tic ip a te d . In response,
e xp e cta tio n s of a fu rth e r d e clin e in U.S. in te re s t rates
faded and the d o lla r s ta rte d to rise. In p a rticu la r, on
June 7, w hen it was re p o rte d th a t May e m p lo y m e n t rose
well above e xp e cta tio n s, the d o lla r rose a lm o st tw o
pfennigs on the day to clo se at DM 1.7720.
D evelopm ents in G e rm a n y d u rin g Ju n e a lso te n d e d to
strengthen the dollar. News of G e rm a n y ’s firs t tra de
d e ficit since 1981, e vid e nce th a t in fla tio n was h ig h er
than pre vio u sly a n tic ip a te d — even before the im p o s itio n
of a c o n su m p tio n tax th a t w ould raise all m a jo r price
indexes fo r the u pcom ing m o n th s — and w hat som e saw
as the reluctance of the B u n d e sb a n k to ra ise o fficia l
in te re st rates all w eighed on the m ark. By late June, the
After generally rising through June and early
July, the dollar eased towards the end of
the period.
1991
German marks per
U.S. dollar
1.9
Japanese yen per
U.S. dollar
150
Chart 2
Three-Month Eurorate Differentials
Foreign Rate minus U.S. Rate
Percentage points
1.8
German mark
— Scale
/
/
1.7
A
v V
\
/ V
1.4 L i
„■
r J/
/
ii
/
\
!
f
\
German
- - . s '
'" X Japanese yen
130
m
11
11
m
I. I I I I I I
J
/
/\
/
r
/
\—' \
1 ,1 1 I J J 125
A
1991
Notes: The top chart shows the percentage change of the
weekly average of daily closing rates from May 1991 through
July 1991. The bottom chart shows the weekly average
closing rates for the German mark a nd the Japanese yen
from January 1991 through July 1991.
r\
^
135
/
/
l...i 1 1 1
s.^ ^
140
r // A "'
Ll i i i.. 1i i l 1 l
Jan
Feb
V*
Japan
i i
Mar
i
1i i i i 1.1111-i 11111
Apr
May
Jun
i i
I
Jul
1991
FRBNY Q uarterly Review/Autum n 1991
71
d o lla r had risen w ell above DM 1.80 in in tra d a y tra d in g .
W ith respect to the yen, the d o lla r showed its greatest
s tre ngth of the re p ortin g period during the first half of
June, breaking above the ¥ 1 4 2 level three tim es. The
dollar-ye n exchange rate reflected not only the more
buoyant o u tlo o k for the U.S. econom y but also concerns
in Jap anese fin a n cial m a rke ts a bout p o ssib le problem s
w ith banking and sto ck m arket practices.
As the d o lla r m oved up to levels not seen fo r more
than a year, m arket p a rtic ip a n ts becam e w ary of the
p o s s ib ility th a t som e action to curb the d o lla r’s rise
m ig ht be decid ed upon at the G -7 m eeting of finance
m in iste rs and ce n tra l bankers, scheduled to be held in
London on June 23. As a result, the m arket becam e
less co nce rn ed abou t the upside risk fo r the dollar, and
the curren cy traded in a narrow range as the m eeting
a pproached.
In the event, the G -7 m in iste rs and gove rn o rs issued
a co m m u niq ue tha t “ reaffirm ed th e ir c o m m itm e nt to
c o o p e ra te c lo s e ly , ta k in g a c c o u n t of th e n eed fo r
o rd e rly m a rke ts, if n e c e s s a ry th ro u g h a p p ro p ria te ly
c o n ce rte d action in foreign exchange m a rke ts.” M arket
p a rtic ip a n ts did not in itia lly construe the G-7 statem ent
as a firm co m m itm e n t to in te rve n e to re sist th e d o lla r’s
rise. But co m m e n ts fo llo w in g the m e e tin g by several
o f f ic ia ls , in c lu d in g J a p a n e s e F in a n c e M in is t e r
H a s h im o to , F rench F in a n c e M in is te r B e re g e v o y and
U.S. Treasury U nder S e c re ta ry M u lfo rd , re in fo rce d the
fe e lin g th a t the p o s s ib ility of in te rv e n tio n was being
a ctively co n sid ere d . R um ors a b o u t o ff-m a rk e t tra n s a c
tio n s b e tw e e n th e B u n d e s b a n k a n d th e F e d e ra l
R eserve, w hich were la te r co n firm e d by th e a u th o ritie s
(see below), were also ta ke n as in d ica tio n s th a t p re p a
rations to co n ta in the d o lla r’s rise were underway.
