Full text of Review (Federal Reserve Bank of St. Louis) : July/August 1993

View original document

The full text on this page is automatically extracted from the file linked above and may contain errors and inconsistencies.

Does the Exchange Hate Regim e Affect
the Economy?
Central Bank Independence and Economic
Perform ance
Hypothesis Testing w ith Near-Unit Roots:
The Case of Long-Run P u rch asin g-P ow er
Parity
The Effect of Mortgage Refinancing on
Money Demand and the Monetary
Aggregates

THE
FEDERAL
A RESERVE
lU N k o t

ST.I/H IS

F e d e r a l R e s e r v e B a n k o f St. L o u is
R eview
July/August 1993

In This Issue . . .
D o e s th e E x c h a n g e R a te R e g im e A ffe ct th e E c o n o m y ?
Terence C. Mills and G eoffrey E. W ood
Teren ce C. Mills and Geoffrey E. Woods exam ine what makes pegged ex
change rates so attractive; and consider empirically the relationship b e
tw een the exchange rate regime and a num ber of key m acroeconom ic
variables, such as output, prices and interest rates, to see w h eth er any
system atic relationship exists betw een the behavior of these variables and
the exchange rate regim e. They focus their study on the United Kingdom,
because the U.K. experienced a wide variety of exchange rate regim es
over the period covered by the data, and because Britain has not en
dured hyperinflation or recessions as severe as those in some oth er coun
tries. Their findings support the conclusion that the exchange rate regime
has not been a source of volatility fo r the m acroeconom ic perform ance of
the British economy.

C e n tr a l B a n k I n d e p e n d e n c e a n d E c o n o m ic P e r f o r m a n c e
Patricia S. Pollard
Central bank independence is becom ing popular, as evidenced by the
num ber of countries that have recently enacted legislation removing their
central banks from governm ent control. Examining the econom ic rationale
for this popularity, Patricia S. Pollard employs em pirical studies to reveal
that countries with independent central banks tend to experience low in
flation with no loss of econom ic grow th. On the other hand, theoretical
studies illustrate that an independent bank may increase policy conflicts
within a country, resulting in poor econom ic perform ance. W eaknesses in
both types of studies, how ever, may limit their ability to prove or disprove
the usefulness of central bank independence regarding econom ic p erfor
m ance. Pollard concludes that the relationship betw een central bank in
dependence and the econom y is not fully understood.

H y p o th e s is T e s tin g w ith N ear-U n it R o o ts : T h e C ase o f L o n g -R u n
P u rc h a s in g -P o w e r P a rity
Michael J. Dueker
As a principle, it has long been asserted that the quantity of goods one
can buy with a given currency, such as the dollar, should be equal across
countries, at least in long-run equilibrium. This condition, known as longrun purchasing pow er parity (PPP), has been subjected to num erous em pir
ical tests. One source of disagreem ent in statistical tests of PPP has been the
choice of null hypothesis: Tests whose null hypothesis is that PPP holds
often fail to reject PPP, while tests whose null hypothesis is that PPP fails

JULY/AUGUST 1993

often com e to the opposite conclusion. Thus, there is a danger in testing
only one null hypothesis fo r a broad set of countries, failing to reject it,
and concluding that the evidence is clearly for or against PPP.
In this article, Michael J. Dueker tests post-1973 monthly data from major
countries using a long-memory model. The advantage of this approach is
that one can test both null hypotheses with the model and dem onstrate
that it is unclear w h eth er long-run PPP holds, because real exchange
rates have near-unit roots, w hich may preclude strong conclusions as to
w hether real exchange rates are m ean-reverting.

T h e E f f e c t o f M o rtg a g e R e f in a n c in g o n M o n ey D e m a n d a n d th e
M o n e ta ry A g g r e g r a te s
Richard G. Anderson
During the last two years, lower interest rates have stimulated extensive
refunding of long-term debt, sharply increasing the relative volume of
financial transactions. Mortgage refinancing has been a highly visible part
of those transactions. Richard G. Anderson examines the effect of recen t
waves of mortgage refinancing on the demand for liquid deposits and
grow th of the m onetary aggregates.
Mortgage servicers may hold unscheduled principal paym ents received
following a refinancing in liquid deposits as long as six weeks prior to
rem ittance to the investors who own the underlying m ortgage-backed
securities. In addition, the grow th of oth er checkable deposits also ap
pears to have been affected by fluctuations in m ortgage refinancing,
perhaps because of households converting hom e equity to cash. The persistance of these increased demands for liquid balances illustrates that all
transactions are not completed instantaneously, as is implicitly assumed.
Anderson finds that the increased m ortgage refinancing accounts for a
great deal of the volatility of M l’s grow th during the last two years,
although not for its continued strong underlying trend.

All non-confidential data and programs for the
articles published in Review are now available
to our readers. This information can be ob
tained from three sources:

1. FRED (F e d e r a l R e s e r v e E c o n o m ic Data),
an e le c tr o n ic b u lle tin b o a rd s e rv ic e .
You can access FRED by dialing 314-6211824 through your modem-equipped PC.
Parameters should be set to: no parity,
word length = 8 bits, 1 stop bit and the
fastest baud rate your modem supports,
up to 14,400 bps. Information will be in
directory 11 under file name ST. LOUIS
REVIEW DATA. For a free brochure on
FRED, please call 314-444-8809.

FEDERAL RESERVE BANK OF ST. LOUIS

2 . T h e F e d e ra l R e s e r v e B a n k o f St. L o u is
You can request data and programs on
either disk or hard copy bv writing to:
Research and Public Information Division,
Federal Reserve Bank of St. Louis, Post
Office Box 442, St. Louis, MO 63166.
Please include the author, title, issue date
and page numbers with your request.
3 . I n t e r - u n i v e r s i t y C o n s o r tiu m f o r
P o l i t i c a l a n d S o c ia l R e s e a r c h
(IC P S R ). M em ber institu tions can r e
quest these data through the CDNet
O rd er facility. N onm em bers should
w rite to: ICPSR, In stitu te fo r Social
R esearch, P.O. Box 1248, Ann A rbor, MI
4 8 1 0 6 , or call 3 1 3 -763-5010.

Terence C. Mills and G eo ffrey E. W ood
Terence C. Mills is a professor at the University of Hull. Geoffrey
E. Wood, a professor at City University Business School in
London, was a visiting scholar at the Federal Reserve Bank of
St. Louis. Kevin White provided research assistance. The authors
would also like to thank Mervyn King and Paul Mizon for their helpful
comments.

Does the Exchange Rate Regime
Affect the Economy?

J . T SEEMS TO BE a general rule that countries
wish to peg their exchange rates but sometimes
have floating rates thru st upon them. On three
occasions during the tw entieth cen tury—the
breakup of the international gold standard in
the 1930s, the breakup of the Bretton Woods
system in the 1970s and m ost recently the exo
dus of countries (notably Britain) from the ex
change rate m echanism (EBM) of the European
Econom ic Community (EEC)—external pressures
led to the demise of fixed rate schem es and
their replacem ent by some degree of exchange
rate flexibility. In each case, the passing of the
fixed rate schem e was m ourned and within
relatively short periods a new fixed rate plan
was advanced to replace its fallen predecessor.
In view of these failures, how ever, it is reason
able to ask: W hat makes pegged exchange rates
so attractive?
Recently, in the context of the ERM, tw o argu
m ents have been advanced. Exchange rate fixity
is, as David Hume described in 1752, a way of
importing another country’s m onetary policy.1
In the case of the ERM, the deutsche mark
served as the system ’s anchor currency, and
'See Hume (1970).
2We do not consider whether a single currency really is a
natural development of fixed exchange rates.

Germany’s low inflation rate was supposed to
spread throughout the EEC. M oreover, the
ERM’s m em ber nations believed that the Bun
desbank’s reputation would provide some credi
bility to the anti-inflation com m itm ent of other
central banks and th erefo re reduce the costs of
lowering inflation throughout the EEC. A se
cond motive fo r adopting fixed exchange rates
has been the claim that they, and ultimately a
single currency, are im portant to the EEC’s Sin
gle M arket Program m e.2 The logic is that the
full benefits that could accrue from the free
intra-European movem ent of goods, labor and
capital will be realized only with a fixed ex
change rate regim e.3 A third argum ent, not em
phasized recently but im portant on earlier
occasions, is that econom ic perform ance—
grow th, inflation or any oth er im portant
m easure—is b etter under a fixed exchange rate
system .4 This third argum ent differs from the
second in that it identifies no specific causal
chain from exchange rate regime to economic
perform ance.
But does the exchange rate regime m atter for
econom ic perform ance? That is the question
“This was an important motivation for Britain’s return to the
gold standard in 1925, for example.

3This argument usually takes as axiomatic that trade crea
tion will outweigh trade diversion.

JULY/AUGUST 1993

addressed in this paper. W e exam ine empirically
the relationship betw een the exchange rate re
gime and a num ber of key m acroeconom ic vari
ables to see w hether any system atic relationship
exists betw een the behavior of these variables
and the exchange rate regime. W e have chosen
to investigate this question for the United King
dom because data over long periods are availa
ble for the variables we wish to exam ine and
because the United Kingdom experienced a
wide variety of exchange rate regim es over the
period covered by these data.

TRADE AND THE EXCHANGE
RATE REGIME
The claim that exchange rate flexibility ham
pers international trade in goods and in capital
and thus depresses w elfare and perhaps grow th
is based on the existence of uncertainty.
It is argued that rem oving the possibility of
exchange rate change will rem ove an im portant
nontariff b arrier, because the possibility of ex
change rate changes will deter some traders
and investors altogether, w hereas others will
have to pay a substantial cost to fix the domes
tic value o f their foreign curren cy receipts.
Floating exchange rates, in oth er words, are b e
lieved to impose additional volatility, and hence
costs, on international m arkets. If this is co r
rect, a case fo r pegged exchange rates exists,
and the case is particularly strong fo r any
group of countries (such as the EEC) that wants
to encourage mutual international trade and in
vestment.
The proposition seem s unexceptional, and for
a num ber of years studies supported the propo
sition. For example, Cushman (1983) and deGrauwe and deBellefroid (1987), w hich are
representative of the early literature, found that
floating rates did impede trade. But as time
passed, an increasing num ber of studies sup
ported it.5
By the early 1990s, not only had evidence
shifted to support the notion that floating ex
change rates do not impede trade, but Feldstein
(1992) even w ent so fa r as to suggest that float
5Examples of these studies are Gotur (1985); the IMF’s
(1984) extension of Cushman (1983) to cover the bilateral
trade of the seven largest industrial countries; Bryant
(1987), Bailey, Tavlas and Ulan (1986 and 1987), Bailey and
Tavlas (1988) and Ascheim, Bailey and Tavlas (1987).
6Haberler (1986) suggested the same thing some years
earlier.

FEDERAL RESERVE BANK OF ST. LOUIS

ing rates are m ore favorable to trade than are
fixed rates.6 Attention is thus directed to other
reasons fo r favoring pegged exchange rates.
T h ere are two rath er distinct types of effects
of exchange rate fixing. The first arises because
if a fixed exchange rate is in place, it is unlikely
to stay fixed without policy actions. These can
take several form s. Most com mon are foreign
exchange intervention and short-term interest
rate manipulation. Accurate figures on official
intervention or the stock of foreign exchange
reserves are not always available. Interest rate
figures, how ever, are available, and several
authors have found that unpredictable interest
rate variability increased after exchange rates
are pegged.7 These actions in tu rn m ake money
grow th m ore volatile, and this can have im por
tant consequences for the economy. It may c re
ate additional u ncertainty about the future
behavior of the price level and thus about real
rates of retu rn, which would affect investment.
If future prices w ere uncertain, wage bargain
ing would be m ore complex because it would
be harder to judge the future purchasing pow er
of an agreed money wage. This u ncertainty
would also affect nominal variables. Bisk-averse
investors would be m ore reluctant to buy
governm ent bonds because they would be un
certain what the coupons would be w orth and
what the capital would be w orth at m aturity.
This would raise nominal interest rates, the cost
of debt service and thus the taxes necessary to
service the debt. All these factors could have an
adverse effect on long-term grow th, depressing
its trend.
In summary, the choice of exchange rate re
gime could affect the long-run behavior of the
economy, influencing trends or cycles in im por
tant m acroeconom ic variables.
If the choice of exchange rate regim e does not
have these long-run consequences, then in
term s of m acroeconom ic effects, all that the
choice of exchange rate regime does is shift the
distribution of short-run fluctuations from one
m arket to another. This is the second type of
effect noted above.
The question we examine is w h eth er any as7See Batchelor and Wood (1982), Wood (1983) and Belongia
(1988). Wood and Belongia’s research was conducted in
the context of the ERM. In Wood (1983) there was an ex
ception to this—Erie (South Ireland) after it joined the
ERM. Unpredictable interest rate variability fell in that
country, although it increased in every other ERM member
country.

sociation exists betw een the exchange rate re
gime and the trend or cyclical behavior of some
key m acroeconom ic variables—in other words,
w hether there is any evidence for the first type
of effect. If no such association exists, then the
only m acroeconom ic consequence of the choice
of exchange rate regim e is the change in the
distribution of short-term volatility betw een the
foreign exchange m arket and the short-term
money m arkets. If, in contrast, such an associa
tion exists, then the choice of exchange rate re
gime may be a m acroeconom ic policy decision
of considerable im portance for national well
being.8
It is now appropriate to present the data we
use fo r exploring this question. W e then exa
mine the properties of those data in light of the
preceding discussion.

THE STOCHASTIC PROPERTIES
OF U.K. MACROECONOMIC SERIES
ACROSS EXCHANGE RATE
REGIMES
In this section we consider the stochastic
properties of five m ajor U.K. m acroeconom ic
series since the mid-nineteenth century. T he ex
change rate regim es since then have encom
passed every possible type except the crawling
peg. Until 1914, the United Kingdom was on the
gold standard. That was suspended (that is, the
United Kingdom left the standard but with the
declared intention of returning) at the outbreak
of W orld W ar I in 1914. A fter the war, the
United Kingdom implemented a deliberate, dis
cussed and announced policy of a retu rn to the
gold standard at the prew ar parity. M onetary
policy and foreign exchange intervention w ere
used to this end, and the policy succeeded in
1925. The United Kingdom left the gold stan
dard in 1931, how ever, and the exchange rate
floated with varying degrees of intervention un
til the outbreak of W orld W ar II in 1939.9 The
rate was then pegged to the U.S. dollar. A fter
8lt is, of course, possible that the exchange rate regime is a
product of the behavior of the economy; it need not be an
exogenous choice.
9For a review and evaluation of explanations that have been
advanced to explain the United Kingdom’s abandonment of
the gold standard, see Capie, Mills and Wood (1986a).

the war, the United Kingdom joined the Bretton
Woods system. Several sterling devaluations oc
curred under Bretton Woods, but sterling did
not finally float until 1972. Again, there w ere
varying degrees of intervention under this re
gime of dirty floating, but the United Kingdom
did not form ally peg sterling until it joined the
ERM in 1990 after shadowing the deutsche
m ark in 1988 and 1989. The United Kingdom
subsequently left the ERM in 1992 to float once
m ore. The series we exam ine across these vari
ous regim es are output, prices, money, and
short- and long-term interest rates.
Our particular interest, and the focus of the
em pirical w ork that follows, is the trend and
the cycle in output and prices primarily, but
also in money and interest rates. W e look to see
how these variables have behaved over our
close to a century-and-a-quarter of data, seeking
changes in trend and changes in cyclical pat
tern. W hen these are identified, we examine
w hether any of these changes are associated
with exchange rate regim e changes and, if so,
consider why this might be.

Output
Annual output in the United Kingdom (meas
ured in logarithms) over the period 1 8 5 5 -1 9 9 0
is shown in figure 1. Detailed econom etric ana
lyses of this series in Mills (1991) and Mills and
Wood (1993) show that it can be represented as
the sum of a segmented linear trend, with
breaks at 1918 and 1921 and a stationary, au
toregressive, cyclical com ponent.10 Thus these
results indicate that if output can be decom
posed as y t = fut + n(, then the trend function
is
(1)

= o + Bt + X lt + A,D2(,
t
tD

w here Dit = ( t - T ) if / > T and zero otherw ise.
The identified breakpoints are at Tt = 64 and
T2 = 67, w hich coincide with 1918 and 1921. The
cyclical com ponent, n(, on the other hand, is
found to be adequately modeled as an AR(2)
separate. The cycle comprises fluctuations about a horizon
tal average; growth is all in the trend. This separation is
consistent with most views of the cycle, but it should be
noted that some scholars see the cycle as an integral part
of the growth process. For an example, see Schumpeter
(1950).

1
0Testing for stationarity has no direct economic significance.
Rather, it lets us separate the cycle from the trend. The
notion is that the trend and the cycle are economically

JULY/AUGUST 1993

Figure 1

Annual U.K. Output (1855-1990)
Logarithms

process, leading to the fitted model (standard
erro rs shown in parentheses),
(2) Yt =

3.474 + 0.01961 - 0.1137D U +
(0.026)
(0.0007)
(0.0103)
0.1170D 2I + n
( 0 . 01 0 1 )

nt = 1.099nt t - 0.346nl 2 + a,
(0.083)
(0.083)'

on the output series in figure 1, and we thus
conclude that, apart from the th ree years im
mediately after W orld W ar I, during w hich the
series fell dramatically, the stochastic process
generating output has rem ained rem arkably sta
ble. Output is a trend stationary process, ir
respective of the exchange rate regime in force.

Prices

This model has some simple properties. Trend
grow th is 1.96 percent per y ear until 1919 and
2.29 percent per year from 1922 on, with the
level of trend output falling 28.3 percent in the
intervening three years. The com ponent n, im
plies that output exhibits stationary cyclical fluc
tuations around the trend grow th path, with
cycles averaging 8.1 years. The residual stan
dard erro r of the equation is 2.33 percent.

Figures 2 and 3 present plots of the (logarith
mic) U.K. price level annually from 1870 to
1990 and monthly from Jan uary 1922 to May
1992, respectively, excluding the w ar years
from 1940 to 1945. Unit root tests, calculated
over various sample periods, provide little or no
evidence against the hypothesis that prices are
difference stationary, that is, 1(1).11 The
post-1973 era may differ and is discussed later.

The trend com ponent is shown superimposed

Tw o aspects of price behavior are w orth fur-

1 Details of these tests and similar tests for the other series
1
investigated are reported in Mills and Wood (1993).

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

Figure 2

Annual U.K. Price Level (1870-1939)
Logarithms

Annual U.K. Price Level (1946-1990)
Logarithms

JULY/AUGUST 1993

Figure 3

U.K. Price Level (1922-1939)
Logarithms

U.K. Price Level (1946-1992)
Logarithms

FEDERAL RESERVE BANK OF ST. LOUIS

th e r investigation. The first is the behavior of
the price level before the United Kingdom aban
doned the gold standard in 1931. Mills (1990)
analyzes the long gold standard period from
1729 to 1931 and obtains an estim ate of the lar
gest autoregressive root of 0.93, identical to that
obtained for the sh orter sample beginning in
1870. The corresponding unit root test, though,
rejects the unit root null hypothesis at the 5
percent significance level, and the process
found to generate the price level (an autoregres
sion of order two) yields cycles of around SO
years, close to the long swings thought to have
characterized prices during this period.12
The second aspect concerns the post-1946 b e
havior of prices. Figures 2 and 3 show the ser
ies to have undergone slope changes around
1973 and 1983; possible explanations fo r these
are discussed in the next paragraph and in the
Interpretation and Conclusions section. Statisti
cally, this behavior is typical of an 1(2) process,
and repeating the unit root tests for the (log
arithmic) price changes, that is, for inflation,
yields some evidence that postw ar prices can be
modeled as an 1(2) process (evidence that infla
tion is nonstationary), particularly fo r the postBretton W oods era beginning in 1973.
The results are th erefore suggestive of the
U.K. price level undergoing two shifts in its
generating process. The first might be associat
ed with the abandonm ent of the gold standard,
shifting the series from 1(0) to an 1(1) process.
(From figure 3 it is in fact clear that prices did
not start a secular increase until m id-1933, some
two years after the move from the gold stan
dard.)13 A stable price level is certainly in acco r
dance with what would be expected under the
gold standard (or, in principle, any commodity
standard). T h ere w ere fluctuations in the supply
of gold, but in countries such as the United
Kingdom, which had developed and stable bank
ing systems, these fluctuations had only modest
price level effects. The system was to some ex
tent self-stabilizing. If prices w ere falling (the
value of m oney rising) because the supply of
gold was falling short of demand, th ere was an
incentive to produce m ore gold. And if prices
w ere rising (the value of m oney falling), then as
the costs of gold production rose relative to
what the m onetary authorities would pay for
12See Cagan (1984) for an extended discussion of this view.
13See the discussion in Capie, Mills and Wood (1986a).

gold, the incentive to produce gold would
diminish.14 The second shift is around 1973 and
could be associated w ith both the move to float
ing exchange rates and the first oil price shock.

M on ey
Figure 4 plots annual observations of the
logarithms of M3 from 1871 to 1912 (the only
aggregate apart from the m onetary base availa
ble for this period), and figure 5 plots monthly
observations of M3 from 1922 to 1989, exclud
ing the w ar years. From a battery of unit root
tests, we found that, for all sample periods in
vestigated, the null hypothesis of a unit root
cannot be rejected. M oreover, the series is in
deed 1(1) because we could not establish that
fu rth er differencing was required for stationarity.

Interest Rates
Figures 6 through 8 plot m onthly observations
of short-and long-term interest rates from 1870
to 1992, excluding w ar years and related peri
ods of interest rate restrictions.
From the results of unit root tests, we find
that since the lifting of restrictions after W orld
W ar II both short- and long-term interest rates
have been 1(1) processes, but their behavior b e
fore 1939 is rath er different. Both are station
ary betw een 1932 and 1939, but during the
1920s long-term rates are stationary (1(0)) and
short-term rates are 1(1), w hereas before 1914
the orders of integration are reversed.15

Trend and Cycle Decom positions
Has the variability about trend of the series
altered across regimes? This is an im portant
question because o f the widespread belief that
floating exchange rates increase volatility in
prices, interest rates, and econom ic activity and
are in some general sense destablizing. To an
sw er this question, w e need to decompose each
series into trend and cycle components. There
are many ways to do this, ranging from using a
predeterm ined moving average to calculate
trend to designing a signal extraction filter
based on the stochastic process generating the
data and a set of assumptions relating to the b e
havior of the unobserved components.
1
5Capie, Mills and Wood (1986b) provides an extended dis
cussion of the behavior of these two interest rate series in
relation to the Stock Conversion of 1932.