T hereafter, the d o lla r rem ained w ell below its e a rlie r
highs a g a in st the yen. The release on June 25 of data
in d ica tin g a la rg e r than exp e cte d rise in U.S. d u ra b le
goods o rd e rs fo r May te m p o ra rily s u p p o rte d the d o lla r
a gainst all cu rre n cie s. But the sp re a d in g ta lk of new
fin a n cial sca n d a ls in Japan was by th is tim e having o ff
se ttin g e ffe cts on the yen. On th e one hand, m arket
p a rtic ip a n ts cam e to e xp e ct th a t the a u th o ritie s m ight
move more q u ic k ly than o th e rw ise to low er in te re st
rates as Ja p a n e se share p rice s co n tin u e d to d e c lin e . In
fact, the Bank of Japan a n n o u n ce d a o n e -h a lf p e rc e n t
age p oint cut in its d is c o u n t rate, to 5.5 p e rce n t, on
Chart 3
Data released during the period first supported the dollar and later led it to ease. The employment report for
May was much stronger than anticipated while that for June was much weaker.
Thousands of jobs
Thousands of jobs
200
200 --------------------D ifference betw een A ctual and E xpected C hanges
150
100
50
-50
100
-150
Apr
1991
May
1991
Jun
Notes: The left panel shows the reported monthly changes in nonfarm payroll employment as of end-July. The right panel shows the actual
minus expected monthly changes in payroll employment: the actual changes used in computing these differences are the preliminary numbers
for April, May, and June 1991 reported on May 3, June 7, and July 5, 1991, respectively, while the corresponding expected changes are based
on surveys of market expectations.
http://fraser.stlouisfed.org/
72 FRBNY Q uarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
July 1. On the oth e r hand, m arket p a rtic ip a n ts viewed
the adverse im pact of the sto ck m a rk e t’s d ecline on
Japanese banks’ ca p ita l ratios as increasing the lik e
lihood that m ajor Ja pane se inve sto rs w ould be re p a tri
ating overseas funds to invest in new subo rd in a te d debt
in stru m e n ts th a t these banks w ould be issuing to shore
up th e ir c a p ita l p o sitio ns. These o ffse ttin g d e ve lop
m ents helped to keep the dollar-yen exchange rate
re la tively steady, tra d in g around ¥ 1 3 8 fo r the rem ainder
of the th re e -m o n th re p ortin g period.
The m ark, however, cam e u nder fu rth e r se llin g pres
sure at the end of June and ea rly July. The d o lla r
in itia lly stre n g th e n e d a g a in st the m ark in response to
the b e tte r than expected U.S. data fo r May durable
goods orde rs released on June 25. Two days later,
w hen, in response to a co n tro ve rsial G erm an co u rt
ruling, G erm an o fficia ls re p o rte d ly suggested th a t a
w ith h o ld in g tax on in ve stm e n t incom e m ight be rein
stated, the dolla r broke de cisive ly through the DM 1.80
level. The idea th at such a ta x — ve ry u n p o p u la r w hen it
was im posed in 1989 and q u ickly w ith d ra w n — m ight
a g a in be u n d e r c o n s id e r a tio n had an im m e d ia te
d e p re ssing im pact both on the m ark and m a rk -d e n o m i
nated assets. The DAX index of share p rices slum ped
2.5 percent the fo llo w in g day, and the d o lla r continued
to rise in the follow ing days to reach its high ag a inst the
m ark for the period and the year of DM 1.8427 in
Table 1
Federal Reserve
Reciprocal Currency Arrangements
in M illions of Dollars
Amount of Facility
Institution
A ustrian National Bank
National Bank of Belgium
Bank of C anada
National Bank of Denm ark
Bank of England
Bank of France
D eutsche B undesbank
Bank of Italy
Bank of Japan
Bank of M exico
N etherlands Bank
Bank of Norway
Bank of Sweden
Swiss National Bank
Bank for International Settlem ents
Dollars against Swiss francs
Dollars against other
authorized European currencies
Total
July 31, 1991
250
1,000
2,000
250
3,000
2,000
6,000
3,000
5,000
700
500
250
300
4,000
600
1,250
30,100
European tra d in g on J u ly 5.
The dollar gives back most of its gains during the
rest of July
Just as the d o lla r was reaching its highs of the period
a g a inst the m ark, m a rke t c o n fid e n c e in the U.S. re co v
ery began to weaken. U.S. econom ic data released during
the m onth no lo n g er p rovided u n a m b ig u o u s e vid e n ce of
e c o n o m ic re c o v e ry . T h e re le a s e on J u ly 5 o f th e
e m p lo ym e n t re p o rt fo r June, in p a rticu la r, show ed an
unexpected drop in em p lo ym e n t.
S im ultaneously, e x p e c ta tio n s began to grow th a t the
B undesbank w ould tig h te n m o n e ta ry p o lic y and p ursue
a more a ggressive m o n e ta ry sta n ce th a n p re v io u s ly
supposed. By then, the release of p rice fig u re s for
several G erm an sta te s th a t s u g g e s te d a sh a rp a c ce le r
ation in prices fo r “ c o re ” item s was seen as g ivin g the
B undesbank more reason fo r an e a rly p o lic y tig h te n in g
move. M arket p a rtic ip a n ts a p peared to be u n c e rta in
only about the e xte n t and tim in g of such a m ove—
w h e th e r it w ould com e before or a fte r the su c c e s s io n of
Dr. S ch le sin g e r to the P residency of the B u n d e sb a n k at
the end of July.
A g a in st this b a ckground, the d o lla r’s rise a g a in s t the
m ark sta lle d , and the e xchange rate flu c tu a te d w ith o u t
d ire ctio n ju s t above DM 1.80. However, on Ju ly 11, w hen
the B u ndesbank did not raise o ffic ia l in te re s t rates at its
b iw eekly m eeting and w hen a sh a rp drop in U.S. w eekly
u ne m p lo ym e n t in su ra n ce cla im s was re p o rte d , the d o l
lar jum ped back up to a lm o st DM 1.84. E arly the next
m orning, as the d o lla r co n tin u e d to move higher, foreign
ce n tra l banks co n d u cte d several rounds of in te rv e n tio n ,
se llin g d o lla rs a g a in st both m arks and o th e r cu rre n cie s.
A fte r the New York m a rke t ope n e d , the U.S. m o n e ta ry
a u th o r itie s a ls o p a r tic ip a te d , s e llin g $ 1 0 0 m illio n
a g a inst m arks. The w id e sp re a d p a rtic ip a tio n of ce n tra l
banks in the co n c e rte d in te rv e n tio n , ahead of th e G-7
sum m it m eeting the next w eek, and the fa ct th a t the
ce n tra l banks co n tin u e d to o p e ra te th ro u g h o u t the day
suggested to m arket p a rtic ip a n ts th a t the ce n tra l banks
were united in th e ir in te n tio n to curb the d o lla r’s rise. As
a result, the d o lla r d e clin e d by a b o u t five p fe n n ig s
during the day to clo se in New York at DM 1.7893. T his
episode of in te rv e n tio n , to g e th e r w ith an in c re a s in g ly
u n ce rta in U.S. e co n o m ic sce n a rio , set the to n e for
tra d in g for the rest of the m onth. T he d o lla r again
reached the DM 1.80 level on tw o o c c a s io n s the next
w eek in response to strong in d u s tria l p ro d u ctio n data
and C h a irm a n G re e n s p a n ’s s ta te m e n t in his s e m i
annual H um phrey-H aw kins te s tim o n y th a t a recovery
was under way, but it fa ile d to move higher.
The co m m u n iq u e released on Ju ly 17 a fte r th e G -7
sum m it m eeting re ite ra te d s u p p o rt fo r close c o o p e ra
tion in foreign e xchange m a rke ts, m o n e ta ry and fiscal
FRBNY Q uarterly Review/Autum n 1991
73
p o licie s to fo s te r low real in te re st rates, and S oviet
eco n o m ic and p o litica l tra n sfo rm a tio n . W hile the co m
m u nique had little im m e d ia te im pact on exchange rates,
it co n trib u te d to an atm osphere in w hich the fe a r of
co n ce rte d in te rve n tio n rem ained. In th a t en viro n m e n t,
the d o lla r did not stre n g th e n even w hen u nexpectedly
favorable housing s ta rts data were released later that
day.
D uring the rest of July, new U.S data releases brought
into q u e stio n the vig o r and even the s u s ta in a b ility of
eco n o m ic recovery. S e n tim e n t also spread am ong U.S.
fin a n cial m a rket a n a lysts th a t a sig n ifica n t d e clin e in
U.S. inflation, both actual and prospective, w ould be
reflected in a d e clin e in long bond yie ld s. Moreover,
sta te m e n ts by a va rie ty of U.S. officials, in clu d in g som e
FOM C m em bers, a b o u t the need to respond if M2
grow th rem ained w eak revived m arket e xp e cta tio n s that
U.S. s h o rt rates m ight still decline. As a result, the
d o lla r fell below DM 1.80 d u ring the third w eek of July
and to levels around DM 1.75 fo r the rest of the period.