14See Barro (1979) and Rockoff (1984) for a discussion of this.

JULY/AUGUST 1993

Figure 4

Annual U. K. Money Supply: M3 (1871-1912)
Logarithms

For output, equation (1) provides the appropri
ate decomposition. Table 1 thus reports the
standard deviations of the cyclical com ponent nt
fo r a variety of sample periods. The sample
periods shown w ere chosen by two quite dis
tinct criteria—output trend change and ex
change rate regim e alteration. The 1922 break
was used because after the 1 9 1 9 -2 2 discontinui
ty, output resum ed a new trend, 1 8 5 5 -1 9 1 3
w ere gold standard years, and 1 9 2 5 -3 1 w ere
years during which the United Kingdom was
either on or com m itted to returning to the gold
standard. The period comprising 1 8 5 5 -1 9 1 3 and
1 9 2 2 -3 1 is the same period omitting w ar and
the postw ar years of the break in output's
trend. The period 1 9 2 2 -3 1 has a stable output
trend com bined with com m itm ent to gold; the
period 1 9 2 2 -3 9 has a stable output trend with a
change in exchange rate regime. The period
1 9 3 2 -9 0 is our whole sample period after gold.
The years 1 9 3 2 -3 9 and 1 9 4 6 -9 0 are, of course,
the same period excluding the W orld W ar II
years. The period 1 9 4 6 -9 0 is simply postwar;
194 6 -7 2 is Bretton Woods; and 1 9 7 3 -9 0 is the
period of various degrees of float. (Further sub
division of the series to examine the association

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

with various exchange rate regim es more
minutely, although appealing, is ruled out by
many of these regim es having too few output
observations for our statistical techniques.)
From all these statistics, one gets the im pression
that variability about trend has increased during
the tw entieth century. In particular, the aban
donment of the gold standard in 1931 seems to
have been accompanied by an increased varia
bility of output about trend, even after the w ar
years are excluded. In summary, the standard
deviation almost doubled (from 2.87 percent to
5.49 percent) after 1931. But it should be noted
that variability fell after the pound floated in
1972. From 1946 to 1972 the standard deviation
was 4.45 percent; from 1973 to 1990 it was 3.64
percent.
For the other series, we have presented evi
dence of shifts in the stochastic processes
generating them, so signal extraction techniques
would be rath er difficult to apply. W e have
chosen therefore to use a technique that has
proved popular in recen t years for re-examining
the stylized facts of m acroeconom ic time series,
namely the detrending filter proposed for use in
econom ics by Hodrick and Prescott and used,

Figure 5

U.K. Money Supply: M3 (1922-1939)
Logarithms

U.K. Money Supply: M3 (1946-1989)
Logarithms

JULY/AUGUST 1993

Figure 6

U.K. Interest Rates (1870-1913)
Percent

Figure 7

U.K. Interest Rates (1922-1939)
Percent

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

Figure 8

U.K. Interest Rates (1954-1992)
Percent

for example, in Kydland and P rescott.16 This is
an alternative to the method used earlier in the
paper for separating a series into trend and cy
cle. It is described in the appendix, which also
contains a summary of w hen this method is ap
propriate and w hen it may be misleading.
Tables 2 through 5 report statistics assessing
the variability of the trend and cycle com po
nents of the price level, money supply and
short- and long-interest rates, and figures 9
through 12 present graphs for these com po
nents. Although these tables report results from
the exam ination of monthly data, the b reak
points are at year ends except for 1992, w hose
data end with June.
This choice of breakpoints reflects two con
siderations. The first relates to w hen an ex
change rate regime changed. Does chan ge for
our purposes relate to w hen the change was
formally announced or to w hen it becam e ex
pected and affected behavior? The latter is the
m ore significant, but it is not clear a p riori
16See Hodrick and Prescott (1980) and Kydland and Prescott
(1990).

w hen it would be. Nor as it turns out does
detailed exam ination of the data case by case
give clear-cut answ ers.17 Accordingly, the simple
expedient of using calendar years as b reak
points was adopted, on the grounds that using
other dates close to these would not change the
results.
For the interw ar years, the trend of the price
level was relatively flat, with a slow decline un
til 1933 and an upward drift th ereafter. The cy
clical component, in contrast, is relatively
volatile, no doubt, in view of the unchanged be
havior of money, reflecting the changes in ex
change rate policy in the United Kingdom, as
well as the disturbed external environm ent. Not
only did the interw ar years include the Great
Depression in the United States, with the as
sociated severely depressing effects on the
prices of commodities, but in continental Eu
rope there w ere inflations—hyperinflations in
some cases—civil w ar and revolutions. M ean
while Britain's exchange rate regim e was chang
ing rapidly. Betw een 1919 and 1925 there was a
17See Mills and Wood (1993). For a subset covering the
years 1870-1939, see Capie and Wood (forthcoming).

JULY/AUGUST 1993

Table 1

Table 3

Variability of the Cyclical Component
of Output_________________________

Component Variability of Money
1922.01-1939.12
1946.01-1992.05
1946.01-1972.12
1973.01-1989.06

Standard
Deviation

Period

1855-1913
1855-1931
1855-1913 and 1922-1931
1922-1931
1922-1939
1932-1939
1932-1990
1932-1939 and 1946-1990
1946-1990
1946-1972
1973-1990

2.69
2.87
2.95
3.45
3.93
3.58
5.49
4.27
4.42
4.45
3.64

7.90
10.04
9.27
11.29

0.10
1.08
0.27
0.67

0.02
0.02
0.02
0.02

sample mean

sx: sample standard deviation
sT: sample standard deviation of trend component
sc : sample standard deviation of cycle component

Table 4

Component Variability of Short-Term
Interest Rates

Table 2

Component Variability of Prices

1922.01-1939.12
1946.01-1992.05
1946.01-1972.12
1973.01-1992.05
x:

3.12
4.60
3.88
5.61

0.09
0.94
0.30
0.51

0.02
0.01
0.01
0.01

sample mean

1870.01-1913.12
1922.01-1931.12
1932.01-1939.12
1954.01-1972.12
1973.01-1992.04
x:

2.79
3.75
0.86
5.42
11.43

1.21
1.07
0.71
1.76
2.44

0.77
0.68
0.51
1.54
1.84

0.84
0.76
0.53
0.66
1.22

sample mean

sx: sample standard deviation

sT: sam p le standard deviation ot trend com ponent

sT: sample standard deviation of trend component

sc: sample standard deviation of cycle component

sc : sample standard deviation of cycle component

com m itm ent to retu rn to gold at the prew ar
parity, and the exchange rate rose steadily
toward that. Gold was abandoned in 1931, and
the exchange rate th ereafter floated with vari
ous degrees of intervention until the outbreak
of w ar in 1939.
A fter 1946, the trend is smooth and m onoton
ic, and the cyclical com ponent is less volatile
than before. Trend money is rath er similar to
trend prices. Its variability is stable throughout
the sample period, supporting the suggestion
that external factors w ere im portant in interw ar
price volatility.
Pre-1914 trend interest rates fluctuate around
18We have noted this result in a series of previous papers.
Mills and Wood (1982) suggested it was due to the stable
price expectations provided by the gold standard. Mills

FEDERAL RESERVE BANK OF ST. LOUIS

3 percent, although the far greater stability of
long-term rates is reflected in the almost cons
tant com ponents of this series relative to short
term rates.18 Volatility is indeed fairly stable un
til 1972, after which both trend and cycle com
ponents becam e considerably m ore variable.

INTERPRETATION AND
CONCLUSIONS
W hen discussing the preceding findings, it is
convenient to consider the trend and cyclical
behavior of each series together. W e start with
output. As noted previously, the trend grow th
and Wood (1992) was unable to reject this hypothesis after
exhaustive testing.

Table 5

Component Variability of Long-Term
Interest Rates
X

1870.01-1913.12
1922.01-1931.12
1932.01-1939.12
1954.01-1972.12
1973.01-1992.04
x:

2.94
4.45
3.31
6.35
11.32

0.23
0.13
0.35
1.71
1.91

0.22
0.11
0.31
1.67
1.74

0.04
0.13
0.14
0.24
0.61

sample mean

sx: sample standard deviation
sT: sample standard deviation of trend component
sc: sample standard deviation of cycle component

o f output changed from 1.96 percent per year
to 2.29 percent per year betw een 1919 and
1922. Speculating on what produced that w el
come change is outside the scope of this paper.
W hat we would note is the stability of the
post-1922 trend in the face of a wide variety of
m onetary experiences and exchange rate re
gimes, a finding clearly consistent with the longru n neutrality of money.
In contrast to that long-run neutrality, the cy
clical behavior w as affected. The variability of
output rose substantially w ith the abandonm ent
of the gold standard. The significance of this is
discussed later.
Turning now to prices, what do we find? The
first notable feature is the essentially flat trend,
with long swings around it, under the gold stan
dard. More dram atic and equally revealing
about the nature of the m onetary regim e is the
post-1946 period. The trend of prices was posi
tive after 1946, accelerated sharply around 1973
and slowed around 1983. T he United Kingdom
w ent to a floating exchange rate in 1972, but at
around the same time there was also the first
oil price shock and the H eath-Barber m onetary
expansion. That the acceleration of prices was
the result of these factors ra th er than the new
exchange rate system is suggested by the slow
ing of prices around 1983, w hen the United
1
9The role of the exchange rate regime in the 1970s episode
is also discussed in Williamson and Wood (1976). The con
clusion that the exchange rate regime was not at fault was
also, by different means, argued there.

Kingdom was still under a floating rate regime
but had a governm ent strongly committed to
reducing inflation by introducing money supply
targets and a com m itm ent to budget balance
over the cycle.19 The cyclical com ponent of
prices becam e m uch sm oother and was un
affected by the exchange rate regim e; its varia
bility was unchanged from 1946 to 1992 and
identical >ver subperiods and the period as a
whole.
And finally, interest rates. The striking con
trast is betw een the behavior in the pre-W orld
W ar II period, w hen long-term rates w ere sta
ble and short-term rates w ere volatile, an obser
vation usually interpreted as reflecting
expectations o f long-run price level stability and
behavior in the post-1972 period, w hen inflation
first accelerated and then slowed, and both in
terest rate series displayed markedly increased
variability.20
How do these findings as a whole b ear on the
hypothesis that the exchange rate regime is not
a source of volatility? They support it. Of the
variety of exchange rate regim es after 1913 (we
turn to the gold standard in a moment), none
seemed to increase the volatility of any series
examined to any significant extent. T he policy
changes necessary to hold rates pegged may
have appeared in foreign exchange reserves, a
series that we did not examine because reliable
data w ere not available. The policy changes did
appear in movements that had higher frequ en
cies than the trends and cycles we isolate.21 In
terest rate cyclical variability did increase with
the move to floating exchange rates in 1972, but
there arc num erous other factors to explain
this. Sho ks to the price of oil disturbed finan
cial m arkets very substantially in this period.
Two other shocks w ere superimposed on the oil
price shocks. T h ere was a com m itm ent to
reduce inflation—particularly after 1979. W hat
this meant in term s of the operation of m one
tary policy was unknow n, so the commitm ent
increased u ncertainty for a time. And further,
m onetary targets w ere adopted. These affected
how the authorities used short-term interest
rates; and as com m itm ent to m onetary targets
planations of the Gibson paradox that depend on slowmoving price expectations.
21See Batchelor and Wood (1982), Wood (1983) and Belongia
(1988).

20See Mills and Wood (1982). Fisher (1930), Friedman and
Schwartz (1982), and Mills and Wood (1992). All have ex

JULY/AUGUST 1993

Figure 9

Price Level Trend and Cycle (1922-1992)
Logarithms

Logarithms

Figure 10

Money Supply: M3 Trend and Cycle (1922-1989)
Logarithms

Logarithms

1922 26 30 34 38 42 46 50 54 58 62 66 70 74 78 821986

FEDERAL RESERVE BANK OF ST. LOUIS

Figure 11

Short Interest Rate Trend and Cycle (1870-1992)
Percent

Percent

3H------- 1------ 1------ 1------ 1------ 1------ 1------ 1------ 1------ 1------ 1------ 1------ r ~ -50
1870

1900

1990

Figure 12

Long Interest Rate Trend and Cycle (1870-1992)
Percent
20

15-

10-

--3 .0
Trend

1
------1-----1---- 1-----1-----1-----1---- 1-----1-----1-----1---- r^-4.o
1870

1900

1990

JULY/AUGUST 1993

becam e increasingly credible, the relationship
betw een m ovements in short- and long-term
rates changed.22
It cannot but be observed that there was
greater stability of output, interest rates and
prices under the gold standard than under any
subsequent exchange rate regime. But, of
course, the gold standard was m ore than an ex
change rate regim e. It was a system, a set of
rules, for the conduct of m onetary policy. As
Bordo (1993) w rote, "The gold standard rule can
be viewed as a form of contingent rule or a
rule with escape clauses. The m onetary authori
ty maintains the standard—that is, keeps the
price of the cu rren cy in term s of gold fixed—
except in the event of a well-understood em er
gency, such as a m ajor w ar or a financial crisis.
In wartim e it may suspend gold convertibility
and issue paper m oney to finance its expendi
tures, and it can sell debt issues in term s of the
nominal value of its cu rren cy on the und er
standing that debt will eventually be paid off in
gold. The rule is contingent in the sense that
the public understands that the suspension will
last only fo r the duration of the w artim e em er
gency plus some period of adjustment. It as
sumes that afterw ard the governm ent will
follow the deflationary policies necessary to re
sume payments at the original parity.” It may be
consistent with this interpretation of the gold
standard that with the floating exchange rate of
the 1970s, output variability fell, but not to
w here it had been under the gold standard. The
argum ent would be that m onetary policy was
now clearly focused on internal objectives and
not subject to the vicissitudes of a multitude of
shocks from the outside world.
All in all, then, it appears clear that the ex
change rate regim e in the United Kingdom has
not been a source of volatility for the main
m acroeconom ic variables. For that reason we
need not consider why exchange rate regim es
might affect real econom ic perform ance—in the
United Kingdom they did not. The case for
a fixed rate regim e in the United Kingdom ap
parently m ust depend only on its traditional
source of support—the desire to import price
level perform ance.
It is, of course, im portant to consider w hether
these results generalize to other economies.
22lnitially, rises in short-term rates produced rises in long
term rates. But as markets became convinced that the
authorities were serious, short- and long-term rates started
to move much more independently of each other.

FEDERAL RESERVE BANK OF ST. LOUIS

There is virtually no feature of the U.K. econo
my to indicate that they should not.23 The com
position of output is not unusual; the U.K.
econom y has always been fairly open. It was a
dominant econom y internationally for only a
modest part of our period, and it has not gone
through hyperinflation or recessions as severe
as those in some other economies, so such
problems cannot have biased our results.
Though we would not claim that our findings
are m ore than those of a case study, we would
suggest that they are findings we would not be
surprised to see roughly repeated in studies of
other countries.

REFERENCES
Aschheim, Joseph, Martin J. Bailey and George S. Tavlas.
“ Dollar Variability, the New Protectionism, Trade and Finan
cial Performance,” in Dominick Salvatore, ed., The New
Protectionist Threat to World Welfare (NorthHolland 1987),
pp. 424-49.
Bailey, Martin J., George S. Tavlas and Michael Ulan. “ Ex
change Rate Variability and Trade Performance:
Evidence for the Big Seven Industrial Countries,”
Weltwirtch Archiv (Volume 122,1986), pp. 466-77.
_______ “ The Impact of Exchange-Rate Volatility on Export
Growth: Some Theoretical Considerations and Empirical
Results,” Journal of Policy Modeling (Spring 1987),
pp. 225-43.
Barro, Robert J. “ Money and the Price Level Under the Gold
Standard,” Economic Journal (March 1979), pp. 13-33.
Batchelor, Roy A., and Geoffrey E. Wood. “ Floating Exchange
Rates: The Lessons of Experience," in Roy A. Batchelor
and Geoffrey E. Wood, eds., Exchange Rate
Policy (St. Martin’s Press, 1982), pp. 12-34.
Belongia, Michael T. “ Prospects for International Policy Coor
dination: Some Lessons for the EMS,” this Review
(July/August 1988), pp. 19-27.
Bordo, Michael D. “ The Gold Standard, Bretton Woods, and
Other Monetary Regimes: A Historical Appraisal,” this
Review (March/April 1993), pp. 123-87.
Bryant, Ralph C. International Financial Intermediation
(The Brookings Institution, 1987).
Cagan, Phillip. “ Mr. Gibson’s Paradox—Was it There?” in
Michael D. Bordo and Anna J. Schwartz, eds., A Retro
spective on the Classical Gold Standard, 1821-1931 (Univer
sity of Chicago Press, 1984), pp. 604-10.
Capie, Forrest H., Terence C. Mills and Geoffrey E. Wood.
“ What Happened in 1931?” in Forrest H. Capie and
Geoffrey E. Wood, eds., Financial Crises and the World
Banking System (St. Martin’s Press, 1986a), pp. 120-48.
23This is also suggested by the similar structure of models
used to explain and predict the economies of a wide range
of countries.

_______ “ Debt Management and Interest Rates: the British
Stock Conversion of 1932,” Applied Economics (Volume 18,
1986b), pp. 1111-26.

King, Robert G., and Sergio Rebelo. “ Low Frequency Filtering
and Real Business Cycles,” University of Rochester Work
ing Paper No. 205 (1989).

Capie, Forrest H., and Geoffrey E. Wood. “ Money in the
Economy, 1870-1939,” in Roderick Floud and Donald
McClosky, eds., An Economic History of Britain (Cambridge
University Press, forthcoming).

Kyland, Finn E., and Edward C. Prescott. “ Business Cycles:
Real Facts and a Monetary Myth,” Federal Reserve Bank
of Minneapolis Quarterly Review (Spring 1990), pp. 3-18.

Cushman, David O. “ The Effects of Real Exchange Rate Risk
on International Trade,” Journal of International
Economics (August 1983), pp. 45-63.
deGrauwe, Paul, and Bernard deBellefroid. “ Long-Run Ex
change Rate Variability and International Trade,” in Sven
Arndt and J. David Richardson, eds., Real Financial Link
ages Among Open Economies (MIT Press, 1987),
pp. 193-212.
Feldstein, Martin. “ The Case Against EMU,” The Economist
(June 13, 1992), pp. 19-22.
Fisher, Irving. The Theory of Interest (Macmillan, 1930).
Friedman, Milton, and Anna J. Schwartz. Monetary Trends in
the United States and the United Kingdom (University of
Chicago Press, 1982).
Gotur, Padma. “ Effects of Exchange Rate Volatility on Trade:
Some Further Evidence,” IMF Staff Papers (September 1985),
pp. 475-512.
Haberler, Gottfried. “ The International Monetary System,” The
AET Economist (July 1986).
Harvey, A.C., and Alfred Jaeger. “ Detrending, Stylized Facts
and the Business Cycle,” London School of Economics
Discussion Paper No. EM/91/230, 1991.
Hodrick, Robert J., and Edward C. Prescott. “ Postwar U.S.
Business Cycles: An Empirical Investigation,” CarnegieMellon University Discussion Paper No. 451 (1980).
Hume, David. “ Of the Balance of Trade, ” in Eugene Rotwein,
ed., David Hume, Writings on Economics (The University of
Wisconsin Press, 1970), pp. 60-77.
International Monetary Fund. “ Exchange Rate Volatility and
World Trade,” Occasional Paper No. 28 (1984).

Mills, Terence C. "Are Fluctuations in U.K. Output Transitory
or Permanent?” Manchester School of Business (Volume
59, 1991), pp. 1-11.
_______ “A Note on the Gibson Paradox During the Gold
Standard,” Explorations in Economic History (July 1990), pp.
277-86.
Mills, Terence C., and Geoffrey E. Wood. “ Capital Flows and
the Excess Burden of the Exchange Rate Regime,” Hull
Economic Research Paper (1993).
_______ “ Econometric Evaluation of Alternative Money
Stock Series, 1880-1913,” Journal of Money, Credit and
Banking (May 1982), pp. 265-77.
_______ “ Money and Interest Rates in Britain From
1870-1913,” in Nick Crafts and Stephen Broadberry, eds.,
Britain in the International Economy 1870-1939 (Cambridge
University Press, 1992), pp. 199-217.
Rockoff, Hugh. “ Some Evidence on the Real Price of Gold,
Its Costs of Production, and Commodity Prices,” in Michael
D. Bordo and Anna J. Schwartz, eds., A Retrospective on
the Classical Gold Standard, 1821-1931 (The University of
Chicago Press, 1984), pp. 613-44.
Schumpeter, Joseph Alois. Capitalism, Socialism, and
Democracy, 3rd ed. (Harper, 1950).
Williamson, John, and Geoffrey E. Wood. “ The British Infla
tion: Indigenous or Imported?” American Economic Review
(September 1976), pp. 520-31.
Wood, Geoffrey E. “ The European Monetary System: Past
Developments, Future Prospects and Economic Rationale,”
in Richard Jenkins, ed., Britain and the EEC (Macmillan,
1983).