The d o lla r closed the M ay-July reporting period about
2 p e rce n t higher ag a in st the m ark and 1 pe rce n t higher
a g a inst the yen. The d o lla r rose about 5.5 percent
a g a inst the S w iss fra n c as expe cta tio n s grew th a t the
m o n e ta ry a u th o ritie s in S w itze rla n d w ould not follow
th o se in G erm any by tig h te n in g m o n e ta ry policy. The
d o lla r eased very slig h tly a g a inst the C anadian d o lla r
as th e m a rke t cam e to b e lie ve th a t th e C a n adia n
a u th o ritie s w ould be more restrained about easing m o n
e ta ry po licy than w ould the U.S. m onetary a u th o ritie s.
D u rin g th e re p o rtin g p e rio d , th e U .S . m o n e ta ry
a u th o ritie s c o n d u c te d o ff-m a rk e t o p e ra tio n s d ire c tly
w ith foreign m o n e ta ry a u th o ritie s to adjust the foreign
cu rre n cy reserve assets of both the Exchange S ta b iliz a
tion Fund (ESF) and the Federal R eserve. The U.S.
a u th o ritie s exchanged $ 8 ,5 4 8 .5 m illion eq u ivale n t of
foreig n cu rre n cie s for d o lla rs:
• On June 25, the U.S. a u th o ritie s purchased a total of
$ 5 ,5 4 8 .5 m illio n a g a in s t G e rm a n m a rk s fro m th e
D eutsche B u nde sb a n k in spot and forw ard tra n sa ctio n s.
The U .S. and G e rm a n a u th o ritie s agreed th a t th e ir
respective h o ld in g s of G erm an m arks and d o lla rs were
in excess of cu rre n t needs and that it was to th e ir
m utual advantage to reduce those holdings. For each of
the se tra n sa ctio n s, 60 p e rce n t of the purchase was to
be executed fo r the a cco u n t of the Federal R eserve and
40 p e rce n t for the acco u n t of the ESF. A spot tra n s a c
tio n of $ 2 ,2 3 0 .5 m illio n s e ttle d on June 27. A forw ard
tra n sa ctio n of $55 6.2 m illio n se ttle d on Ju ly 29. The
rem aining forw ard tra n s a c tio n s are to be se ttle d during
the re m a in d e r of the ca le n d a r year.
• On Ju ly 16, the U.S. a u th o ritie s purchased a total of
http://fraser.stlouisfed.org/
74 FRBNY Q uarterly Review/Autumn 1991
Federal Reserve Bank of St. Louis
$3,000 m illio n a g a in s t a n o th e r fo re ig n c u rre n c y in spo t
and fo rw a rd tr a n s a c tio n s w ith a fo re ig n m o n e ta ry
a u th o rity. T he d o lla rs p u rc h a s e d w ere s p lit e q u a lly
betw een the ESF and the Federal R eserve. A spot
tra n sa ctio n of $1,000 m illio n se ttle d on Ju ly 18. Forw ard
tra n s a c tio n s to ta lin g $ 2 ,0 0 0 m illio n are to be se ttle d
during the next q u a rte rly re p o rtin g p eriod.
In a d d ition , in July, the ESF sold a to ta l of $130.2
m illio n e q u ivale n t of m arks a g a in s t SD R s. T he ESF also
purchased a to ta l of $ 2 3 0 .4 m illio n a g a in s t s a le s of
SDR s in tra n s a c tio n s w ith fo re ig n m o n e ta ry a u th o ritie s
in need of SDR s e ith e r fo r p a ym e n t of IM F c h a rg e s or
fo r repurchases. Both the sa le s and p u rch a se s of S D R s
were a rranged by the IMF.
P rim arily because of its a c q u is itio n of d o lla rs in the
fo re ig n c u rre n c y e x c h a n g e s a n d S D R tr a n s a c tio n s
d e scrib e d above, the ESF was able, a fte r the end of the
re p orting period, to re p u rch ase $ 2 ,5 0 0 m illio n e q u iv
a lent of foreign cu rre n cy w arehoused w ith th e F ederal
R eserve. These re p u rch ase s reduced the a m o u n t of
ESF foreign cu rre n cy b a la nce s w a re h o u se d w ith the
F ederal R e se rve fro m $ 4 ,5 0 0 m illio n e q u iv a le n t to
$2,000 m illio n e q u iva le n t as of th e end of A ug u st.