Appendix
The Hodrick-Prescott Filter
The filter proposed by Hodrick and Prescott
(1980) has a long tradition as a m ethod o f fitting
a smooth curve through a set of points, ver
sions of it being used as an actuarial graduation
formula. Given the traditional decomposition
y, = M + n,> the trend series /i( is obtained as the
,
solution to the problem of minimizing
T

(2) y, = A[^/+2- 4 M + r6 + A-J)M
,+J
,

+ M ,J

Using the lag operator B, defined such that
=
this can be w ritten as
( 3 ) Yt = X[B 2- 4 B 1+ ( 6 + X 1 - 4 B + B % t
)
= m

- B f a - B 1
)2+ \

( 1)

so that if an infinite series of y values w ere
available,
would be given by the two-sided
moving average
co

with respect to
\2,
x
\ir The first ord er con
dition for this minimization problem is

( 4) u

—

Lj
j.

or Y
j

, a

< -j ’

— or

-j

JULY/AUGUST 1993

w here the weights can be calculated from

(5) a(B) = [A ( 1 - B H 2 -B _1)2+ 1 ] " !.

by the structural model
(6)

y t = nt + n,
M = M,-. + v ,-,
,

King and Bebelo (1989) provide expressions for
the ay w hich do not take a simple form . For
tunately, Hodrick and Prescott (1980) provide an
algorithm that rem oves the need to calculate
the moving average weights and so allows the
trend to be computed w hen only a finite num
ber of y observations are available. This al
gorithm was employed to com pute the
decompositions used here, noting that the cycli
cal com ponent can be obtained by residual as
nl = y t-\xt. Typically, following Hodrick and Pres
cott, A is set at 100 if annual data are used or
1,600 if quarterly or monthly data are used.1
Harvey and Jaeger (1991), fo r example, show
that the filter a(B)yt can be interpreted as being
the optimal estim ate of
w hen y t is generated

1See Hodrick and Prescott (1980).

FEDERAL RESERVE BANK OF ST. LOUIS

V, =
n'NID (0,0;), £~MD (0,ct|), A = o 2lo*
n
An observed series may not be generated even
approximately by such a model, and even if it
is, the ratio of the tw o innovation variances
may be very different from the assumed value
of A Harvey and Jaeg er argue that the Hodrick.
Prescott filter may create spurious cycles, dis
tort the estim ates of the com ponents or both.
King and Bebelo argue in similar vein, although
they focus on the calculation of sample moments
of the estim ated trend and cycle components.
Given these strictures, we emphasize that our
use of the filter is purely for exploratory pur
poses outside the confines of any explicit model.

Patricia S. Pollard
Patricia S. Pollard is an economist at the Federal Reserve
Bank of St. Louis. Heather Deaton and Richard D. Taylor
provided research assistance.

Central Bank Independence and
Economic Performance

J n RECENT YEARS MANY countries have
adopted or made progress tow ard adopting
legislative proposals removing their central
banks from governm ent control, that is, making
them independent. Between 1989 and 1991,
New Zealand, Chile and Canada enacted legisla
tion that increased the independence of their
central banks. T he 1992 Treaty on European
Union (M aastricht Treaty) requires European
Community (EC) m em bers to give their central
banks independence as part of establishing the
European M onetary Union. As a result, EC
countries that do not yet have strongly indepen
dent central banks have introduced legislation
or announced their com m itm ent to make their
central banks m ore independent.1 Furtherm ore,
in recen t months the governm ents of Brazil and
Mexico have announced their intentions to in
troduce legislation to create m ore independent
cen tral banks.
In view of these developments, it might seem
reasonable to conclude that unambiguous links
had been established betw een econom ic p erfor
1To meet the level of independence prescribed by the Maas
tricht Treaty, a central bank must be prohibited from taking
instructions from the government. The term for central
bank governors must be set at a minimum of five years,
although it can be renewed. In addition, the central bank
must be prohibited from purchasing debt instruments
directly from the government (that is, in the primary
market) and from providing credit facilities to the govern
ment. Both Denmark and the United Kingdom have
reserved the right to decline membership in the European
Monetary Union. Thus neither country has introduced

m ance and the degree of central bank indepen
dence. Interestingly, how ever, the two postW orld W ar II star perform ers among the indus
trialized econom ies—Germany and Japan—have
d ifferent levels of cen tral bank independence.
The German Bundesbank is viewed as one of
the most independent central banks in the
world, w hereas the Bank of Japan is seen as
m ore subject to governm ent control. Thus the
contrast betw een the movement to grant central
banks m ore independence and widely different
degrees of independence across the m ajor econ
omies raises several questions. Among these are:
W hy is the idea of an independent central bank
popular? Are th ere econom ic benefits of having
an independent central bank?
This paper exam ines em pirical and theoretical
studies of central bank independence to address
these questions. Empirical research ers have de
vised m easures of independence to focus on the
relationship betw een central bank independence
and a country's econom ic perform ance. T h eo
retical studies have modeled the strategic belegislation to ensure conformity of their central banks with
the Maastricht provisions.
For a detailed analysis of the institutional status of the
central banks of the EC countries, see the Committee of
Governors of the Central Banks of the Member States of
the European Economic Community (1993).

JULY/AUGUST 1993

havior of m onetary and fiscal policym akers to
be able to com pare an econom y’s perform ance
w hen policym akers cooperate in setting policies
with its perform ance w hen they do not
cooperate.
The next section of this paper presents a sur
vey and evaluation of em pirical studies.-Next,
theoretical studies are presented and evaluated.
The final section exam ines the extent to which
these studies eith er explain the cu rren t move
ment tow ard greater central bank independence
or highlight unresolved questions in this debate.

EMPIRICAL STUDIES: CENTRAL
HANK INDEPENDENCE AND
ECONOMIC PERFORMANCE
Inflation and Central Bank
Independence
As a broad generalization, interest in central
bank independence was motivated by the belief
that, if a central bank was free of direct politi
cal pressure, it would achieve low er and m ore
stable inflation.2 Bade and Parkin (1985) con
ducted one of the first em pirical studies of this
link. The authors used data for 12 Organization
for Econom ic Cooperation and Development
(OECD) countries in the post-Bretton W oods era
and m easured the degree of central bank in
dependence according to the extent of govern
m ent influence over the finances and policies of
the central bank.3 The degree of financial in
fluence on the central bank was determ ined by
the governm ent’s ability to set salary levels for
m em bers o f the governing board of the central
bank, to control the cen tral bank’s budget and
to allocate its profits. The degree of policy in
fluence was determ ined by the governm ent’s
ability to appoint the m em bers of the central
bank governing board, governm ent rep resen ta
tion on this board, and w h eth er the govern
ment or the central bank was the final policy
authority. Countries w ere given a rank of one
through fou r in each category, with fou r being
the highest level of central bank independence.
Bade and Parkin concluded that the degree of
financial independence of the central bank was
2Buchanan and Wagner (1977) point out that even an in
dependent central bank may not be immune from political
pressures and thus exhibit an inflationary bias.
3The 12 OECD countries are Australia, Belgium, Canada,
France, Germany, Italy, Japan, Netherlands, Sweden, Swit
zerland, United Kingdom, and United States.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

not a significant determ inant of inflation in the
post-Bretton Woods period. Policy indepen
dence, how ever, was seen as an im portant de
term inant of inflation because the two countries
with the highest degree of policy independence
(Germany and Switzerland) had inflation rates
significantly below those of all other countries
in the sample. They found no significant d iffer
ences in inflation perform ance among countries
with low er rankings of independence in the
post-Bretton Woods era.
Alesina (1988) used the Bade and Parkin (1985)
index but added the following fou r countries:
Denm ark, New Zealand, Norway and Spain. He
found, as hypothesized, that th ere was generally
an inverse relationship betw een average infla
tion rates and the level of central bank in
dependence.
Grilli, Masciandaro and Tabellini (1991) created
two indexes of central bank independence—one
based on econom ic m easures of independence
(with a scale ranging from zero to eight), and
the other based on political m easures of in
dependence (with a scale ranging from zero to
seven).4 The political factors w ere similar to
those identified by Bade and Parkin. The eco
nomic factors considered w ere the ability of the
governm ent to determ ine the conditions under
w hich it can borrow from the central bank and
the m onetary instrum ents under the control of
the cen tral bank. The data set com prised 18
OECD countries over the period 1 9 5 0 -8 9 .5 For
the period as a whole, Grilli, M asciandaro and
Tabellini found that econom ic independence
was negatively related to inflation. Political in
dependence also had a negative correlation with
inflation, but the relationship was not statistical
ly significant. Breaking the data into fou r
decade-long subperiods, they found that neither
m easure of independence had a significant e f
fect on inflation in the first tw o decades. In the
1970s both m easures of independence w ere sig
nificant, w hereas in the 1980s only the econom
ic independence m easure was significant.
Alesina and Summ ers (1993) calculated a
m easure of central bank independence by aver
aging the indexes created by Bade and Parkin,
4ln both measures the scale is increasing in the level of in
dependence.
5Grilli, Masciandaro and Tabellini add Austria, Denmark,
Greece, New Zealand and Portugal to Bade and Parkin’s
group of countries and eliminate Sweden.

Figure 1

Average Inflation: 1955-1988
Percent

•I■Spain
8-

New Zealand
7-

Italy
^United Kingdom
Australia *
* Denmark
*** France / Norway / Sweden

.j. Japan
Canada

54-

Belgium •h

The
Netherlands

United States
Switzerland
Germany

-----------------1
------------------------- 1----------------

2
3
Index of Central Bank Independence
Source: Alesina and Summers (1993).

and Grilli, Masciandaro and Tabellini.6 The
countries included w ere the same as in Bade
and Parkin with the addition of Denm ark, New
Zealand, Norway and Spain. The sample period
was 1 9 5 5 -8 8 .7 As in the previous studies, they
found a negative correlation betw een the level
of central bank independence and the rate of
inflation (figure 1). They also found that the
m ore dependent a central bank was, the greater
the variability in inflation (figure 2). This, they
argued, was a result of a correlation betw een
the level and variability of inflation.
Cukierman (1992) provided an extensive analy
sis of central bank independence and its rela

tionship to inflation perform ance using data for
1 9 5 0 -8 9 . Unlike previous studies, he used not
only legal m easures of central bank indepen
dence, but also practical m easures of the level
of independence. One such m easure was the
frequency of turnover of the central bank
governors. Another m easure of practical in
dependence was based on answ ers from a ques
tionnaire completed by qualified individuals at
the central banks.8 Cukierm an’s analysis is the
most com prehensive to date, not only because it
incorporates inform ation about the actual level
o f independence a central bank enjoys in prac
tice, but also because it includes a sample of 70
countries.9 Cukierman concluded that "central

6See Bade and Parkin (1985) and Grilli, Masciandaro and
Tabellini (1991).
7See Alesina (1988). Alesina and Summers report that the
results of their study are the same if the data period is res
tricted to 1973-1988, the post-Bretton Woods era.
8The sample period for the questionnaire data was 1980-89.
9The questionnaire data were available for only 24
countries.

JULY/AUGUST 1993

Figure 2

Variance of Inflation: 1955-1988
Percent
40
•fr Italy
30Spain +
•I* United Kingdom
Australia / France
•I* Japan

New Zealand
20-

j,Sweden
Norway j .
4 Canada
B e lg iu m *
Denmark

10 -

•fr United States

*T h e
Netherlands

^Switzerland
Germany

2
3
Index of Central Bank Independence

Source: Alesina and Summers (1993).

bank independence affects the rate of inflation
in the expected direction.”10 This result was also
found by Cukierm an, W ebb and Neyapti
(1992).11

Central Bank Independence and
the Real E conom y

dence and econom ic output. If an independent
central bank can produce low er inflation than a
dependent central bank, does this com e at the
cost of low er output? Conversely, are dependent
central banks attem pting to exploit a short-run
Phillips Curve relationship, accepting higher in
flation in order to achieve higher output?

Although most of the em pirical w ork focused
on the relationship betw een central bank in
dependence and the rate of inflation, some
studies examined the link betw een indepen-

Grilli, Masciandaro and Tabellini (1991) found
no system atic effect of central bank indepen
dence (using either of their two indicators) on
the grow th rate of real output. Alesina and

10Cukierman did not actually use the rate of inflation, but the
rate of depreciation of the real value of money, defined by
the following formula:
d, =

1 + n,
where n, is the inflation rate in period t. The use of d, as
noted by Cukierman, moderates the effects of hyper
inflation on the results.
11Capie, Mills and Wood (1992) also studied the link between
inflation and central bank independence. Their data set
consisted of 12 countries, with the data series beginning
between 1871 and 1916 and ending in 1987. Central

FEDERAL RESERVE BANK OF ST. LOUIS

banks were classified as either dependent or independent
according to the extent of their control over monetary poli
cy. The authors examined the relationship between the sta
tus of the central bank and inflation over the entire sample
period and four subsample periods—pre-World War I, the
Interwar Years, Bretton Woods and post-Bretton Woods.
Periods of hyperinflation, however, were excluded from the
data. In all sample periods, the countries with independent
central banks were in the low inflation group. Nevertheless,
some of the dependent central banks were also in this
group. The authors concluded that independence may be a
sufficient condition for low inflation but not a necessary
one.

Figure 3

Average Real GNP Growth: 1955-1987
Percent
* Japan

S p a in *

Australia/
Norway * Canada

" a 'v *

F ran ce

j . The Netherlands
.j. Germany
Denmark
?Sw eden
* United States
•I* Switzerland
•I* United Kingdom

n . i . i , i m i T<

New Z e a la n d *

2 “ I-------------------------1
------------------------- 1
------------------------- 1
------------------------- 1
----------------------

2
3
Index of Central Bank Independence

Source: Alesina and Summers (1993).

Summers (1993) likewise found no correlation
betw een average econom ic grow th or the varia
bility of grow th and the level of cen tral bank
independence (figures 3 and 4).12
De Long and Sum m ers (1992) looked at the
relationship betw een central bank independence
and output per w ork er while trying to eliminate
differences betw een countries that w ere due
solely to convergence effects.13 To do this, they
exam ined the grow th rate of real gross domes
tic product (GDP) per w ork er during 1 9 5 5 -9 0 ,
controlling for the level of GDP per w orker in
1955 .14 This procedure showed a positive rela1
2The results are the same if per capita gross national
product (GNP) is used rather than GNP.
,3Standard neoclassical growth models suggest that growth
rates of economies tend to converge over time. Thus given
two countries, the one with the lower per capita output will
have a higher growth rate than the other until their levels
of real output per capita converge.
14GDP per worker levels are based on the Summers and
Heston (1991) estimates, which use purchasing power
parity conversions.

tionship betw een central bank independence
and econom ic grow th.15 More precisely, they
found that holding constant the 1955 level of
real output per w orker, a unit increase in their
index of cen tral bank independence was as
sociated with a 0.4 percentage point increase in
grow th per y ear.16
In contrast, Cukierman, Kalaitzidakis, Summers
and W ebb (1993) found that output grow th in
industrialized countries was unrelated to central
bank independence even after controlling for
structural factors that might influence grow th.
The factors they considered w ere the initial level
15This study does not take into account that the degree of in
dependence of the central bank of New Zealand changed
dramatically in 1989. Furthermore, all of the studies, with the
exception of Alesina (1988), do not take into account that
there was an institutional change in the structure of the
Bank of Italy in 1981 that increased its independence. The
latter change, however, was not as substantial as the former.
16De Long and Summers regress the average growth rate of
GDP per worker over the period 1955-90 on GDP per worker
in 1955 and the central bank independence index.

JULY/AUGUST 1993

Figure 4

Variance of Real GNP Growth: 1955-1987
Percent
* Japan

•I* Spain
•I" Switzerland
-T h e Netherlands
^ Denmark
^.

New Zealand •b

. . .. * Belgium
Australia
.
United Kingdom / F ra n c e *
Canada
•I* Sweden

*!• Germany

United States

Norway

------------------------- 1
------------------------- 1
------------------------- 1
------------------------- 1------------------------

2
3
Index of Central Bank Independence

Source: Alesina and Summers (1993).

of a country’s GDP, its initial enrollm ent rates for
primary and secondary education, and changes
in its term s of trade. The authors did find,
however, using the turnover rate of central
bank governors as a proxy for independence,
that central bank independence did have a posi
tive effect on grow th in developing countries.
The difference in the results for industrialized
countries versus developing countries, they ar
gue, may imply that “dependence on political
authorities is bad for grow th only when the lev
el of independence is sufficiently high."17 Cen
tral bank independence is higher in all the
industrialized countries than in most of the de
veloping countries.

Central Bank Independence and
Fiscal Deficits
Another area of empirical study has been the
relationship betw een central bank independence
17See Cukierman, Kalaitzidakis, Summers and Webb (1993),
p. 42.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

and fiscal deficits. The motivation for these
studies is the belief that independent central
banks should be b etter able to resist govern
ment efforts to have them monetize deficits.
Thus governm ents realizing that th ere may be
some limit on their ability to issue bonds con
tinuously to finance deficits may decide to limit
deficit spending.
Parkin (1987) investigated this question for the
same 12 countries as Bade and Parkin for the
period 1 9 5 5 -8 3 .18 He found that there was some
evidence of a negative relationship betw een cen
tral bank independence and the long-run be
havior of governm ent deficits as a percent of
gross national product (GNP). The deficits of
Switzerland and Germany, the countries w ith the
highest levels of central bank independence, had
long-run equilibrium values n ear zero w ith little
variance. However, other countries, notably
France, that had low levels of central bank in18See Bade and Parkin (1985).

dependence also had small long-run deficits as a
p ercent of GNP.
Masciandaro and Tabellini (1988) looked at fis
cal deficits as a percent of GDP in Australia,
Canada, Japan, New Zealand and the United
States during the period 1 9 7 0 -8 5 .19 They found
that New Zealand, w hich had the lowest level of
central bank independence of the five countries
during this period, had the highest fiscal deficit
as a percent of GDP. The United States,
how ever, w ith the highest level of central bank
independence among this group of countries,
had a deficit/GDP ratio similar to those of the
other countries.
Grilli, Masciandaro and Tabellini (1991) found
that th ere was generally a negative correlation
betw een the deficit/GNP ratio and the degree of
central bank independence. However, if political
factors, as well as central bank independence,
w ere included in their regression, the latter
variable was insignificant.20 Thus they conclude
that an independent m onetary authority appar
ently does not discourage the governm ent from
running fiscal deficits.
A fu rth er exam ination of the relationship b e
tw een fiscal deficits and central bank indepen
dence, w hich is consistent with the w ork done
by Alesina and Sum m ers and De Long and Sum
m ers, is presented h ere.21 Using the same index
of central bank independence and the same 16
countries as these previous papers, th ere is
some evidence of a negative correlation b e
tw een average deficits as a percent of GDP and
central bank independence for the period
19 7 3 -8 9 , as shown in figure 5 .22 The degree of
independence, however, is not a statistically
significant (at a = .05) determ inant of the
deficit/GDP ratio. The variability of deficits as a
percent of GDP is also negatively correlated
w ith central bank independence (figure 6) and

this relationship is statistically significant.

EVALUATION OF THE EMPIRICAL
STUDIES
At first glance, these studies seem to indicate
that a country that wants to low er its inflation
rate and do so w ithout hurting grow th should
create an independent cen tral bank. Such a cen
tral bank apparently could also help reduce fis
cal deficits and increase output. These benefits
would explain the recen t popularity of indepen
dent central banks. Thus Grilli, M asciandaro
and Tabellini commented:
Having an independent central bank is almost
like having a fre e lunch; th e re are b enefits but
no apparent costs in term s o f m acroeconom ic
perfo rm ance.23

Alesina and Sum mers (1993) w ent a step fu rth er
in concluding their findings: "Most obviously
they suggest the econom ic perform ance m erits
of central bank independence.”24
A m ore careful analysis of these studies,
however, indicates w eaknesses that highlight
the need fo r fu rth er evidence before one
should believe that creating an independent cen
tral bank will improve a country's economic
perform ance. The following four weaknesses
are considered: 1) the difficulty in m easuring
central bank independence; 2) the possibility of
a spurious relationship betw een independence
and econom ic perform ance; 3) the possible en
dogeneity of central bank independence; and 4)
the inclusion of the fixed exchange rate period
in the sample data of some of the studies.
The m easures of central bank independence
used in em pirical studies have been determ ined
by establishing a set of factors thought to be
relevant for independence and then analyzing
central bank ch arters and laws for compliance
with these factors. W ith the exception of the in-

19The deficits are as a percent of GNP for Japan.

23See Grilli, Masciandaro and Tabellini (1991), p. 375.

20These political factors include the frequency of government
changes, significant changes in the government and the
percent of governments in a given period supported by a
single majority party.

24See Alesina and Summers (1993), p. 159. Even the press
has picked up the banner of central bank independence. A
recent headline in The Washington Post proclaimed: “ More
Independence Means Lower Inflation, Studies Show.” See
Berry (1993).

2 See Alesina and Summers (1993) and De Long and Sum
1
mers (1992).
22The 1989 ending date was chosen because of the change
in the status of the Bank of New Zealand, which occurred
in 1989. All data are from the International Monetary Fund,
International Financial Statistics.

JULY/AUGUST 1993

Figure 5

Average Deficit as a Percent of GDP: 1973-1989
Percent
12.5
Italy
10

•{• Belgium

7.5

New Zealand^
T

, .
* Japan
Spain +

2.5

AThe Netherlands
j,
t Canada
,
Sweden /
* United States
United Kingdom
Australia / France
.j. Germany

Norway*^

* Denmark
-I* Switzerland

2
3
Index of Central Bank Independence

Figure 6

Variance of Deficit as a Percent of GDP: 1973-1989
Percent

•I* Sweden

*1* Denmark
New Zealand •i'

*** Norway

•fr Belgium

United Kingdom .
Spain *
*
+ T h e Netherlands

Italy
Japan
A ustria*
Canada
+ France
0

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

j,United States
^.Germany
4* Switzerland

2
3
Index of Central Bank Independence

dex created by Cukierman, all of the indexes of
independence apply equal w eight to each factor.
For instance, the Grilli, M asciandaro and Tabelli
ni index based on political m easures of indepen
dence gives a country one point if no one on
the central bank board is appointed by the
governm ent and one point if the policy form u
lated by the central bank does not require ap
proval by the governm ent. Although the latter
certainly places a greater constraint on the ac
tions of the central bank than the form er, the
two are treated the same empirically.
Another concern is that the studies are based
on a legal m easure of independence that may
not reflect a bank's d e fa c t o level of indepen
dence. If th ere is a difference betw een legal and
practical independence, studies based on the
form er type of m easures may provide m islead
ing results. Cukierman (1992), in an attem pt to
address this possibility, uses central bankers'
responses to a questionnaire to determ ine the
actual degree of independence in the 1980s. He
finds that the correlation betw een the legal in
dex and this practical index of independence is
0.33 for developed countries, 0.06 for devel
oping countries and 0.04 overall.25 This finding
indicates, as Cukierman notes, that a legal index
of independence is not useful for studying de
veloping countries. It also indicates that a legal
index may be a weak m easure of actual in
dependence fo r the developed countries.
T h ere also may be bias in the factors selected
to m easure independence. For example, Grilli,
M asciandaro and Tabellini include: “statutory re
quirem ents that central bank pursues m onetary
stability amongst its goals” in their index.26 Like
wise, a central bank is m ore independent under
Cukierm an’s system if price stability is its only
objective than if price stability is one of a num
b er of objectives or not an objective at all. Us
ing the goal of price stability as a m easure of
central bank independence may result in a bias
betw een the m easure of independence and the
inflation rate.

Table 1

Comparison of Relative Rankings of
Central Bank Independence________
Alesina

Australia
Belgium
Canada
Denmark
France
Germany
Italy
Japan
Netherlands
New Zealand
Norway
Spain
Sweden
Switzerland
United Kingdom
United States

Alesina and
Summers

Cukierman

14
5
5
5
5
1
13
3
5
14
5
14
5
1
5
3

8
8
4
4
8
1
14
4
4
16
8
15
8
1
8
3

7
14
5
4
9
2
12
15
6
10
16
13
10
1
7
3

relative rankings as given by Alesina, Cukierman,
and Alesina and Sum m ers.27 All agree that Swit
zerland and Germany have the m ost independent
central banks of the countries studied. T h ere
are, however, a few countries w hich are ranked
quite differently by the authors. For example,
Japan has the second lowest level of indepen
dence of all 16 countries, according to Cukier
man, w hereas Alesina, and Alesina and Summers
give it a m uch higher level of independence.

The problem s in developing precise m easures
of central bank independence are less im portant,
however, if there is a consensus in ranking cen
tral banks within broad levels of independence.
Table 1 lists 16 OECD countries along with their

This discrepancy over the degree of indepen
dence of the Bank of Japan is not due solely to
differences in factors considered in m easuring
independence. The index used by Alesina is
based on the criteria of independence created
by Bade and Parkin (1985). The index used by
Alesina and Sum mers is constructed by averag
ing the indexes created by Alesina, and Grilli,
Masciandaro and Tabellini. Bade and Parkin
claim that the Bank of Japan is independent
from the governm ent in form ulating and im
plementing m onetary policy, and Grilli, M ascian
daro and Tabellini claim that th ere are no
provisions for handling policy conflicts betw een
the Bank of Japan and the governm ent. In con
trast, Cukierman claims that the Bank of Japan
and the governm ent form ulate policy jointly and

25The correlations are based on the weighted indexes.
Giving each factor related to independence an equal
weight in the indexes results in a correlation of 0.01 for de
veloped countries and 0.00 for developing countries.

27The measure of independence developed by Cukierman is
based on more factors than the measure used by Alesina,
and Alesina and Summers. Thus Cukierman’s rankings are
more delineated than the other two.

26See Grilli, Masciandaro and Tabellini (1991), p. 368.

JULY/AUGUST 1993

fu rth er notes that in the case of a policy con
flict, the executive bran ch of the governm ent
has final authority.28
Since most of the em pirical studies consider
only central bank independence as a d eter
m inant of econom ic perform ance, it is possible
that if oth er factors are accounted for, these
results could be spurious. Grilli, M asciandaro
and Tabellini attem pt to account for other fac
tors that could affect the rate of inflation by in
cluding political variables. They find that after
accounting fo r political factors, central bank in
dependence was still negatively related to infla
tion in the countries studied over the period
1 9 5 0 -8 9 . The incorporation of political variables
is a step in the right direction, but other factors
also should be considered. As noted by Cukier
man, “m onetary policy is generally sensitive to
shocks to governm ent revenues and expendi
tures, employment, and the balance of pay
m ents.”29 The types of shocks that a country
experienced over the sample period and the
reaction of the central bank to these shocks can
affect its econom ic perform ance. A study by
Johnson and Siklos (1992) found that the reac
tions of central banks (as m easured by changes
in interest rates) to shocks to unemploym ent, in
flation and w orld in terest rates w ere not closely
related to standard m easures of central bank in
dependence.
Empirical use of these indexes may be proble
matic if central bank independence is an en
dogenous variable in the sense that countries
with a com m itm ent to price stability may have a
g reater propensity fo r independent central
banks. If this is true, the m ere establishm ent of
an independent bank without a com m itm ent to
price stability will not bring inflation benefits to
a country. In fact, a public aversion to inflation
predates the establishm ent of many independent
central banks. This was true fo r the creation of
the Bundesbank and m ore recently with respect
to central banks in Chile and New Zealand. New
Zealand had one of the highest inflation rates of
all industrialized countries in the 1980s. In 1989
legislation was passed to increase the indepen
dence of its central bank substantially. This
28Aufricht (1961) reproduces the Bank of Japan charter and
subsequent changes in its governing regulations, which
support the conclusion reached by Cukierman.
29See Cukierman (1992), p. 438.
30See McCallum (1989), pp. 285-88, for an explanation of the
limitations on monetary policy under a fixed exchange rate
system.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

change is often credited with bringing inflation
down to near zero. Though the legislation c e r
tainly formalized the country’s com m itm ent to
price stability, New Zealand had succeeded in
reducing its inflation rate from nearly 16 p er
cent in 1987 to 6 percent b efore the creation of
an independent central bank.
In theory, the degree of independence of a
central bank should not be a determ inant of a
country’s inflation perform ance under a fixed
exchange rate system because m onetary policy
cannot be set exogenously.30 During the BrettonWoods era, it is not clear that any central bank
(with the possible exception of the U.S. Federal
Beserve) could be considered independent in the
sense of an ability to pursue an independent
m onetary policy.31 Thus the em pirical finding of
a negative relationship betw een independence
and inflation w hen the sample period extends
over both the Bretton Woods and post-Bretton
Woods eras may indicate a flaw in these studies.
To assess the effect of central bank indepen
dence on inflation, the data used in these
studies could be divided into tw o periods. If no
evidence of a relationship betw een indepen
dence and inflation is found in the Bretton
Woods period, this would strengthen the under
lying argum ent of these studies that central
bank independence is a prim ary determ inant of
a country’s inflation perform ance.32 If, how ever,
evidence is found of a relationship betw een cen
tral bank independence and inflation in the
Bretton W oods period, this would conflict with
theory and could indicate that the em pirical
findings are spurious.

THEORETICAL MODELS OF FIS
CAL AND MONETARY POLICY IN
TERACTIONS
In contrast to the em pirical studies, the theo
retical studies of cen tral bank independence and
econom ic perform ance concentrate on the con
flicts that can arise w hen m onetary and fiscal
policy are delegated to independent institutions.
In this literature an independent central bank is
one that does not cooperate with the fiscal au3 Indeed, the primary argument in favor of a flexible ex
1
change rate system was that such a system would permit
individual countries to pursue independent monetary poli
cies. See, for example, Friedman (1953) and Johnson
(1969).
32This is Grilli, Masciandaro and Tabellini’s finding (1991).

thorities in setting econom ic policy. A depen
dent central bank is one that cooperates with
the fiscal authority in setting policy.
In examining the theoretical implications of
central bank independence, this paper focuses
on models in w hich the policymaking process is
decentralized.33 The basic fram ew ork of these
models is as follows. The governm ent controls
fiscal policy, and the central bank controls
m onetary policy. Both parties set goals for the
econom y (generally inflation and output targets)
and assign priority to these goals. The goals and
priorities may differ across the policymakers.
Each institution uses the instrum ents available
to it in an attempt to reach its goals. In most
models the central bank controls the grow th
rate of the m onetary base and the governm ent
controls fiscal spending. T h ere is an underlying
model of the econom y that indicates how fiscal
and m onetary policy will affect the relevant eco
nomic variables. All of the models assume that
th ere are no stochastic shocks to the economy.
The governm ent and the central bank can
either cooperate in implem enting their policies
or choose not to cooperate. If they do not co
operate, they either can set policies simultane
ously, or one party can set its policies first and
the other then adopts its policies in reaction to
these.
Consider Andersen and Schneider’s (1986) sim
ple model in w hich the governm ent and the
central bank establish targets for inflation and
output.34 The fu rth er the actual level of output
and rate of inflation are from their respective
targets, the m ore disutility each authority r e
ceives. Thus, using the following equations,
each authority can be modeled as setting policy
to minimize its respective loss functions:35
(1) Lf = a J y - Vj)2 + bfln - nfl2
(2) L m = a m ( y - y m )2 + b m ( n - n m )z
V
J

af > bf
bm — am
>

(3) nf > nm, yf > y m
33There have been studies concentrating solely on monetary
policy that have shown that better economic outcomes
result from the policymaker placing a greater weight on in
flation than society as a whole. Rogoff (1985) argues that
these results indicate the economic benefits of central
bank independence. These studies ignore the interaction of
fiscal and monetary policy in determining economic out
comes and thus are not discussed here.
34Generally it is assumed that the government places more
weight on meeting its output target than its inflation target,
whereas the opposite holds for the central bank. Further

where:
L f is the fiscal authority’s loss function
L m is the m onetary authority’s loss function
y is output
n is inflation
yf is the fiscal authority’s output target
y m is the m onetary authority's output target
nf is the fiscal authority's inflation target
nm is the m onetary authority’s inflation target
a is the weight placed on the output target
b is the weight placed on the inflation target
Andersen and Schneider com pare the economic
outcomes under cooperation vs. noncooperation
given th ree different models of the economy.
The first model is Keynesian in nature. This is a
short-run model with price sluggishness so that
even anticipated changes in policy affect ag
gregate demand. The level of output and the
rate of inflation prevailing in the econom y are
affected by both fiscal and m onetary policies,
which can be shown in a simple reduced form
model with the following equations:
w here f is the fiscal policy instrum ent and m is
the m onetary policy instrum ent.36
( 4 ) y = r 0f + y i m

0< y,< y0

(5) n = dQ + #,m
f

0 < 6 0< 6 V

In the second model, which Andersen and
Schneider re fe r to as Keynesian-New Classical,
anticipated m onetary policy is neutral; it can af
fect only inflation. Thus in a world of certainty,
equation (4) becom es the following:
(V y = Y ,f
In the third model, the econom y is New Classical
in nature, characterized by p erfect price flexibi
lity and rational expectations. Anticipated policy,
both fiscal and m onetary, affects only inflation,
not output. T he econom y is modeled by the fol
lowing equations:
( 7) n = 1 7 /

r ) ,m

more, it is generally assumed that the inflation and output
targets set by the government are greater than or equal to
the targets set by the central bank.
35The quadratic nature of the loss functions, which is stan
dard in the macroeconomic game theory literature, implies
that deviations on either side of the targets produce an
equal loss to the policymaker.
36The restrictions in equations (4) and (5) imply that fiscal
policy has a greater (lesser) effect on output (inflation) than
does monetary policy.

JULY/AUGUST 1993

(8) y = 7i - tT
(9) n - ne = r j j f - f e) + r ] ^ m - m e),
w here y now refers to output relative to capaci
ty and the superscript e refers to the expecta
tion of the variable. Output can be increased
above capacity only through unanticipated infla
tion, and unanticipated inflation can occu r only
through unanticipated changes in fiscal policy,
m onetary policy or both.
The relevant issue for policy is the size of the
loss to each policym aker under cooperation and
noncooperation. Cooperation in the determ ina
tion of m onetary and fiscal policies is modeled
by the governm ent and the central bank choos
ing the policy variables (f and m) to minimize a
weighted average of their loss functions:
(10) min Lc = pLf + ( l - p ) L m

0<p<l

= p\afi y - V j f + b fln - t t/I
+ (1 - p )[a m( y - y j + b j n - n j ] ,
w h ere the w eight placed on each loss function
is determ ined by the relative bargaining strength
of the two parties. Solving this minimization
problem yields the equilibrium values for output
and inflation, w hich can be substituted into the
loss functions for the governm ent, equation (1),
and the central bank, equation (2), to determ ine
the loss to each.
As noted above, noncooperation can be model
ed in two ways. In the first, fiscal and m onetary
policies are chosen simultaneously; that is, the
governm ent selects a level of spending to mini
mize its loss function, equation (1), taking as
given the actions of the central bank. At the
same time, the central bank chooses the grow th
rate of the m onetary base to minimize its loss
function, equation (2), taking as given the actions
of the governm ent. This stru ctu re is referred to
as a Nash game and the resulting equilibrium is
called a Nash equilibrium. In a Nash equilibrium,
neither authority, taking the actions of the other
as given, can decrease its loss by unilaterally
changing its policy.
In the second model of noncooperation, one
policy is set b efore the other is determ ined.
This process is known as a Stackelberg game,
and the policym aker who moves first is known
37See Andersen and Schneider (1986), p. 188.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

as the Stackelberg leader, w hereas the other
policymaker is known as the Stackelberg follow
er. The leader chooses its policy, and the fol
lower sets its policy in reaction. Furtherm ore,
the leader, in choosing its policy, knows how
the follower will react.
Although the equilibrium level of output and
the rate of inflation vary depending on which
model of the econom y is used, in all three
models the cooperative solution is Pareto superi
or to the noncooperative solution. This result is
invariant to the stru ctu re of noncooperation—
Nash or Stackelberg. The perform ance of the
econom y is b etter under cooperation in the
sense that the losses to the governm ent and the
central bank are each low er than they are un
der noncooperation. This result holds even if
the governm ent and the central bank each place
the same weight on meeting their inflation tar
gets relative to their output targets (af = a m and
b f= b m but m aintain different targets.
)
Andersen and Schneider sum marize these
results by noting the following:
When we have two independent authorities
who act in their own selfish interest, then we
quite often observe a conflict over the "right”
policy direction. This result should be kept in
mind when quite often the argument is put for
ward that an independent monetary authority
should be created. ... Two independent policy
makers do not automatically guarantee a policy
outcome which is preferred to other outcomes
under different institutional solutions.37
Alesina and Tabellini (1987) show that adding
one m ore target to the loss functions of the
governm ent and the central bank also does not
change the nature of the results. Noncooperation
is once again suboptimal.
Adding a time dimension to the model also
does not change the basic result that coopera
tion can improve the outcom e from the p er
spective of both policym akers. Pindyck (1976)
presents one of the first dynamic models
analyzing the strategic interaction of m onetary
and fiscal policy. He argues that the
separation of monetary and fiscal control may
considerably limit the ability of either authority
to stabilize the economy, particularly when the
conflict over objectives is at all significant.38
Petit (1989) exam ines the issue of policy coor
dination in a continuous time model. The
38See Pindyck (1976), p. 239.

governm ent sets targets for output and inflation,
giving higher priority to output. The central
bank targets inflation and the level of in tern a
tional reserves, giving higher priority to infla
tion.39 As is standard, the governm ent sets the
level of public expenditures to minimize its loss
function, w hereas the central bank sets the
grow th of the m onetary base to minimize its
loss function.
In this model, policies are set at the beginning
and are unchanged over the period considered.
Once again, cooperation is Pareto superior to the
Nash and the Stackelberg equilibriums. F u rth er
more, cooperation in this dynamic system leads
to a decrease in the variability of the targets
(particularly prices and international reserves),
and raises the speed of adjustm ent of the sys
tem. The latter indicates that, given a shock to
the system, the econom y will retu rn more
quickly to its long-run values of output and in
flation if the governm ent and the central bank
are coordinating their policies. Thus Petit con
cludes that policym akers should coordinate their
policies.40
Other studies con cen trate on the interaction
of the governm ent and the central bank in
financing fiscal deficits w here the deficit must
be financed through bonds, seignorage or
both.41 Under the assumption that th ere is some
limit on the ability of a governm ent to continu
ally issue bonds to finance its deficit, the need
fo r inflation revenues becom es im portant.42 Sar
gent and W allace (1981) conducted the seminal
research on this question and showed that if
the governm ent em barks on a path of unsus
tainable deficits, the central bank might eventu
ally be forced to inflate to fund the deficits. If
the public realizes that the governm ent debt is
on such a path, it will expect inflation to in
crease, w hich may cause inflation to increase
39The target for international reserves reflects a balance of
payments objective.
40Hughes Hallett and Petit (1990) also model the interaction
of fiscal and monetary policy in a dynamic setting, reach
ing this same conclusion.
4 Seignorage is the revenue received from the creation of
1
money. It occurs because base money costs only a fraction
of its face value to produce.
42As the public debt grows, there may be increasing concern
among bondholders that the government will be unable to
repay the bonds.
43As Sargent and Wallace note, if money demand today de
pends on inflationary expectations, then the price level to
day is a function of not only the current money supply, but
also expectations of the future levels of the money supply.

well b efore the debt limit has been reached.43
This outcom e is a result of the governm ent b e
ing able to set its policies and the central bank
having to react to those policies (a Stackelberg
game).44
In general, a conflict over the public debt can
arise at any time w hen the governm ent and the
central bank are allowed to adopt independent
policies. Tabellini (1986) develops a dynamic
model in w hich the central bank sets targets for
changes in the m onetary base and the stock of
outstanding public debt while the governm ent
sets targets for the fiscal deficit net of interest
payments and the stock of outstanding public
debt. The target value of public debt is the
same for both authorities. In choosing the level
of the m onetary base and the fiscal deficits, the
two authorities are constrained by the govern
ment's dynamic budget constraint.45 The stock
of public debt as a proportion of income is con
sidered too high by both the fiscal and m one
tary authorities. In the noncooperative setting,
however, each authority ignores the benefit to
the other of its own actions to reduce the level
of debt. In the cooperative setting these benefits
are internalized, resulting in a lower level of
debt.
Tabellini (1987) and Loewy (1988) provide two
m ore examples of models examining the conflict
betw een central banks and governm ents over
fiscal policy. Both show that such a conflict can
lead to an increase in governm ent debt. As not
ed by Blackburn and Christensen (1989), a con
flict will always arise betw een a central bank
whose goal is to maintain price stability and a
governm ent whose objective is to increase out
put and is pursuing this goal by running a
stream of large deficits. Such a m acroeconom ic
program is infeasible; one party will have to re
vise its strategy (give in). The conflict creates
44The concern that undisciplined fiscal policies could result
in inflation was recognized by the EC in drafting the Treaty
on European Monetary Union. In the regulations concern
ing the proposed European Central Bank, the bank is pro
hibited from financing fiscal deficits of the member
countries.
As pointed out by Sargent and Wallace, and expounded
on by Darby (1984), the need for the central bank to mone
tize government debt through an inflationary policy is
based on the assumption that the rate of growth of the real
economy is less than the real rate of interest.
45Note that monetary base and fiscal deficits in this model
are both instruments and targets.

JULY/AUGUST 1993

problem s for the econom y because of the un
certainty over the future course of policy: the
public can expect higher inflation or higher tax
es, depending on w hich policym aker gives in.46

EVALUATION OF THE
THEORETICAL LITERATURE
The theoretical studies indicate that noncoor
dination of fiscal and m onetary policies will
result in a suboptimal econom ic perform ance
from the perspective of both the governm ent
and the central bank. Policy targets are m ore
closely met w hen coordination occurs. Thus an
independent central bank is not conducive to
achieving b etter policy outcomes.
However, the theoretical w ork, like the em pir
ical studies, has its weaknesses. One criticism is
that the models are too simplistic. Neither the
p referen ce structures of the two authorities,
nor the models of the economy, are completely
specified. Furtherm ore, most of the models
operate in a world o f certainty. Policy, however,
is not made in a world of certainty. Extrinsic
u ncertainty—shocks to the econom y—can drive
a wedge betw een the im plem entation of policy
and its outcom e. Intrinsic u ncertainty—lack of
knowledge of the p referen ces of a policym aker—
is incorporated only in Tabellini and Loew y’s
models.47 As these two models illustrate, adding
uncertainty can increase the policy conflict b e
tw een an independent central bank and fiscal
authority.
In addition to assuming certainty, the models
also omit one im portant player in these policy
gam es—the public. Public perception of the
credibility of a m acroeconom ic program is im
portant to its results because the public can
limit the ability of policym akers to take advan
tage of an inflation/output tradeoff. If an in
dependent central bank can increase the public
perception of the credibility of policy, this in
turn should produce b etter econom ic results.48
Another deficiency of this literature is its
failure to address the feasibility of the policymak
e rs’ goals. The output goals set by the govern
46A government may adopt a strategy of running deficits,
through decreasing taxes, to force future governments to
cut expenditures. Under this strategy, the government
would prefer an independent central bank, which will re
fuse to monetize the deficits and thereby increase the
likelihood that fiscal spending will be reduced. See Sargent
(1985) for a discussion of this type of strategy.

FEDERAL RESERVE BANK OF ST. LOUIS

m ent, fo r example, may not be sustainable
w ithout accelerating inflation. Tax and expendi
tures plans, which lead to a stream of deficits,
may also raise questions about the sustainability
of fiscal policy. In this environm ent, an indepen
dent central bank could be useful if its credible
com m itm ent to price stability forced the govern
ment to evaluate the sustainability of its policy
goals. In contrast, centralization of policies
might reduce the long-run econom ic p erfo r
m ance of a country w hen the governm ent’s fo
cus is short-run perform ance.