D uring the M ay-July p e rio d , th e Federal R e se rve re a l
ized profits of $147.5 m illio n from th e o ff-m a rk e t foreign
c u rre n c y e x c h a n g e s d e s c rib e d above. T he T re a su ry
realized profits of $ 60.3 m illio n , of w h ich $18.8 m illion
was from the o ff-m a rk e t fo re ig n c u rre n c y and SDR
e xchanges and $41.5 m illio n was from th e renew al of
ce rta in w arehousing o p e ra tio n s.
The Federal R eserve and the ESF re g u la rly in vest
Table 2
Net Profits ( + ) or Losses ( - ) on
United States Treasury and Federal Reserve
Foreign Exchange Operations
In M illions of Dollars
Valuation profits
and losses on
outstanding assets
and lia b ilities as of
April 30. 1991
Realized
A pril 30, 1991 to
July 31, 1991
Valuation profits and
losses on
outstanding assets
and liab ilitie s as of
July 31, 1991
Federal
Reserve
U.S. Treasury E xchange
S tabilizatio n Fund
+ 2,316.3
+ 570.6
+ 147.5
+ 60.3
+ 1,919.9
+ 321.4
Note: Data are on a value-date basis.
their foreign currency balances in a variety of instru
ments that yield market-related rates of return and that
have a high degree of quality and liquidity. A portion of
the balances are invested in securities issued by for
eign governments. As of the end of July, holdings of
such securities by the Federal Reserve amounted to
$7,807.7 million equivalent, and holdings by the Trea
sury amounted to the equivalent of $7,540.2 million
valued at end-of-period exchange rates.
FRBNY Quarterly Review/Autumn 1991
75
RECENT FRBNY UNPUBLISHED RESEARCH PAPERS*
9119.
H ickok, Susan, Jam es Orr, M.A. Akhtar, and K u rt C. W u lfe ku h ler. “ The
U ruguay Round of GATT Trade N e g o tia tio n s.” June 1991.
9120.
Kasm an, B ruce, and A nthony R o drigues. “ F in a n cia l R eform and M o n e ta ry
C o n tro l in Japan.” M ay 1991.
9121.
H ickok, Susan, Juann H. Hung, and K urt C. W ulfekuhler. “An E xpanded,
C o in te gra te d M odel of U.S. Trade.” June 1991.
9122.
Lown, C ara S. “An In d ica to r of Future Inflation E xtra cte d from the S te e p
ness of the In te re st Rate Y ield C urve along Its E ntire L e n g th .” W ith Je ffre y
A. Frankel. June 1991.
9123.
G aske, M. E llen. “ S ources of F lu ctu a tio n s in Long-Term E xp e cte d Real
R ates of Interest: E vidence from th e U.K. Indexed B ond M a rk e t.” J u ly 1991.
9124.
H arris, Ethan S. “ Tracking the E conom y w ith the P urchasing M a n a g e rs’
Index.” A u g u st 1991.
9125.
S kile s, M arilyn E. “ S ta b iliz a tio n and F inancial S e cto r R eform in M e xico .”
A u g u st 1991.
9126.
C h a rre tte , S u sa n . “ C a p ita l F lig h t from D e b to r N a tio n s W h e n L a b o r Is
M o b ile .” S e ptem ber 1991.
9172.
S a n ta m a ria , M a rco . “ P riv a tiz in g
N ovem ber 1991.
9 128.
W izm an, T h ie rry A. “ R eturns on C a p ita l A sse ts and V a ria tio ns in E conom ic
G row th and V o la tility: A M odel of B ayesian L e a rn in g .” N ovem ber 1991. W ith
C onnel R. Fullenkam p.
9 129.
Akhtar, M .A., and E than S. H arris. “ The S up p ly-S id e C o n se q u e n ce s of U.S.
F iscal Policy in the 1980s.” N ovem ber 1991.
S o cia l S e c u rity : T he
C h ile a n C a s e .”
fS in g le co p ie s of these papers are available upon request. W rite R esearch P apers,
Room 901, R esearch Function, Federal Reserve Bank of New York, 33 Liberty S treet,
New York, N.Y. 10045.
FRBNY Q uarterly Review/Autumn 1991
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