CENTRAL BANK INDEPENDENCE
AND THE ECONOMY— W H AT DO
WE KNOW?
This paper began with two questions: W hy is
the idea of an independent central bank as pop
ular as it is? Are th ere econom ic benefits to be
gained from having an independent central bank?
Unfortunately, the em pirical and theoretical
studies surveyed do not provide clear answ ers.
The em pirical studies find that th ere is a nega
tive correlation betw een central bank indepen
dence and long-run average inflation. They also
show a negative correlation betw een indepen
dence and long-run average governm ent deficits
as a percent of GDP. In general, they find no
evidence of a positive correlation betw een out
put grow th and central bank independence.
These results all point in the same direction yet
do not provide unequivocal evidence that an in
dependent central bank will low er inflation and
governm ent deficits and raise a cou ntry’s
output.
In sum, these empirical studies provide evi
dence of a negative correlation betw een central
bank independence and inflation and central
bank independence and fiscal deficits, but they
do not provide evidence of causality. Countries
with an aversion to inflation may formalize this
aversion through the creation of an independent
central bank. If this is true, it is the inflation
aversion, not the independence of the central
bank, that is the prim ary causal factor behind
the low inflation result. The empirical m easures
them selves are biased tow ard the finding that
47See Tabellini (1987) and Loewy (1988). In Tabellini’s model
the government is initially unaware of the preferences of
the central bank. In Loewy’s model both parties are initially
unaware of the preferences of the other.
48This issue has been studied in the literature that focuses
only on monetary policy. See Blackburn and Christensen
(1989) for a survey of this literature.

independence prom otes low inflation. This is b e
cause the m easures place m uch weight on legal
requirem ents that a central bank pursue price
stability and place this goal above all others.
Cukierman is explicit in stating that his m easure
of independence:
is not the independence to do anything that the
central bank pleases. It is ra th er the ability of
the bank to stick to the price stability objective
even at the cost o f oth er short-term real ob jec
tives.49

Given such a definition of independence, it is
not surprising that independence is equated
with low inflation.
Theoretical studies indicate that an indepen
dent central bank can increase policy conflicts
with the governm ent w henever the preferen ces
of the two differ and, in so doing, w orsen the
econom ic perform ance of a country. These
studies, how ever, do not provide overwhelming
support for the idea that countries should place
m onetary policy in the hands of the executive or
legislative branches of governm ent. The simple
stru ctu re of these models ignores some factors
that affect the outcom e of policy decisions—
for example, the role of the public and the
overall credibility of policy. Central bank in
dependence may enhance credibility and thus
the overall effectiveness of a policy program.
In sum then, in the em pirical studies, em pha
sis on price stability and freedom to pursue this
goal are prim ary determ inants of independence.
In the theoretical studies independence is eq
uated with noncooperation betw een the fiscal
and m onetary authorities in policy im plem enta
tion. These different definitions of indepen
dence may partly explain the d ifferent results.
Fu rtherm ore, countries that may be classified as
independent using the em pirical definition may
be classified as dependent using the theoretical
definition. New Zealand is one such example.
T he 1989 Reserve Bank of New Zealand Act
made price stability the only goal of the central
bank, and the central bank is free to adopt poli
cies to achieve that goal. Thus according to the
em pirical definition of independence, the 1989
act created an independent central bank in New
Zealand. The central bank’s inflation target,
however, is established by the governm ent fo r a
multi-year period. The governor of the central
bank signs an agreem ent pledging the bank to
adopt policies to m eet this target. Such coopera

tion betw een the m onetary and fiscal policy
m akers is consistent with a dependent central
bank in the theoretical models.
Altogether these studies indicate that we are
far from fully understanding the role of central
bank independence in producing favorable eco
nomic outcomes.

REFERENCES
Alesina, Alberto. “ Macroeconomics and Politics,” in Stanley
Fischer, ed. NBER Macroeconomics Annual (MIT Press,
1988), pp. 13-52.
, and Lawrence H. Summers. “ Central Bank In
dependence and Macroeconomic Performance: Some
Comparative Evidence,” Journal of Money, Credit and Bank
ing (May 1993), pp. 151-62.
_______ , and Guido Tabellini. “ Rules and Discretion with
Noncoordinated Monetary and Fiscal Policies,” Economic
Inquiry (October 1987), pp. 619-30.
Andersen, Torben M., and Friedrich Schneider. “ Coordination
of Fiscal and Monetary Policy Under Different Institutional
Arrangements,” European Journal of Political Economy
(February 1986), pp. 169-91.
Aufricht, Hans. Central Banking Legislation (International
Monetary Fund, 1961).
Bade, Robert, and Michael Parkin. “ Central Bank Laws and
Monetary Policy,” unpublished manuscript, University of
Western Ontario, 1985.
Berry, John M. “ More Independence Means Lower Inflation,
Studies Show,” The Washington Post (February 17, 1993).
Blackburn, Keith, and Michael Christensen. “ Monetary Policy
and Policy Credibility: Theories and Evidence,” Journal of
Economic Literature (March 1989), pp. 1-45.
Buchanan, James M., and Richard E. Wagner. Democracy in
Deficit (Academic Press, 1977).
Capie, Forrest H., Terence C. Mills, and Geoffrey E. Wood.
“ Central Bank Dependence and Inflation Performance: An
Exploratory Data Analysis,” Centre for the Study of Monetary
History, City University Business School, Discussion Paper
No. 34 (March 1992).
Committee of Governors of the Central Banks of the Member
States of the European Economic Community. Annual
Report 1992 (April 1993).
Cukierman, Alex. Central Bank Strategy, Credibility, and In
dependence: Theory and Evidence (MIT Press, 1992).
_______, Steven B. Webb, and Bilin Neyapti. “ Measuring
the Independence of Central Banks and Its Effect on Policy
Outcomes,” The World Bank Economic Review (September
1992), pp. 353-98.
_______ , Pantelis Kalaitzidakis, Lawrence H. Summers, and
Steven B. Webb. “ Central Bank Independence, Growth, In
vestment, and Real Rate,” Carnegie-Rochester Conference
Series on Public Policy (autumn 1993).
Darby, Michael R. “ Some Pleasant Monetarist Arithmetic,”
Federal Reserve Bank of Minneapolis Quarterly Review
(spring 1984), pp. 15-20.
De Long, J. Bradford, and Lawrence H. Summers. “ Macroeconomic Policy and Long-Run Growth,” Federal Reserve
Bank of Kansas City, Economic Review (fourth quarter
1992), pp. 5-30.

49See Cukierman (1992), p. 370.

JULY/AUGUST 1993

Friedman, Milton. “ The Case for Flexible Exchange Rates,”
in Milton Friedman, ed. Essays in Positive Economics
(University of Chicago Press, 1953), pp. 157-203.

Petit, Maria Luisa. “ Fiscal and Monetary Policy Co-ordination: A
Differential Game Approach,” Journal of Applied Economet
rics (April-June 1989), pp. 161-79.

Grilli, Vittorio, Donato Masciandaro, and Guido Tabellini. “ Po
litical and Monetary Institutions and Public Financial Poli
cies in the Industrial Countries,” Economic Policy 13
(October 1991), pp. 341-92.

Pindyck, Robert S. “ The Cost of Conflicting Objectives in
Policy Formulation,” Annals of Economic and Social Meas
urement (May 1976), pp. 239-48.

Hughes Hallett, Andrew and Maria Luisa Petit. “ Cohabitation
or Forced Marriage? A Study of the Costs of Failing to
Coordinate Fiscal and Monetary Policies,” Weltwirtschaftliches Archiv (1990), pp. 662-89.
Johnson, David R., and Pierre L. Siklos. “ Empirical Evidence
on the Independence of Central Banks,” unpublished
manuscript, Wilfrid Laurier University (March 1992).
Johnson, Harry G. “ The Case for Flexible Exchange Rates,
1969,” this Review (June 1969), pp. 12-24.
Loewy, Michael B. “ Reaganomics and Reputation Revisited,”
Economic Inquiry (April 1988), pp. 253-63.
Masciandaro, Donato and Guido Tabellini. “ Monetary Regimes
and Fiscal Deficits: A Comparative Analysis,” in H. Cheng,
ed. Monetary Policy in the Pacific Basin Countries (Kluwer
Academic Publishers, 1988), pp. 125-52.
McCallum, Bennett T. Monetary Economics: Theory and Policy
(MacMillan, 1989).
Parkin, Michael. “ Domestic Monetary Institutions and
Deficits,” in J. M. Buchanan et al., eds. Deficits (Basil
Blackwell, 1987), pp. 310-37.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

Rogoff, Kenneth. “ The Optimal Degree of Commitment to an
Intermediate Monetary Target,” Quarterly Journal of Eco
nomics (November 1985), pp. 1169-90.
Sargent, Thomas J. “ Reaganomics and Credibility,” in Alberto
Ando et al, eds., Monetary Policy in Our Times (MIT Press,
1985), pp. 235-52.
Sargent, Thomas J., and Neil Wallace. “ Some Unpleasant
Monetarist Arithmetic,” Federal Reserve Bank of Minneapo
lis, Quarterly Review (fall 1981), pp. 1-17.
Summers, Robert, and Alan Heston. “ The Penn World Table
(Mark 5): An Expanded Set of International Comparisons,
1950-88,” Quarterly Journal of Economics (May 1991),
pp. 327-68.
Tabellini, Guido. “ Money, Debt and Deficits in a Dynamic
Game,” Journal of Economic Dynamics and Control
(July 1986), pp. 427-42.
. “ Central Bank Reputation and the Monetization of
Deficits: The 1981 Italian Monetary Reform,” Economic In
quiry (April 1987), pp. 185-200.

Michael J. D uek er
Michael J. Dueker is an economist at the Federal Reserve
Bank of St. Louis. Richard I. Jako provided research
assistance.

Hypothesis Testing with Near-Unit
Roots: The Case o f Long-Run
Purchasing-Power Parity

' I HE HYPOTHESIS THAT the purchasing
pow er of a given currency, like the dollar, will
be equal across countries has strong appeal. If
the hypothesis is true, then inflation and ex
change rate m ovements will be such that a given
cu rren cy will, over time, lose equal am ounts of
its purchasing pow er in all countries. The se
quence of events by w hich deviations from
purchasing-pow er parity would be eliminated
can best be illustrated by example: If the dollar
could purchase m ore goods in other countries
than in the United States, then U.S. consum ers
would purchase m ore goods from abroad,
w hich would raise the demand for foreign cu r
rencies relative to the dollar and lead to a
depreciation of the dollar and eventual equaliza
tion o f the dollar’s purchasing pow er across
countries. Despite the intuitive appeal of such
argum ents for long-run purchasing-pow er pari
ty, statistical tests have been mixed. This paper
argues that previous test results have conflicted
because tests of purchasing-pow er parity have
relatively low pow er under both the null hypoth
esis that it holds, and the null that it fails. Hence
this paper contains tests of both null hypotheses
and shows that frequently neither is rejected
for monthly data from five m ajor industrialized
countries. This result serves as a caution against
testing only one null hypothesis, finding that the
null hypothesis cannot be rejected for a broad

set of countries and concluding that there is
robust evidence for o r against the theory of
long-run purchasing-pow er parity.
The theory of long-run purchasing-pow er par
ity (PPP) implies that a currency's purchasing
pow er is equal across countries in long-run
equilibrium, but does not specify how long devi
ations from this equilibrium can last. Large and
persistent departures from PPP in the last 20
years, how ever, have cast doubt on the validity
of PPP. As we will discuss later, there is a liter
ature w hich tests w hether long-run PPP holds,
that is, w h eth er departures from PPP are transi
tory. This article aims to reconcile some of the
disparate results from previous studies by using
a long-memory model, w hich can do m ore than
classify deviations from PPP as tem porary or
perm anent: it can provide specific m easures of
their persistence. Such m easures are useful b e
cause large, persistent differences in a cu rren
cy’s purchasing pow er across countries can
greatly affect trade flows and the allocation of
resources.
Empirically, long-run PPP holds if the real ex
change rate, w hich equals the nominal exchange
rate multiplied by the ratio of the domestic and
foreign price levels, is m ean-reverting. This arti
cle will conform with the m ajority of the em pir
ical PPP studies by using consum er price indexes

JULY/AUGUST 1993

to calculate the real exchange rate. If price in
dexes m easured the prices of identical baskets
of goods across countries, absolute PPP would
hold if P* = S x P , w here P is the domestic
price of the goods basket, P* is the foreign
price and S is the exchange rate in term s of
units of foreign cu rren cy per unit of domestic
currency. Because consum er price indexes do
not m easure the cost of identical baskets of
goods across countries, how ever, relative PPP
modifies the relationship to account for the ra
tio of the values of the tw o distinct baskets of
goods: P* = kSP, w here k is the ratio of the
value of the foreign basket to the domestic
basket. The domestic country’s real exchange
rate with the foreign country is then 1/k and
equals SP/P*.1
The conventional approach to testing for longrun purchasing-pow er parity consists of testing
for a unit root in the real exchange rate: Longru n PPP holds if the real exchange rate is meanreverting but not if it has a unit root. Previous
tests have shown little pow er to reject w hichever
of the two null hypotheses is employed. Tests
w hose null hypothesis is that the real exchange
rate contains a unit root generally fail to reject,
w hereas tests w hose null is that long-run PPP
holds also often fail to reject. These disparate
findings are reconciled, however, if th ere is
long m em ory in the real exchange rate, which
enables both acceptance and rejection of longrun PPP at conventional significance levels.2
This paper employs long-memory models to
obtain estim ates of the orders of integration of
real exchange rates on a continuous scale. The
advantage of estim ating the order of integration
on a continuous scale is that we can confirm
that long-memory time series behavior in real
exchange rates is a possible source of the dis
crepancies betw een previous tests of long-run

'Summers and Heston (1991) tabulate the costs of nearlyidentical baskets of goods across countries, rather than
use existing price indexes. They define the PPP nominal
exchange rate to be P*/P and use this implied exchange
rate, rather than the market exchange rate, to make cross
country comparisons. The Summers and Heston measures
of the price levels could take the place of commonly used
consumer price indexes when testing long-run PPP, as
could wholesale price indexes. The Summers and Heston
data, however, are only available on an annual basis and
include data extrapolated between five year data collection
periods. The analysis in this paper will be limited to the
use of consumer price indexes to facilitate comparison with
previous studies.
2Long memory, as will be discussed later, means that
the order of integration of a time series process is greater

FEDERAL RESERVE BANK OF ST. LOUIS

PPP. The finding of long m em ory in real ex
change rates also allows us to judge w hether
the real exchange rate reverts to its m ean within
an econom ically meaningful tim e fram e.

W HY PURCHASING-POWER
PAR ITY MIGHT NOT HOLD
Before discussing statistical tests of PPP, it is
w orth repeating Engel’s (1992) list of possible
reasons fo r the em pirical failure of PPP:
1. Barriers to trade such as tariffs and tran s
portation costs.
2. D ifferent consumption preferen ces across
countries.
3. The presence of non-traded goods in price
indexes.
4. Prices which are sticky in term s of the cu r
rency in w hich the good is consumed.
B arriers to trade, such as tariffs, are an obvious
reason why the same goods do not sell at the
same price throughout the world. D ifferent con
sumption p referen ces, on the oth er hand, would
lead consum ers in each country to choose
different baskets of goods. Because price indexes
are constructed for baskets of goods designed
to rep resent a particular cou ntry’s consumption,
an apparent failure of PPP could be due to
different rates of price inflation across two
country’s distinctive baskets of consumption
goods, rath er than different prices fo r the same
goods across countries. W hen included in price
indexes, non-traded goods can also muddle the
interpretation of the real exchange rate, because
non-traded goods can be idiosyncratic and are
thus not directly com parable across countries.
Nevertheless consum er price indexes will be
used in this paper, despite the presence of nontradeables, because wholesale price indexes can
fail to reflect the underlying rate of inflation ac
curately.3 The fourth source of failure, sticky

than zero. If the order of integration is greater than 0.5,
the series is not covariance stationary and if the order of
integration is greater than one, the series does not have a
mean. Long memory is not the same as an autoregressive
near-unit root, because a series with a near-unit au
toregressive root is still integrated of order zero, and is not
considered a long-memory process.
3For example the wholesale price index for Japan suggests
that Japan has had deflation on average from 1980 to the
present, whereas the GDP deflator and CPI show moderate
inflation.

prices, can best be explained by an example:
Japanese autos sold in Japan and also exported
to the United States have sticky prices in yen
w hen sold in Japan and sticky prices in dollars
when sold in the United States. Any exchange
rate fluctuations would cause the yen (or dollar)
price of the same model of ca r to differ across
the Pacific. Thus autos might contribute to the
failure of PPP in the true sense: the same good
being sold at d ifferent prices (net of taxes)
across countries.

PREVIOUS TESTS OF
PURCHASING-POWER PARITY
Tests of PPP in the literature can be classified
according to many criteria. In this b rief review
of a large literature, three features will receive
attention: 20th century annual data vs. post-1973
monthly data; the use of consum er price index
es versus wholesale price indexes in the calcula
tion of the real exchange rate; and w hether or
not price levels are assumed to be measured
with e rro r.4 The aim of this review is to illus
trate the lack of consensus that has em erged
from studies of long-run PPP and identify which
modeling choices might have influenced the out
com es of those tests.
Coughlin and Koedijk (1990) conduct unit-root
tests on real exchange rates using post-1973
monthly data and consum er prices and find that
the unit-root null hypothesis cannot be rejected.
They also exam ine w h eth er the real exchange
rates are cointegrated with factors thought to
determ ine the real exchange rate. Cheung and
Lai (1993b) use post-1973 monthly data on con
sum er prices and allow for m easurem ent erro r
in prices. They use a Johansen (1991) likelihood
ratio test for cointegrating vectors, in which
long-run PPP is the null hypothesis, and general
ly fail to reject long-run PPP.5 Thus the studies
o f Coughlin and Koedijk (1990) and Cheung and
Lai (1993b) illustrate the im portance of the null
hypothesis in testing long-run PPP. Edison and
Fisher (1991) and Fisher and Park (1991)
represent another pair of studies that differ in
the null hypothesis employed and the general
conclusions about long-run PPP. Cheung and
4For a thorough introduction to what the authors call the
purchasing-power parity assumption, see Caves, Frankel
and Jones (1990).

Lai (1993a) study 20th century annual data and
find evidence that the real exchange rate has
long memory, but not a unit root for most
countries studied. Pippenger (1993) uses w hole
sale price indexes and finds evidence that PPP
betw een Switzerland and various countries ap
pears to hold in the long run.
Overall a lack of consensus em erges from em
pirical tests of long-run PPP. The choice of null
hypothesis is one source of discrepancy, and it
appears that results from a long-memory model
can reconcile the apparently conflicting results
o f several previous studies of long-run PPP, which
differ prim arily in their choice of null hypothesis.
Consequently, tests of each null hypothesis will
be highlighted in the estim ation results below.

TESTING PURCHASING-POWER
PARITY
Much research on w hether PPP holds in the
long run consists of perform ing a unit-root test
on the real exchange rate. It is well known,
however, that unit-root tests, especially those
having a unit root as the null hypothesis, like
Dickey-Fuller, have little power against longmemory alternatives. Such unit-root testing con
sists of classifying econom ic variables as either
integrated of order zero [1(0)1 o r one [1(1)1.6 In
contrast, we use a param etric long-memory
model in w hich data series, like the real ex
change rate, are modeled as integrated of order
d, denoted 1(d), w here d does not have to be an
integer. Any series that is integrated of order
d < l will retu rn eventually to its mean (or its
determ inistic trend), so shocks to the real ex
change rate are not perm anent if the real ex
change rate is integrated of order d < l .
This paper also provides inform ation about
the sources of PPP failure by examining the
com ponents of the real exchange rate. T he ratio
of the price levels may have a higher order of
fractional integration than the nominal exchange
rate, or vice versa. If r is the real exchange
rate, s is the nominal exchange rate, p is the
domestic price level, and p* is the foreign price
level (all in natural logs), then r = s + (p-p’ ). If s is
I(dl), (p-p *) is I(d2), then r will generally be in6lntegration of order one means that a variable’s first differ
ences are stationary, whereas its levels are not.

5Provided that the hypothesis of zero cointegrating vectors
has been rejected, the null hypothesis for subsequent
hypothesis tests using the Johansen procedure is that
there is at least one cointegrating vector.

JULY/AUGUST 1993

tegrated of order m a x {d l,d 2 }. If r is 1(b) w here
b < max {d1 ,d 2 }, then the nominal exchange
rate and the price ratio (p-p*) are fractionally
cointegrated, that is, they share the same
stochastic trend to some exten t.7 T he real ex
change rate does not have a unit root if b < l .
W e can also exam ine the point estim ates of d l
and d2 and see w hich com ponent appears to
have the strongest trend. This com parison an
sw ers critics of flexible exchange rates who ar
gue that floating rates actually have caused the
real exchange rate to be less stable than it would
have been under fixed nominal exchange rates.8
If d l > d 2 , then shocks to the nominal exchange
rate are m ore persistent than shocks to the
relative price levels. The latter is som ewhat cu
rious, because proponents of the switch to a
flexible exchange rate regime envisioned flexible
exchange rates as sources of real exchange rate
stability in a world in w hich countries might
have persistent d ifferences in inflation rates. Yet
if shocks to the nominal exchange rate are m ore
persistent than shocks to the relative price levels,
then the nom inal exchange rate has persistence
above w hat is potentially useful in reducing the
variance of the real exchange rate. In the em
pirical results that follow, the possibility of ex
cess persistence in the nominal exchange rate
will be examined.

BACKGROUND ON LONGMEMORY MODELS
For many time series, autoregressive movingaverage (ARMA) models serve as a parsimonious
way to sum m arize the autocovariance stru ctu re
of the data. One limitation of such models is that
ARMA processes are integrated of order zero,
and the autocovariances die o ff relatively quickly,
even when a root in the autoregressive polyno
mial is n ear one. Thus ARMA models can be
called short-m em ory models, because a shock
affects the level of the series fo r a relatively
short time.
Long-memory models, in contrast, are suitable
for data that have slowly decaying coefficients
7For reasons outlined below, the restriction that the coeffi
cients on s and (p-p*) equal one is relaxed, so that the
order of integration of a general linear (cointegrating) com
bination of s, p and p* is assumed to be the order of in
tegration of the real exchange rate. The concept of
cointegration has been generalized [Granger (1986)] to in
clude cases in which series have stochastic trends that
only partially offset each other. This is called fractional
cointegration. Originally, cointegration meant that a partic
FEDERAL RESERVE BANK OF ST. LOUIS

in their moving-average representations. The
fractional ARMA model can serve as a longm em ory model, yet it adds only one param eter
to a standard ARMA model. To illustrate, we b e
gin with a simple ARMA(1,1) applied to the first
difference of a data series y, w here L is the lag
operator, £ is a m ean-zero disturbance, p is the
AR coefficient, and 0 is the MA coefficient:
(1) (1 - pL )(l - L)y = ( 1 + d L k
A fractional ARMA model is simply an ARMA
model applied to fractionally differenced data:
(2) (1 - pL )(l - L)dy = (1 + 0L)£
The fractional differencing operator is evaluated
by taking a Taylor series expansion around
L = 0:9
(3) (1 - L P = 1 - d L + d ( d - l ) L 2 2

d i d - l ) ( d - 2 ) L 3 + ...
3!
Two characteristics of fractionally integrated
data are w orth noting. First, a series that is in
tegrated of order d (1(d)) with d < l reverts to
its mean (or at least to its determ inistic trend).
Second, if d < .5 , the series is covariance station
ary. At first glance, it might seem counter-intui
tive that a m ean-reverting series can fail to be
covariance stationary. W ith long m em ory, how
ever, the departures from the m ean can be
sufficiently persistent that the variance of the
series is infinite.
Furtherm ore, two commonly assumed datagenerating processes fit within the subset of
fractional integration: trend and difference stationarity. Fractional integration offers a bridge
betw een the controversial assignment of a data
series as either trend or difference stationary,
so that questions about stationarity assumptions
ular linear combination of two strongly trending series was
1 0 ).
(

8For example, Aliber (1993) notes that “ the U.S. dollar ap
preciated from 1979 to 1985 even though the U.S. inflation
rate was higher than the inflation rates in Germany and
Japan.”
9The concept of fractional differencing was developed by
Granger and Joyeux (1980) and Hosking (1981).

may be avoided. For example, if y is trend sta
tionary, we might model y as

the real exchange rates. The general form of
the fractional ARMA model is

(4) y, = M + £,
f

(8) A(L)(2 - L)d t = B(L)et
y

In differences, equation (4) looks like

w here y is 1(d) and t is assumed to have zero
mean, no serial correlation and variance a 2. A(L)
is an autoregressive polynomial of order p and
B(L) is a moving-average polynomial of order q:

(5) (2 - L ) y t = fi + ( l- L ) £ ,
If y is difference stationary, then
(6) ( 1 - L ) y t = /i + £(
Now suppose that y is fractionally integrated of
order d. The first differences of y are then
equal to
(7) (1 - L ) y t = ju + ( 1 - L Y X
w here fl is proportional to )u. Clearly trend stationarity (d = 0) and d ifference stationarity ( d = l )
are bridged by fractional integration, which al
lows for interm ediate cases. An intuitive way to
understand why fractional integration is an in
term ediate case betw een trend and difference
stationarity is to interpret each shock in a
difference-stationary process to be a perm anent
shift away from any previous trend; a shock in
a trend-stationary process is a short-lasting shift
from the trend; a shock in a fractionally inte
grated process is a long-lasting shift from the
trend. This paper uses estim ates of the order of
fractional integration to discrim inate, if possible,
betw een long-lasting shifts from the mean real
exchange rate and perm anent shifts in the real
exchange rate (the unit-root case).

(9) A(L) = l - p ^ L - p 2L 2- . . . - p L P
(10) B(L) = 1 + 0,L + 0,L2 + ... + Q Li
Estimation was carried out using the Fox and
Taqqu (1986) frequency-dom ain estim ator of
fractional ARMA models. The estim ator is based
on an approxim ation to the likelihood. Dahlhaus
(1988, 1989) has analyzed the Fox and Taqqu es
tim ator and has shown that it shares the same
asymptotic efficiency as exact maximumlikelihood estimation. Fu rth er details regarding
the estim ator appear in the appendix.
Before presenting estim ation results, we must
discuss how the real exchange rate was calculat
ed. Any m ism easurem ent of the price levels can
lead to spurious changes in the mean of the
real exchange rate and bias tests toward reje c
tion of long-run PPP. To minimize the possibility
of spurious rejections of long-run PPP, the real
exchange rate was calculated by estimating a
fractionally cointegrating relationship betw een
the nominal exchange rate (s), the domestic
price level (p) and the foreign price level (p*):11
( 11) s ~ a 0- a p t + aj>*t + £ ,

ESTIMATES OF LONG MEMORY IN
REAL EXCHANGE RATES
The data used in this article consist of 234
monthly observations of the nominal exchange
rates and consum er price indexes for the Unit
ed States, Great Britain, Germany and Japan
from June 1973 to November 1 9 9 2 .10 Thus six
bilateral relationships will be examined.
Fractional ARMA models are used to estim ate
the orders of integration of the nominal ex
change rates, the ratios of the price levels and

10Koedijk and Schotman (1989) find that the real exchange
rates between 15 industrialized countries are fairly well
spanned by the real exchange rates between the United
States, Japan, Germany and Great Britain.

The residuals from equation (11) w ere then
treated as the real exchange rate for unit-root
testing with the fractional ARMA model. Cheung
and Lai (1993b) and Pippenger (1993) also esti
m ate a general cointegrating relationship, rath er
than impose a t = a , = 1. They both argue that,
because of m easurem ent erro r in price indexes
and unequal weights attached to the same good
in different indexes, it is undesirable to impose
unit coefficients on the price indexes w hen
studying w hether the real exchange rate is
m ean-reverting. The Phillips and Hansen (1990)

"Cheung and Lai (1993a) discuss the asymptotic theory be
hind estimating regressions where the residuals are frac
tionally integrated.

JULY/AUGUST 1993

Table 1

Estimated Orders of Integration_________________________
Real exchange
rate (b)

t-statistic
for 1 -b

Nominal exchange
rate (d1)

Price ratio
(d2)

United StatesUnited Kingdom

1.06
(.039)

-1.54

1.12
(.031)

1.60
(.043)

United StatesGermany

.903
(.067)

1.45

1.17
(.028)

1.21
(.081)

United StatesJapan

1.24
(.067)

-3.58

1.22
(.070)

1.62
(.080)

United KingdomGermany

.872
(.047)

2.72

1.01
(.052)

1.57
(.079)

United KingdomJapan

1.19
(.034)

-5.59

1.15
(.039)

1.12
(.039)

GermanyJapan

1.13
(.207)

-.628

1.19
(.142)

1.50
(.091)

Countries

NOTE: Standard errors are in parentheses.

method of estim ating cointegrating relationships
was used for equation (11). This method ac
counts for simultaneity in the determ ination of
left- and right-hand side variables. Cheung and
Lai (1993b) suggest using estim ation procedures
that take into account interactions betw een leftand right-hand side variables. In fact, the
Phillips-Hansen estim ates of a } and a 2 indicate
that they should not be restricted to equal one.
For example, for the nominal exchange rate b e
tw een Britain and the United States, the estim ates
of a 1 and a 2 are 1.45 and 1.22, respectively.
The next point of focus is the null hypothesis
to be tested: PPP holds as the null hypothesis if
the hypothesis that b < 1, w here b is the order
of integration of the real exchange rate, is not
rejected; the alternative null hypothesis that PPP
fails is not rejected if the null that b > l is not
rejected. Using the fractional ARMA model, it is
easy to test both null hypotheses and show how
the results depend on the choice of the null.
Table 1 contains the main results on the esti
mated orders of integration of the relevant ser
ies. Simple t-tests can be used to test for
long-run PPP by dividing one minus the estim at
ed order o f integration of the real exchange
rate by its standard erro r. Doing this, we see
that the null hypothesis that b < 1, w here b is
12Orders of integration greater than one for data in logs im
ply that the growth rates of the series have long memory.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

the order of integration of the real exchange
rate, is rejected for only tw o of the six pairs:
United States/Japan and Britain/Japan. These are
the significantly negative t-statistics in table 1.
Thus long-run PPP is not rejected as a null
hypothesis in four of six cases. If we reverse
the null, how ever, we can reject the null that
b > 1 for only one pair: Britain/Germ any.12 This
is the significantly positive t-statistic in table 1.
The results for United States/Germany are b o r
derline with a t-statistic of 1.45, but this is not
significant at the usual 5 percent level of sig
nificance in a one-tailed t-test, w here the critical
value is 1.658.
Overall, the orders of integration of real ex
change rates are often close enough to one that
neither null hypothesis is rejected. This explains
some discrepancies betw een past tests of longrun PPP. Coughlin and Koedijk (1990) used
Dickey-Fuller unit-root tests on real exchange
rates and could not reject the null that long-run
PPP fails to hold. Cheung and Lai (1993b), on
the other hand, tested the null that long-run
PPP holds and did not find many rejections of
long-run PPP. The results from the long-memory
model reconcile these findings.
The estim ates from the long-memory models
do more than give unit root tests, how ever, by

providing an estim ate of the order of integra
tion of the real exchange rate on a continuous
scale. W hichever null is used, one result is
clear: Even if long-run PPP holds, it is very slow
in developing. Assuming that the order of in
tegration of the real exchange rate is 0.9, 73
percent of a shock is still present after 12
months; 68 percent after 24 months; 65 percent
after 36 months; and 63 percent after 48 months.
As a practical m atter, it seem s fair to conclude
that PPP does not hold within a time horizon
that is econom ically relevant. Uncovering this
type of inform ation is the chief advantage of es
timating the order of integration on a continu
ous scale. W ith other unit-root testing methods,
we are forced to view a series as being either
1(0) or 1(1). Such a polar characterization may
not provide practical inform ation about the p er
sistence of the shocks.
Table 1 also provides inform ation about the
persistence of shocks to the nominal exchange
rate relative to shocks to the ratio of the price
levels. In five of six cases, point estim ates sug
gest that the nominal exchange rate has a lower
order of integration than the price ratio.13 For
Britain/Japan the point estim ate of the order of
integration is 1.15 fo r the nominal exchange
rate vs. 1.12 for the price ratio, but this d iffer
ence does not appear to be statistically signifi
can t.14 Thus the conjecture that nominal
exchange rates in the post-Bretton W oods era
have shown excess persistence appears to be
false. In general, g reater persistence in the
nominal rate would be needed to offset the p er
sistence in the price ratios for the real exchange
rate to be rendered 1(0). This is because no 1(0)
linear combination can exist, for example, b e
tw een a series that is I(.8) and one that is I(.2).
The series that is I(.2) does not have enough of
a tren d with w hich to offset the relatively
strong trend in the variable that is I(.8).
A nother finding from table 1 is that inflation
differentials are fractionally cointegrated in
some cases. For example, the estim ates indicate
that (p us- p JP ) is 1(1.62) and (p LIS- p U ) is 1(1.60),
N
K
but the d ifference (pUK- p JPf) is only 1(1.12). This
means that the inflation differentials betw een
the United States and Britain and betw een the

13An order of integration above one for variables in logs
means that the growth rates display long memory. In the
case of the price ratio, the corresponding growth rate is
the inflation differential across the two countries. For the
nominal exchange rate, it is the rate of exchange rate ap
preciation.

United States and Japan appear m uch m ore p er
sistent than the inflation differential betw een
Britain and Japan. In other words, inflation
rates in Britain and Japan com e closer to sh ar
ing a comm on trend with each other than with
inflation in the United States.
Tables 2 through 4 rep ort the fractional
ABMA param eter estim ates fully only for the
bilateral relationships for the United States for
the sake of brevity. The key result in these ta
bles is that in fractional ABMA models the fra c
tional differencing param eter can capture the
long-run behavior of the data, freeing AR
param eters to match the short-run dynamics. If,
on the other hand, an ARMA model instead of a
fractional ABMA model w ere fit to the data, the
autoregressive polynomial would be forced to
have a near-unit root.
In table 2 several AR param eters are negative,
and the largest equals 0.53 in the fractional
ABMA model of the ratio of the price levels b e
tw een the United States and Britain. In table 3
both estim ated AB param eters for the nominal
exchange rate betw een the United States and
Germany are negative, implying that all positive
dependence in the exchange rate beyond the
first lag is due to the large positive value of the
fractional-differencing param eter. The largest
root in an AR polynomial in table 3 is 0.37,
w hich is far from the unit circle, in the real ex
change rate betw een the United States and Ger
many. Estimates of the model of the real
exchange rate betw een the United States and
Japan, found in table 4, also show two negative
AR coefficients. In fact all of the AR polynomi
als in table 4 have roots with real parts that are
very far from the unit circle. They are - .0 5 ,
- .0 3 , and .08, respectively, for the real ex
change rate, the nominal exchange rate and the
price ratio. W ith the inclusion of the fractional
differencing param eter, the AR param eters can
take values which allow the fractional ARMA
model to capture both long-run dependence and
short-run dynamics in the data. The shaded in
sert and figures 1 through 3 provide a visual
check of the m atch betw een the covariance
stru ctu re of the data and that implied by the es
timated fractional ARMA model.

14A formal test would require joint estimates of the two frac
tional ARMA models, however.

JULY/AUGUST 1993

Table 2

Fractional ARMA Models: United States-United Kingdom
Real Exchange Rate
Parameter

Order of integration
AR
MA

b
^1
p2
0,

Parameter value

1.06
-.2 2 8
.012
.534

Standard error

.039
.096
.063
.084

Nominal Exchange Rate

Order of integration
AR
MA

di
P\
P2
01

1.12
-.161
.002
.452

.031
.100
.045
.090

Ratio of Price Levels

Order of integration
AR

*2
**1

1.60
.529
-.0 8 9
-.809

.043
.051
.043
.027

Table 3

Fractional ARMA Models: United States-Germany
Real Exchange Rate
Parameter

Order of integration
AR
MA

b
*>
1
"2
«1

Parameter value

.903
.508
-.050
-.176

Standard error

.067
.132
.060
.140

Nominal Exchange Rate

Order of integration
AR
MA

d1
Pi
P2
01

1.17
-.3 3 2
-.017
.463

.028
.099
.039
.105

Ratio of Price Levels

Order of integration
AR
MA

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

d2
Pi
P2
01

1.21
.326
.044
-.1 2 7

.081
.162
.071
.134

Table 4

Fractional ARMA Models: United States^lapan
Real Exchange Rate
Parameter

Order of integration
AR
MA

Parameter value

Standard error

1.24
-.105
-.191
-.2 2 4

.067
.126
.066
.172

Nominal Exchange Rate

Order of integration
AR
MA

p-l
Pz

1.22
-.065
-.174
.226

.070
.113
.073
.150

Ratio of Price Levels

Order of integration
AR
MA

d2
Pa
Pz
•i

1.62
.159
-.2 8 8
-.749

.080
.108
.058
.019

A Visual Check of the Results
O nce a fractional ARMA model has been es
timated, a visual diagnostic check of the ade
quacy of the specification and estim ates can
be obtained by looking at a plot of the periodogram of the data alongside a plot of the
spectral density implied by the specified fra c
tional ARMA model. If the model fits the data
well, the two graphs will have the same
general turning points. The area under the
spectral density equals the variance. Thus the
height of the spectral density at any point in
dicates how much of the variance is due to
cycles of that frequency. In this way, the
spectral density sum m arizes the autocovari
ance structure. Figures 1 through 3 provide a
look at the three data series fo r the United
States/Germany: the real exchange rate, the
nominal exchange rate and the price ratio.
The figures show that the estim ated fractional
ARMA models roughly envelope the smoothed
1ln theory the distinction is clear: A series with a nega
tive order of fractional integration will have a spectral
density value of zero at frequency zero; a series with a
positive order of fractional integration will have a spec
tral density value of infinity at frequency zero. For
figures 2 and 3, it is clear from the periodogram that
the series have positive orders of integration. This is not
clear for the real exchange rate in figure 1.

periodograms. Figure 1 shows why the out
come of the unit root test on the U.S./
Germany real exchange rate is borderline: It
is unclear w hether the upturn in the smoothed
periodogram at very low frequencies is sig
nificant or w hether the point estim ate of a
negative order o f integration for the differ
enced real exchange rate is c o rre ct.1 One
other thing to note is that without ARMA
param eters the model would be able to fit
o n ly s e r ie s that h a v e g lo b a lly c o n c a v e (o r
globally convex) spectral densities.2 W ith
ARMA param eters, how ever, figure 1 illus
trates that a fractional ARMA can generate
turning points in the spectral density. The
best-fitting globally concave spectral density
would obviously provide a much less satisfac
tory fit of the U.S./Germany real exchange
rate than the one with turning points shown
in figure 1.
2Without ARMA parameters, the fractional ARMA model
is called the fractional noise model. Its spectral density
is globally concave if the fractional differencing
parameter from equation (2) is negative; it is globally
convex if the fractional differencing parameter is
positive.

JULY/AUGUST 1993

Figure 1

Spectrum of Differenced Real Exchange Rate: U.S./Germany
ARFIMA (2,d,1) Spectrum

Smoothed Periodogram

Frequency

Figure 2

Spectrum of Differenced Nominal Exchange Rate: U.S./Germany
ARFIMA (2,d,1) Spectrum

http://fraser.stlouisfed.org/
FEDERAL St. Louis
Federal Reserve Bank ofRESERVE BANK OF ST. LOUIS

Smoothed Periodogram

Frequency

Figure 3
Spectrum of Differenced Price Level Ratio: U.S./Germany
ARFIMA (2,d,1) Spectrum

Smoothed Periodogram

Frequency

CONCLUSIONS
This article illustrates the key role played by
the null hypothesis in testing fo r unit roots in
real exchange rates. If b is the order of integra
tion of the real exchange rate, then the null
that f a > l is difficult to reject, in which case one
would presum e that long-run purchasing-pow er
parity does not hold. W hen the null is that
b < 1, we also find few rejections, so long-run
PPP apparently holds. W hen this type of am
biguity appears, it is helpful to estim ate the ord
er of integration on a continuous scale. The
fractional ARMA models presented here do this
and the standard erro rs on b for the six real
exchange rates studied show that even if b < 1,
it is not fa r enough away from one to make a
strong case that purchasing-pow er parity is em
pirically relevant.

Cheung, Yin-Wong. “ Long Memory in Foreign Exchange
Rates,” Journal of Business and Economic Statistics (Janu
ary 1993), pp. 93-101.
Cheung, Yin-Wong, and Kon S. Lai. “A Fractional Cointegra
tion Analysis of Purchasing Power Parity," Journal of Busi
ness and Economic Statistics (January 1993a), pp. 103-12.
_______ “ Long-Run Purchasing Power Parity during the Re
cent Float,” Journal of International Economics (February
1993b), pp. 181-92.
Chowdhury, Abdur R., and Fabio Sdogati. “ Purchasing Power
Parity in the Major EMS Countries: The Role of Price and
Exchange Rate Adjustment,” Journal of Macroeconomics
(Winter 1993), pp. 25-45.
Coughlin, Cletus C., and Kees Koedijk. “ What Do We Know
About the Long-Run Real Exchange Rate?” this Review
(January/February 1990), pp. 36-48.
Dahlhaus, Rainer. “ Small-Sample Effects in Time Series
Analysis: a New Asymptotic Theory and a New Estimate,”
Annals of Statistics (Volume 1b, 1988), pp. 808-41.
_______ “ Efficient Parameter Estimation for Self-Similar
Processes,” Annals of Statistics (Volume 17, 1989), pp.
1749-66.

REFERENCES

Dickey, David A., and Wayne A. Fuller. “ Likelihood Ratio
Statistics for Autoregressive Time Series with a Unit Root,”
Econometrica (July 1981), pp. 1057-72.

Aliber, Robert Z. “ The Case for Flexible Exchange Rates
Revisited,” unpublished manuscript, University of Chicago
(April 1993).

Diebold, Francis X., and Glenn D. Rudebusch. “ Long
Memory and Persistence in Aggregate Output,” Journal of
Monetary Economics (September 1989), pp. 189-209.

Caves, Richard E., Jeffrey A. Frankel and Ronald W. Jones.
World Trade and Payments: An Introduction (Scott, Foresman, 1990).

Edison, Hali J., and Eric O'N. Fisher. “A Long-Run View of
the European Monetary System,” Journal of International
Money and Finance (March 1991), pp. 53-70.

JULY/AUGUST 1993

Engel, Charles. “ Real Exchange Rates and Relative Prices:
An Empirical Investigation,” NBER Working Paper No. 4231
(December 1992).

Johansen, Soren. “ Estimation and Hypothesis Testing of
Cointegrating Vectors in Gaussian Vector Autoregressive
Models,” Econometrica (November 1991), pp. 1551-80.

Engel, Charles, and James D. Hamilton. “ Long Swings in the
Dollar: Are They in the Data and Do Markets Know It?”
American Economic Review (September 1990), pp. 689-713.

Koedijk, Kees, and Peter Schotman. “ Dominant Real Ex
change Rate Movements,” Journal of International Money
and Finance,” (December 1989), pp. 517-31.

Fisher, Eric O’N., and Joon Y. Park. “ Testing Purchasing
Power Parity Under the Null Hypothesis of Co-Integration,”
Economic Journal (November 1991), pp. 1476-84.

McNown, Robert, and Myles S. Wallace. “ National Price Lev
els, Purchasing Power Parity, and Cointegration: A Test of
Four High Inflation Economies,” Journal of International
Money and Finance (December 1989), pp. 533-45.

Fox, Rort, and Murad S. Taqqu. “ Large Sample Properties of
Parameter Estimates for Strongly Dependent Stationary
Gaussian Time Series,” Annals of Statistics (Volume 14,
1986), pp. 517-32.

Phillips, Peter C.B., and Bruce E. Hansen. “ Statistical Infer
ence in Instrumental Variables Regression with 1(1)
Processes,” Review of Economic Studies (January 1990),
pp. 99-125.

Granger, C.W.J. “ Developments in the Study of Cointegrated
Economic Variables,” Oxford Bulletin of Economics and
Statistics (August 1986), pp. 213-28.

Pippenger, Michael K. “ Cointegration Tests of Purchasing
Power Parity: The Case of Swiss Exchange Rates,” Journal
of International Money and Finance (February 1993),
pp. 46-61.

Granger, C.W.J., and Roselyne Joyeux. “An Introduction to
Long-Memory Models and Fractional Differencing,” Journal
of Time Series Analysis (Volume 1, 1980), pp. 15-29.

Sowell, Fallaw. “ Maximum Likelihood Estimation of Stationary
Univariate Fractionally Integrated Time Series Models,”
Journal of Econometrics (July-September 1992a),
pp. 165-88.

Hakkio, Craig S. “ Does the Exchange Rate Follow a Random
Walk? A Monte Carlo Study of Four Tests for a Random
Walk,” Journal of International Money and Finance (June
1986), pp. 221-29.

_______ “ Modeling Long-Run Behavior with the Fractional
ARIMA Model,” Journal of Monetary Economics (April
1992b), pp. 277-302.

Hosking, J.R.M. “ Fractional Differencing,” Biometrika (Volume
1b, 1981), pp. 165-76.

Summers, Robert, and Alan Heston. “ The Penn World Table
(Mark 5): An Expanded Set of International Comparisons,
1950-1988,” Quarterly Journal of Economics (May 1991), pp.
327-68.

Hsieh, David A. “ The Determination of the Real Exchange
Rate: the Productivity Approach,” Journal of International
Economics (May 1982), pp. 355-62.

Taylor, Mark P “An Empirical Examination of Long-Run Pur
.
chasing Power Parity Using Cointegration Techniques,” Ap
plied Economics (October 1988), pp. 1369-81.

Huizinga, John. “An Empirical Investigation of the Long-Run
Behavior of Real Exchange Rates,” Carnegie-Rochester
Conference on Public Policy (Autumn 1987), pp. 149-214.

Whitt, Joseph A. Jr. “ Nominal Exchange Rates and Unit
Roots: A Reconsideration,” Journal of International Money
and Finance (December 1992), pp. 539-51.

Appendix:
The Fox and Taqqu Estimator
Dahlhaus (1988) discusses why the Fox and
Taqqu (1986) frequency domain estim ator is an
approxim ate maximum-likelihood estim ator,
sharing the same optimality properties as exact
maximum-likelihood estimation. The Fox and
Taqqu estim ator is derived from the following
minimization problem:
v

—i

(12) min y

/(A.)

, (/n(cr/U ;0)) + ---------------]
o 2/ ( Xk;6)

w here /(Afc is the vector of periodogram o r
)
dinates of the data and o 2 k) is the spectral
f(\

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

density function implied by the param eterized
model. For the fractional ARMA model in equa
tion (8), the spectral density equals
(13) j U) = ^ (e" ^ (2 - e ‘x)~d (1 - e ~ ' x)~d,

IA(e-*)|2
w here A and B are polynomials defined in equa
tions (9) and (10). T he objective function is
minimized with respect to 6 and o2- An intuitive
description of the objective function is that one
w ants to choose param eters that will m ake the
spectral density function implied by the model
look like the periodogram of the data.

Richard G. Anderson
Richard G. Anderson is a research officer with the Federal
Reserve Bank of St. Louis. Heather Deaton provided
research assistance.

The Effect o f Mortgage
Refinancing on M oney Demand
and the Monetary Aggregates

. ONEY SERVES AS A medium of exchange
fo r transactions involving financial instrum ents
as well as real goods and services. Unfortunately,
the total volume of transactions in the economy
is not observable. As a result, econom ic analyses
of money demand typically focus on the relation
ship betw een the quantity of m oney demanded
and the production of new goods and services,
m easured by either gross domestic product or
personal consumption expenditures. Because ag
gregate volumes of financial and nonfinancial
transactions likely move in parallel with the out
put of new goods and services, the use of out
put rath er than the volume of transactions may
cost little in term s of understanding movements
in the m onetary aggregates. In some periods,
however, events occu r which rem ind us that
this is not always the case. This article examines
the effect of one such ongoing recen t event—
the refinancing of residential m ortgages—
on money dem and.1
'Other recent examples include the Tax Reform Act of 1986,
which boosted household liquid deposits in late 1986 and
early 1987, and the closure of large numbers of thrifts by
the Resolution Trust Corporation. Recognizing that special
factors can significantly distort growth of the monetary ag
gregates, the Bach commission recommended that the
Federal Reserve regularly undertake and publish studies of
the effects of special factors; see Report of the Advisory

Simple models of the demand for money as a
medium of exchange often implicitly assume that
the purchase or sale of a good or service is com
pleted within a relatively b rief period. Unlike
the transactions in these models, the refinancing
of a residential m ortgage that has been securi
tized in the secondary m arket initiates a sequence
of transactions that may continue for four to six
weeks, or m ore. During this time, the quantity
of liquid deposits demanded increases. W hen the
last transaction in the sequence is concluded,
the quantity of deposits demanded falls back
ceteru s p a rib u s to its earlier level.
Mortgage refinancing is an im portant phenom e
non in the United States because most homes
are financed with long-term, fixed-rate amortized
mortgages that contain a “put” option, allowing
the borrow er to repay the outstanding principal
amount of the loan at any tim e without penalty.
Homeowners typically exercise that option when
mortgage rates fall significantly (1-2 percentage
Committee on Monetary Statistics (1976). The Bank of
England regularly publishes such analyses; see Pepper
(1992, 1993) and Topping and Bishop (1989).

JULY/AUGUST 1993

points) below recen t previous levels by taking
out a new m ortgage loan to repay the old.
As shown in figure l , 2 extensive m ortgage r e
financing has occu rred during two periods in
the last decade, 1 9 8 6 -8 7 and 1 9 9 1 -9 3 . In the
form er, an initial surge in refinancing during
1986 was interrupted by a pause, b efore fears
of rising m arket rates launched a second round
in 1987. In the latter, th ree waves of refinancing
—of increasing m agnitude—m irrored the halting
fall in long-term m arket interest rates. During
1992, fo r example, nearly one-fifth of all hom e
ow ners refinanced their m ortgages.3 In 1993,
the volume of refinancing activity will surpass
1992’s record pace.
The next section of this article describes the
changes in the grow th and volatility of liquid
deposits and M l that have occurred during
periods of extensive m ortgage refinancing. The
article then exam ines the extent to w hich these
changes may be related to increases in mortgage
securitization. Finally, it explores w h eth er recen t
fluctuations in the grow th of oth er checkable
deposits (OCDs) since 1991 also may be related
to m ortgage refinancing.

during late 1991, the third q uarter of 1992 and
the second quarter of 1993, liquid deposit grow th
accelerates. As refinancings continue at the
higher rate, deposit levels converge to the new
desired level and deposit grow th slows. W hen
refinancing activity subsides—as in mid-1992
and early 1993—liquid deposit grow th slows fu r
th er and deposits may run off.
Through its effect on liquid deposits, mortgage
refinancing sharply increased the volatility of
M l during both 1 9 8 6 -8 7 and 1 9 9 1 -9 3 , as shown
in figure 3.4 At the same time, the volatility of
the broader aggregate M2, shown in figure 4,
apparently was only slightly affected. In large
part, the low er sensitivity of M2 to mortgage
refinancing reflects the m uch sm aller share of
transaction deposits in M2 (about 20 percent)
than in M l (about 70 percent). The small
changes that do appear in the volatility of M2
closely resem ble changes in its non-M l com
ponent.5

The increases in liquid deposits that have ac
companied accelerations in mortgage refinan c
ing since m id-1990 are shown in figure 2. The
link betw een m ortgage refinancing and liquid
deposit grow th is a stock adjustm ent process
w herein the stock of liquid deposits responds to
changes in the flow of refinancings. W hen the
pace of mortgage refinancing increases, as it did

The ability of increases in m ortgage refinancing
to affect the level and volatility of liquid deposits
and M l is in part due to the borrow ed reserves
operating procedure used by the Federal Reserve
to control the grow th of M2. During the last de
cade, this operating procedure has largely
evolved into one that closely stabilizes the fed er
al funds rate about a level thought to be consis
tent with the desired amount of discount
window borrow ing and the grow th of M2. To
maintain the desired levels of the federal funds
rate and discount window borrow ing, transitory
increases in the demand fo r reserves are auto
matically accom m odated with increases in the
supply of nonborrow ed reserv es.6

2ln the figure, the volume of refinancing activity is proxied
by liquidations of mortgage-backed securities. This concept
is explored further in this article.

No such correlations between refinancing-related deposit
inflows and nontransaction funding sources are apparent in
the data, however.

3Nineteen percent of the homeowners interviewed in Fannie
Mae’s 1993 national housing survey had last refinanced
their mortgage between January 1992 and March 1993. An
additional 3 percent had refinanced during 1991 and 1990.

6For an analysis of the borrowed reserves procedure and its
relationship to federal funds rate targeting, see Thornton
(1988). For a careful discussion of why and how reservesbased targeting procedures evolve into federal funds rate
targets, see Meulendyke (1990).

MORTGAGE REFINANCING AND
MONEY DEMAND

4The coefficient of variation shown in the figure equals the
ratio of the standard deviation to the mean of the series,
each calculated from the most recent 12 months of data.
The coefficient of variation indicates whether the variability
of the data has increased or decreased over time relative
to its average level.
5The volatility of M2 differs little from that of its non-M1
component. It is feasible that banks’ cash management
practices might account for the insensitivity of M2 volatility.
Increases in liquid deposits provide additional funds to
banks. If bank cash managers respond by reducing their
issuance of overnight repurchases (RPs), the change in the
volatility of M2 might be considerably less than that of M1.

FEDERAL RESERVE BANK OF ST. LOUIS

F ig u re

M ortgage Interest Rate and Refinancing Activity
Percent

Billions of dollars
Monthly data, January 1984-September 1993

Contract rate on 30-year, fixed
-40

-30

-2 0

-10

1993

Liquidations of federal-agency-guaranteed mortgage-backed securities.

F ig u re

Refinancing Activity and Liquid Deposits
Billions of dollars
Monthly data, seasonally adjusted

50-

Index of mortgage
refinancing activity

40C h an g e in

Liquid Deposits

3020-

10-

[JU

Liquid D e p o s its= D em and D e p o s its+ O th e r C h e c k a b le D e p o s its * S a v in g s and MMDA d e p o s its
-

10-

i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i i

J ASON DJ FMAMJJASONDJ FMAMJJASONDJ FMAMJJ AS
1990
1991
1992
1993
Liquidations of federal-agency-guaranteed mortgage-backed securities.

JULY/AUGUST 1993

Figure 3
Refinancing and the Volatility of M1
Billions of dollars

Index

Monthly data, seasonally adjusted

Shaded areas are periods of heavy refinancing activity.

Figure 4
Refinancing and the Volatility of M2
Billions of dollars
Monthly data, seasonally adjusted

92
Shaded areas are periods of heavy refinancing activity.

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

1993

THE ROLE OF MORTGAGE
SECURITIZATION
The increase in mortgage securitization during
the last decade has increased the potential for
mortgage refinancing to affect the grow th of the
m onetary aggregates.7 The sale of m ortgages in
the secondary m arket creates an additional finan
cial instrum ent—the m ortgage-backed security,
or MBS—and involves a num ber of additional
firm s in the m ortgage process, including the
originators of the mortgages, the assem bler of
the m ortgage pool (who also issues the MBSs),
the servicer of the mortgage pool (who collects
monthly payments and disburses funds to inves
tors) and, typically, at least one governm ent
agency. The refinancing of securitized mortgages
thus becom es a circuitous calling and refunding
of relatively large amounts of long-term, publicly
held debt. Elevated levels of liquid deposits may
persist for four to six weeks or m ore, until all
related transactions are settled.
Legally, mortgage securitization entails com
bining a fixed pool of m ortgages into a trust.
The m ortgages serve as collateral for MBSs sold
against the trust. The servicer of the MBSs, as a
trustee, collects payments from hom eow ners
and passes them through w ithout taxation to
the holders of the MBSs. Liquidity of the MBSs
is enhanced by obtaining a third-party guaran
tee covering the payments that will be due to
investors if hom eow ners pay at the scheduled,
minimum contract rate. T h ree federal-government-sponsored enterprises, known as "agen
cies,” dominate that business.8 For a fee, these
agencies guarantee the paym ent of principal
and interest on securities backed by pools of
specified mortgages. The Governm ent National
Mortgage Association (Ginnie Mae, or GNMA),
7See Duca (1990) for an analysis of the interactions between
demand deposits and mortgage refinancing during 1986-87.
8 small amount of MBSs is issued without agency guaran
A
tees. Bank of America issued the first such private mort
gage pool in 1977. In 1992, private mortgage pools
represented only 8 percent of all outstanding pools. For
background, see Downs (1985) and Pavel (1986).
9The precise nature of the guarantee varies somewhat by
agency. GNMA and FNMA guarantee timely (within the
month) payment of principal and interest, regardless of
payments by the borrower. FHLMC guarantees timely pay
ment of interest and eventual (within the year) payment of
principal. In addition to issuing guarantees on MBSs backed
by privately assembled mortgage pools, FNMA and FHLMC
may purchase mortgages outright and market MBSs
backed by pools of those mortgages. In 1992, for example,
FNMA “ issued” (guaranteed) $194 billion in MBSs. Of that
amount, about $13 billion were originated by FNMA itself;

a part of the D epartm ent of Housing and Urban
Development, guarantees payments on MBSs
backed by pools of Federal Housing Administra
tion (FHA) and Veterans Administration (VA)
mortgages. The Federal National Mortgage
Association (Fannie Mae, or FNMA), a federally
chartered, privately owned stock corporation,
and the Federal Home Loan Mortgage Corpora
tion (Freddie Mac, or FHLMC), a wholly owned
subsidiary of the federally chartered Federal
Home Loan Bank System, guarantee payments
on MBSs backed by pools of conventional
m ortgages.9
Absent refinancings or home sales, MBS in
vestors receive a monthly payment that includes
the scheduled am ortization of the pool’s m ort
gage principal plus the accum ulated interest.
Refinancings, home sales and an occasional ex
tra payment by a hom eow ner retu rn additional
(or unscheduled) principal pro rata to the
holders of the MBSs backed by that mortgage
pool. The monthly liquidation fo r a mortgage
pool is the sum of the scheduled and unscheduled
principal payments retu rned to investors. Note
that MBSs aren ’t "called" in the traditional sense
associated with corporate bonds, but rath er are
only proportionately liquidated or repaid.
As shown in the upper panel of figure 5, the
outstanding stock of MBSs increased about six
fold during the last decade, m uch m ore rapidly
than M l o r M2. W ith few changes in mortgage
servicing rules and practices during the last
decade, the rapid grow th of securitization sug
gests that the transactions incurred in refinancing
securitized m ortgages will have larger effects on
the m onetary aggregates in the 1990s than they
did in the m id-1980s. Annual liquidations of
MBSs, shown in the low er panel of the figure,
the balance was originated by private lenders under a
FNMA guarantee plan. FNMA’s 1992 Annual Report em
phasizes the off-balance-sheet contingent risk nature of
these securities: “ MBS are not assets of the corporation
[FNMA], except when acquired for investment purposes,
nor are the related outstanding securities recorded as lia
bilities. However, the corporation is liable under its guaran
tee to make timely payment of principal and interest to
investors. The issuance of MBSs creates guaranty fee in
come with Fannie Mae assuming credit risk, but without
assuming any debt refinancing risk on the underlying
pooled mortgages.” In 1992, FNMA recorded $834 million
in guaranty fees.

JULY/AUGUST 1993

Figure 5
M ortgage-Backed Securities O utstanding atYear-E nd
Billions of dollars
1400—-----------------------------------------------------------------------------------------------------------------------------------1
----------------------Data through September 1993
r—
■
1200-

I■

GNMA
GNMA

1000-

■
■

FHLMC
FHLMC

800-

□
□

FNMA
FNMA

|-|

600-

400
200

--------------------------------------

0
1972 73 74 75 76 77 78 79 80 81

82 83 84 85 86 87 88 89 90 91 92 1993

End-of-year level

Annual Liquidations of M ortgage-Backed Securities
Billions of dollars
350
D ata th ro u g h S e p te m b e r 1993

300

■

GNMA

250

FHLMC

□

FNMA

200
150
100

50
0
1972 73 74 75 76 77 78 79 80 81

FEDERAL RESERVE BANK OF ST. LOUIS

82 83 84 85 86 87 88 89 90 91 92 1993

Annual total

have on balance increased in proportion to the
outstanding stock except for significant surges
during periods of refinancing. Annual liquida
tions jumped to about 17 percent of the out
standing stock of MBSs during 1 9 8 6 -8 7 and 19
percent during 1 9 9 1 -9 2 . More recently, liquida
tions during Ju n e through Septem ber 1993
averaged nearly $44 billion a month, almost a
40 percent annual rate. Recent fu rth er decreases
in mortgage rates portend continuing high liqui
dation rates during late 1993 and early 19 9 4 .10
The increase in deposits that follows an in
crease in mortgage refinancing activity may in
part be traced to the m echanics of mortgage
securitization and servicing. Mortgage servicers’
handling of the unscheduled principal payments
associated with refinancings is governed by the
rules of the federal agency that guarantees the
MBSs issued against the m ortgage pool. In gener
al, these rules require that m ortgage servicers
hold unscheduled principal payments in special
custodial accounts during the interval betw een
receipt from hom eow ners and disbursem ent to
MBS investors. GNMA requires that these cus
todial accounts be non-interest-bearing demand
deposits. FNMA allows funds to be held in
interest-bearing accounts as long as they are im
mediately available without prior notice of w ith
drawal. FHLMC’s rules are similar to FNMA's.
A surge in refinancing greatly increases the
monthly average amount of funds held in liquid
deposits by a m ortgage servicer. In a typical
month without refinancing, a servicer holds a
hom eow ner's mortgage paym ent for a relatively
b rief period of time (up to 15 days) before
rem ittance to investors. Following a mortgage
refinancing, how ever, the servicer will hold the
unpaid principal balance of the extinguished
mortgage loan—an amount perhaps 10 to 100
(or more) times as large as the hom eow ner’s
regular monthly principal paym ent—in a cus
todial account for a m uch longer period, often
10While it is always risky to forecast financial market activity,
recent decreases in mortgage rates (through October 1993)
are likely to trigger substantial further increases in re
financing and MBS activity during late 1993 and early
1994. In addition to older mortgages issued during the
1980s, mortgages that were issued as little as 12 to 18
months ago at 7 to 7-1/2 percent rates now may profitably
be refinanced. Rather than the pace of refinancing slowing
and related distortions to the monetary aggregates diminish
ing as the outstanding stock of seasoned MBSs are rolled
over, recent rate decreases have placed nearly the entire
outstanding stock of MBSs “ in the money” for rollover.
"Homeowners typically make monthly mortgage payments
between the 1st and 15th of the month, with the servicer
remitting these funds to MBS investors on the 15th. Follow

two to six weeks (see the shaded insert).11
Estimates of the size of this effect on monthly
grow th rates of demand deposits, M l and M2,
are shown in figure 6 .12 W hen MBS liquidations
accelerate, the grow th rates of demand deposits
and M l after rem oving the MBS effect are
smaller than the published grow th rates. Con
versely, w hen MBS liquidations slow, the MBSadjusted grow th rates are larger than the pub
lished rates. Overall, the estimated differences
in grow th rates equal in some months as much
as one-half of the change in M l. From D ecem ber
1991 to M arch 1992, for example, inflows to
mortgage servicers’ custodial accounts are esti
mated to have added betw een 5 to 10 percentage
points to the monthly grow th rates of demand
deposits. The largest estimated effects w ere in
O ctober 1992 and May 1993, when MBS-related
inflows likely accounted fo r four-fifths and
three-fifths, respectively, o f demand deposit
grow th. In both cases, deposit grow th slowed
sharply in later m onths w hen deposit levels had
increased enough to support the accelerated
pace of mortgage activity. Subsequently, during
the first q uarter of 1993, runoffs of servicers’
custodial balances likely depressed monthly
average deposit grow th by as much as 10 p er
centage points.
These patterns show through to M l (see the
cen ter panel of figure 6) but are muted. C urren
cy and OCDs, w hich comprise two-thirds of M l,
are unlikely to be affected by MBS activity.13
Nonetheless, the distortions to demand deposits
are sufficient that monthly grow th rates of M l
since mid-1992 appear to have been distorted
by as m uch as 5 to 7 percentage points. Similar
estim ates for M2 that include estim ated effects
on money m arket demand account (MMDA)
balances are shown in the bottom panel of the
figure.
Overall, fluctuations in mortgage servicers’
ing a refinancing, the funds received by the servicer from
the homeowner (at any time within the month) are placed
in a custodial account. These funds are remitted by the
servicer to MBS investors after the middle of the following
month. The exact date, however, depends on the contract
specifications of the agency guarantee program under
which the MBSs backed by the mortgage pool that con
tained the extinguished mortgage were issued. See, for ex
ample, Karcher (1989).
1Construction of these estimates is discussed in the ap
pendix.
1
3The next section raises the possibility that OCD balances
also might have been affected by refinancing since 1991,
albeit not through MBS-related transactions.

■
IULY/AUGUST 1993

Extra Deposits:
Where Do They Come From? Where Do They Go?
T he im pact of surges in refinancing-related
transactions on the demand for liquid de
posits and the m onetary aggregates depends
on the m onetary control operating procedure
being used by the Federal Reserve. The
refinancing of securitized m ortgages gen er
ates tem porary increases in the demand for
liquid deposits. Since these deposits are sub
ject to non-zero reserve requirem ent ratios,
the Federal Reserve accom m odates this de
mand under its cu rren t m onetary control
procedures by furnishing additional reserves
as necessary to maintain the federal funds
rate near the level expected to be consistent
with desired longer-run grow th of M2. W hen
final paym ents to MBS investors are com plet
ed, both the quantity of liquid deposits de
manded and banks’ required reserves fall. To
again m aintain the desired level of the fed er
al funds rate, the Federal Reserve w ithdraws
the now-surplus reserves from the market.
Consider, fo r example, a 10 percent m ort
gage with an unpaid principal balance of
$100,000. Assume that the hom eow ner

'$877*15/30. The implied intramonthly pattern of fluctua
tions in transaction deposits and reserves is character
istic of liquid deposits. Aggregate transaction deposits
tend to increase sharply at the beginning of each month

custodial deposits likely account for about onehalf of the recen t increase in M l volatility. It is
unlikely that these estim ates are too large, since
they are based on legal restrictions imposed on
m ortgage servicers by federal agencies and
realistic but conservative assumptions regarding
intra-m onth patterns of mortgage closings and
deposit behavior.
The estim ates may be biased downward,
how ever, for a num ber of reasons. The most
im portant perhaps is the omission of any in
crease in deposits held by issuers of new MBSs.
As some issuers draw on bank warehouse
credit lines to fund the purchase of m ortgages
to be assembled into new MBS pools, they may
offset part of the bank charges for these lines
via earnings credits based on their deposit lev

FEDERAL RESERVE BANK OF ST. LOUIS

deposits some funds (a paycheck, for exam
ple) into a demand deposit account on the
first day of a typical month. On the fifth day
of the month, the hom eow ner mails a check
for $877 to his m ortgage servicer. The serv
icer receives the check on the 10th and
places the funds in a demand deposit ac
count. On the 15th, the funds are paid to
MBS investors who immediately move the
funds out of demand deposits and into new
earning assets. The extent to w hich this
transaction affects the average daily levels of
demand deposits and M l depends on what
type of asset/deposit was held by the firm
prior to paying the employee and on what
type of asset/deposit is held by the MBS in
vestor after receipt of the funds. Assuming
that the firm held the funds prior to the first
of the month in an asset not included in M l,
and assuming that investors move funds
promptly into assets not included in M l on
the 15th, this single transaction contributes
about $439 to the average monthly level of
demand deposits.1

and run off during the latter weeks of the month. The
implied fluctuations in reserve demand are accommo
dated by the Federal Reserve, perhaps through an RP
by the Open Market Desk.

els. Also omitted are any increases in liquid
deposits that arise because of the significant
volume of additional transactions used to pur
chase and sell large quantities of mortgages and
MBS.

HOUSEHOLD DEPOSITS AND
REFINANCING
In addition to demand deposits, changes in
OCDs since mid-1991 also have reflected the
ebbs and flows in the pace of m ortgage
refinancing (see the upper panel of figure 7).
The apparent increase in the correlation of
OCDs with demand deposits contrasts with its
behavior b efore 1991 and during 1 9 8 6 -8 7 , the
latter shown in the low er panel of figure 7.

To illustrate the magnitude of refinancingrelated payments, suppose that the hom e
ow ner now refinances the m ortgage on the
25th day of the month, with the servicer
receiving funds on the 30th and holding them
(as a fiduciary) in a demand deposit custodial
account until rem ittance to investors around
the middle of the following month. The
refinancing, w hen it closes on the 25th, c re
ates $100,000 of demand deposits that didn’t
previously exist, reflecting the new mortgage
loan extended to the hom eow ner. If the
transaction is subject to Regulation Z’s rightof-rescission provisions, the $100,000 deposit
likely will be held by the settlem ent agent or
new lender for the first three days following
the mortgage closing. If the mortgage has
been securitized via federal-agency-guaranteed
MBSs, the funds subsequently will be rem itted
to the servicer of the extinguished mortgage.
If not, the funds will be paid to the current
ow ner of the original mortgage. Since the
outstanding MBSs backed by the old mortgage

have not yet been extinguished, the new
mortgage (and new deposits) rep resent a tem
porary net increase in the amount of out
standing credit in the economy. Both the new
deposit and the MBSs backed by the old
m ortgage continue to exist until about the
middle of the following m onth.2
The mortgage refinancing of $100,000 con
tributes $16,666 to the average level of de
mand deposits during the cu rren t month and
$50,000 to the average level of the following
month, assuming that the servicer rem its
funds to investors on the 15th and investors
immediately tran sfer the funds from demand
deposits.3 W hen investors do so, the ag
gregate level of demand deposits drops and
the Federal Reserve will drain reserves from
the m arket if necessary to maintain discount
window borrow ings and the federal funds
rate near the desired levels.

2Note that financial market participants (and federal
agencies) record an MBS as being liquidated on the
last day of the month in which the refinancing occurred,
even though investors will not receive the underlying
funds until after the middle of the following month.
3
What asset might investors buy with the demand
deposit? One possibility is new MBSs backed by the
new mortgages. What happens to the demand deposits
that they use to purchase these new MBSs? They van
ish, in textbook multiple-expansion-of-deposits fashion,
accompanied by the Fed’s withdrawal of reserves.

Should some portion of the OCD fluctuations
during 1 9 9 1 -9 3 be attributed to m ortgage re
financing activity? If so, and if the impact of re
financing on OCDs w ere similar to its effect on
demand deposits, then their com bined effects
could account for as much as three-quarters of
M l's grow th during a num ber of m onths since
1991.

to mortgage activity is necessarily less direct
and more circum stantial than that for demand
deposits. Tracing direct links betw een house
hold deposits and econom ic activity is generally
not possible, since the Federal Reserve collects
deposit data from the issuers of deposits such as
banks and thrifts rath er than from the ow ners of
deposits, including households and firm s.14

The recen t parallel monthly m ovements in
these two types of liquid accounts is compelling
but puzzling. Any evidence linking these deposits

W hy might a household increase its OCD
balances following a m ortgage refinancing? One
possibility could be the conversion of home eq-

14Although the Federal Reserve Board’s flow of funds ac
counts present a fairly complete balance sheet for the
household sector, few items are directly observed. Most en
tries are calculated as residuals, inferred from the double
entry nature of the accounts and from balance sheet data

for firms and government. See Guide to the Flow of Funds
Accounts, p. 120.

JULY/AUGUST 1993

Figure 6

Published Growth Rates Less Rate Adjusted for MBS Activity
Demand Deposits
Percent

Monthly data

3020-

10-

20-

____

noDD _ _

I ^ H I
■

_ -„ n [l_

w ^m

c
I

E
C
C

10-

J AS ONDJF MAMJJASONDJF MAMJJASONDJFMAMJJAS

1990

1991

1992

1993

M1
Percent
Monthly data

JASONDJ

FMAMJ J A S O N D J

1990

FMAMJ J A S O N D J F M A M J

1991

1992

JAS

1993

M2
Percent
Monthly data

j l

"B-

l l

NOTE: Includes MBS effects on demand deposits only.
JASONDJFMAMJ

1990

JASONDJFMAMJ J ASONDJFMAMJ

1991

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

1992

1993

JAS

Figure 7

Average Monthly Change in Demand Deposits and OCDs,
by Quarter
Billions of dollars

OCDs
Demand Deposits

IV 1 i

1 M r m— r ~iv~

1990

1991

1992

III

1993

Average Monthly Change in Demand Deposits and OCDs,
by Quarter
Billions of dollars

1985

1986

1987

1988

JULY/AUGUST 1993

uity into cash at the time of refinancing. If
operative, this factor should be m uch stronger
during the 1990s than during 1 9 8 6 -8 7 , for two
reasons. First, many households have been res
tructuring their balance sheets, seeking to reduce
the levels of debt (and debt service) that they
took on during the 1980s. Home equity convert
ed to cash at refinancing allows them to repay
other outstanding debt and reduce monthly
debt service. Second, households generally ex
perienced large capital gains on houses during
the 1980s. For many, capital gains in housing
appeared largely as a windfall, accruing more
rapidly than had been anticipated w hen the
hom e was purchased and without any overt ef
fort by the hom eow ner. As such, these in
creases in wealth likely w ere not optimally
deployed (from a portfolio standpoint) across all
household asset categories. For other hom e
ow ners who might have p referred to consume
the increased w ealth rath er than save it, the
capital gain appears as a type of forced saving
in the form of home equity. W hile a home equi
ty loan may increase the liquidity of home equi
ty, it doesn’t perm it the household to consum e a
windfall increase in home equity, since the loan
must be repaid. Hence, there may be some
pent-up demand by hom eow ners for redirecting
part of their home equity tow ard balance sheet
restru cturing (reducing other consum er debt),
consum ption or perhaps redeploym ent into
m ore liquid assets.
Although no direct data on cash withdrawals
at m ortgage refinancings are available, recent
evidence is supportive. Fannie M ae’s 1993 na
tional housing survey asked households w hether
their prim ary motivation in refinancing was to
shorten the m aturity of the loan (thereby build
ing equity m ore quickly) or to reduce their
monthly payments. W hile a shorter m aturity
was the motive m ore frequently stated, in fact
15These provisions do not apply to home purchases, nor to
refinancings with the same lender for an amount equal to
or less than the unpaid principal balance. The Act exempts
from right-of-rescission provisions “ residential mortgage
transactions,” which are defined in the Act as extensions of
credit to acquire a principal residence. In May 1987, at the
request of mortgage market participants, refinancings with
the same lender were exempted from Regulation Z. At the
time, it was felt that this change likely would significantly
reduce the number of refinancings subject to right-ofrescission provisions.
16On the eligibility of lawyers to hold a client’s funds in OCD
deposits, see section 2-341 of the Fed’s Regulation D in
Federal Reserve Regulatory Service (1993). Client funds
also may be placed in MMDA deposits, although the rul
ings contained in section 2-341 perhaps suggest a prefer

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

at refinancing m ore households tended to forego
a shorter m aturity in favor of low er monthly
payments, consistent with reducing the im por
tance of home equity in their portfolios. (Unfor
tunately, the survey did not ask about the
withdrawal of home equity at refinancing.)
Home equity lending at banks, shown in figure
8, also has been w eak since mid-1991, with
reports suggesting that hom eow ners are indeed
repaying outstanding home equity loans with
cash w ithdraw n at the time of a mortgage
refinancing.
W hile the grow th in OCDs likely reflects
changes in households' deposits, some profes
sionals and small businesses also may account
for a portion of the increase. Some real estate
payment practices tend to increase the demand
for OCDs w hen m ortgage activity increases. The
1969 Tru th in Lending Act, for example, im
plem ented through the Federal Reserve’s Regu
lation Z, requires a three-day, right-of-rescission
period for any new credit transaction secured
by the borrow er's principal residen ce.15 During
this period, settlem ent agents typically hold
funds in a liquid deposit, or perhaps in the
form of cashier's and officers' checks. If the
funds are held solely for the beneficial interest
of the household, they may be placed in an
OCD accou nt.16 Cashier's and officers’ checks
issued by banks are included as demand deposits
in M l, while such checks issued by thrifts typi
cally are included in OCDs.
This supportive yet largely circum stantial evi
dence leaves a num ber of unansw ered questions.
If a household extracts funds at refinancing to
repay a home equity loan, how long will it keep
the funds in a liquid deposit? And isn’t the
amount of funds almost surely far smaller than
the amounts held by m ortgage servicers, associ
ated with MBS refunding activity? If so, can the
ence to hold the funds as OCDs. OCDs have no restrictions
on the number of third-party withdrawals per month. While
both OCDs and MMDA deposits are included in M2, data
on MMDAs have not been collected by the Federal
Reserve System since September 1990. Banks and thrifts
began reporting that month only a combined total for all
savings and MMDA deposits. Hence, no separate analysis
of MMDA deposits is shown in this article.

Figure 8
Refinancing Home Equity Lending
Billions of dollars

Billions of dollars
Monthly data

-60
-50
-40
-30
■20
10

-10

1990

1991

increasingly parallel m ovem ents in OCDs reason
ably be attributed to refinancing activity? On
balance, while the sharp increase in the correla
tion betw een the changes in OCDs and demand
deposits since 1991 suggests an underlying rela
tionship to mortgage refinancing, the magnitude
of any effect on the m onetary aggregates rem ains
uncertain and a convincing explanation elusive.

1992

1993

rules governing the custodial accounts of m ort
gage servicers. The m echanism generating
parallel high-frequency m ovements in OCDs,
however, is far less clear. The coincidence of its
timing with changes in refinancing activity and
the onset of unusual weakness in home equity
lending in 1992 suggest that it may be related
to the ongoing restru cturing of household
balance sheets during the 1990s.

SUMMARY
Any factors that increase the demand for tran s
action deposits can distort the grow th of the
m onetary aggregates over significant periods of
time. Recent waves of mortgage refinancing ac
tivity have caused significant fluctuations in li
quid deposits and M l. Under cu rren t Federal
Reserve operating procedures for controlling the
grow th of M2, such transitory changes in the
demand for liquid deposits, like those associated
with mortgage refinancing, are automatically ac
commodated through changes in bank reserves,
leading to increased volatility of M l.
A large portion of this increased volatility of
demand deposits can be traced to fiduciary

REFERENCES
Board of Governors of the Federal Reserve System. Federal
Reserve Regulatory Service, volume 1 (October 1993).
_______ Guide to the Flow of Funds Accounts (Board of
Governors of the Federal Reserve System, 1993).
_______ “ Improving the Monetary Aggregates,” Report of the
Advisory Committee on Monetary Statistics (1976).
_______ “ Money Stock Revisions,” supplement to the Federal
Reserve statistical release, H.6 (February 4, 1993).
Downs, Anthony. The Revolution in Real Estate Finance (The
Brookings Institution, 1985).
Duca, John V. “ The Impact of Mortgage Activity on Recent
Demand Deposit Growth,” Economics Letters (February
1990), pp. 157-61.

JULY/AUGUST 1993

Federal National Mortgage Association. 1992 Annual Report
(Federal National Mortgage Association, 1992).
. Fannie Mae National Housing Survey 1993 (Federal
National Mortgage Association, 1993).
Karcher, Louis J. Processing Mortgage-Backed Securities
(New York Institute of Finance, 1989).

Pepper, Gordon. “ Controls and Distortions: Monetary Policy in
the 1970s,” National Westminster Bank Quarterly Review
(February 1992), pp. 58-72.
_______ “ Monitoring the Money Supply and Distortions to
Monetary Data,” National Westminster Bank Quarterly
Review (February 1993), pp. 40-58.

Meulendyke, Ann-Marie. “A Review of Federal Reserve Policy
Targets and Operating Guides in Recent Decades,” in Inter
mediate Targets and Indicators for Monetary Policy (Federal
Reserve Bank of New York, 1990), pp. 452-73.

Thornton, Daniel. “ The Borrowed-Reserves Operating Proce
dure: Theory and Evidence,” this Review (January/February
1988), pp. 30-54.

Pavel, Christine. “ Securitization,” Federal Reserve Bank of
Chicago Economic Perspectives (July/August 1986),
pp. 16-31.

Topping, S.L., and S.L. Bishop. “ Breaks in Monetary Series,”
Bank of England Discussion Paper, Technical Series No.
23 (February 1989).

Appendix
Estimates of Mortgage Servicers’ Custodial
Account Balances
This appendix employs methodology suggested
by Duca (1990) to estim ate the im pact of r e
financing on the am ount of liquid deposits held
by m ortgage servicers.1 At refinancing, the out
standing principal of an extinguished securitized
mortgage is retu rned to the m ortgage servicer.
Following rules established by the federal agency
that guaranteed the MBSs issued against the pool
containing the mortgage, servicers place incom
ing unscheduled payments in custodial accounts
(liquid deposits) until rem itted to the holders of
the MBSs around the middle of the following
month. Since the servicing rules of the three
agencies differ, the overall increase in custodial
deposits that follows an increase in refinancing
depends on the agency composition of MBS liq
uidations. D ifferences in this composition during
1 9 9 1 -9 3 relative to earlier periods have attenu
ated the deposit im pact of recen t MBS liquida
tions from what might have been expected.
GNMA-guaranteed issues made up about onehalf of aggregate liquidations during 1 9 8 6 -8 7 ,

1The model in this appendix differs from Duca’s in some
respects, including assuming a more uniform rate of mort
gage closings during each month and that funds remain in
liquid deposits somewhat longer during the month follow
ing the refinancing before they are withdrawn by investors.
2
The exact monthly scheduled amortization rate is a func
tion of the outstanding balances, rates and terms on the
mortgages in the pool. Such calculations require extensive
databases well beyond the scope of this study. An alterna

http://fraser.stlouisfed.org/
FEDERAL RESERVE BANK OF ST. LOUIS
Federal Reserve Bank of St. Louis

for example, but only one-quarter in 1 9 9 1 -9 3 .
The largest volume of liquidations during
1 9 9 1 -9 3 , on balance, has been FHLMC issues
that have a smaller impact, dollar for dollar, on
liquid deposits than liquidations of GNMA- or
FNMA-guaranteed MBS.
The increase in liquid deposits due to MBS liq
uidations is estimated from a simple simulation.
The param eters are:
T he proportion of MBS liquidations during a
m onth that result from scheduled am ortization
of principal (n o r m __ liq). Separation o f sched
uled from unscheduled paym ents m atters fo r
reasons explained in the text. Estim ates in this
article assum e that scheduled principal pay
m ents equal 1 percent (at an annual rate) of the
outstanding stock of M BSs.2
T he average nu m ber o f days, expressed as a pro
portion of the m onth, th at unscheduled principal
paym ents are held in custodial accounts during
the m onth in w hich the refinancing occu rred
(G N M A __ th is ___ month, F N M A _ _ th is ___

tive set of estimates that assumed scheduled monthly
amortization equal to 2 percent of outstanding aggregate
principal had a relatively large number of months wherein
actual principal payments were less than estimated sched
uled payments and, hence, was rejected as implausible.

month, FHLMC__th is ___month). Payments
received by servicers early in the m onth have
larger im pacts on month-average deposit levels
than paym ents received late in the month.
Herein it is assumed that m ortgages close at a
uniform rate during the month. Under this as
sumption, the weighted average holding period
fo r payments received by GNMA and FNMA
_
servicers is 0.50 m onths, i.e., GNMA__th is _
month = FNMA__t h is _ month = 0.50. For
_
FHLMC servicers, who generally hold unsche
duled principal paym ents fo r five days or less,
an average holding period o f 0.15 m onths is
assumed.
T h e average num ber of days, expressed as a pro
portion o f the m onth, th at funds due to MBS
investors are held in liquid deposits during the
m onth following the refinancing (GNMA__last
__month, FNMA_ la s t___month). Under
_
GNMA and FNMA servicing rules, unscheduled
principal paym ents received by servicers during
the preceding month are rem itted to investors
on the 15th and 18th of the cu rren t month, re
spectively. Funds may b e on deposit longer if
investors do not w ithdraw them immediately.
Values of 20 days and 23 days, corresponding
to GNMA__la s t _ month = 0.67 and FNMA_
_
_
la s t __month = 0.77, are used in th e calcula
tions below . T hese som ew hat longer periods
w ere suggested by exam ination of daily deposit
data reported to the Federal Reserve by several
large banks. For FHLMC, this is set equal to
zero.
For FNMA servicers, the proportion of incom
ing funds placed in MMDAs ra th er than demand
deposits (MMDA__share). A value o f 0.25 is as
sumed below .3 Funds in MMDAs are assumed to
rem ain on deposit fo r the same num ber of days
as funds placed in demand deposit accounts.

Monthly liquidation o f GNMA-guaranteed MBSs
equals, by definition, the am ount of GNMAguaranteed MBSs issued during the month minus
the change in the amount of GNMA-guaranteed
MBSs outstanding as of the end of each month:
GNMA _

liq = GNMA _

iss - AGNMA _

stk.

In turn, the am ount of unscheduled principal pay
m ents received by GNMA servicers during a
month is assumed to equal the liquidations of
GNMA-guaranteed MBSs minus 1 percent of the
amount of GNMA-guaranteed MBSs outstanding
at the end of the previous month:
G N M A __un = G N M A ___ liq - n o r m ___liq*
G N M A __s t k ___lag.

Liquidations and unscheduled principal pay
ments for FNMA and FHLMC are calculated in
the same m anner.
The amount of demand deposits that are cus
todial account balances due to GNMA mortgage
servicers is calculated as:
GNMA __dda = G N M A ___ t h i s __ month*GNM A
__un + G N M A ___ l a s t ___ m on th *G N M A ___ un
—

la
g.

For FNMA servicers, the amount is:
FNMA _ dda = (1 -MMDA _ share)*(FNM A
11]is__m on th * FNMA ___ un + F N M A ___ last
__m on th * F N M A ___ u n ___lag);
and for FHLMC it is:
FHLMC _
FHLMC _

d d a = FHLMC _
un

this _

m onth *

A similar calculation is made fo r the holdings of
MMDAs by FNMA servicers.
An MBS-adjusted, not seasonally adjusted
(n.s.a.) demand deposit series is obtained by
subtracting the sum (GN M A__ dda + F N M A ___
dda + F H L M C __dda) from published n.s.a.
monthly levels of demand deposits. The resulting
demand deposit series is seasonally adjusted us
ing the seasonal factors for demand deposits
published by the Federal Reserve Board staff in
M oney Stock R evisions (1993). (Seasonal factors
are recovered from the published data by divid
ing the n.s.a. level by the s.a. level.) The differ
ences in grow th rates of demand deposits and
M l shown in the upper two panels of figure 6
are calculated from published and these adjust
ed data.
An MBS-adjusted, non-M l com ponent of M2 is
obtained by subtracting the estimated amount
of MBS-related MMDA deposits from the pub
lished, seasonally adjusted, non-M l com ponent
of M2. (Since the non-M l com ponent of M2 is
seasonally adjusted by the Federal Reserve
Board staff as a whole, and separate data on
MMDA are not available, the seasonally adjusted
series was adjusted by MBS effects.) The grow th
rates shown in the low er panel of figure 6 are
calculated from published M2 and from the sum
of the MBS-adjusted M l and non-M l com po
nents of M2.

3The value of 0.25 is from Duca (1990).

JULY/AUGUST 1993

F e d e ra l R e s e rv e B a n k o f St. L o u is
Post Office Box 442
St. Louis, Missouri 63166

The R e v ie w is published six
times per year by the Research
and Public Information
Department o f the Federal
Reserve Rank o f St. Louis.
Single-copy subscriptions are
available to the public fr e e o f
charge. Mail requests f o r
subscriptions, back issues, or
address changes to: Research
and Public Information
Department, Federal Reserve
Bank o f St. Louis, P.O. Box 442,
St. Louis, Missouri 63166.
The views expressed are those
o f the individual authors and do
not necessarily reflect official
positions o f the Federal Reserve
Rank o f St. Louis or the Federal
Reserve System. Articles herein
may be reprinted provided the
source is credited. Please provide
the Bank’s Research and Public
Information Department with a
copy o f reprinted material.

Full text of Review (Federal Reserve Bank of St. Louis) : July/August 1993

FRASER