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Will the Removal of Regulation Q
Raise Mortgage Interest Rates?
R. ALTON GILBERT

EGISLATION passed in March 1980 calls for
the gradual phase-out of interest rate ceilings on
deposits by 1986. Some critics of this change have
claimed that banks and thrift institutions will charge
their borrowers higher interest rates once these
deposit interest rate ceilings are removed. Accord
ing to these critics, lenders will raise their lending
rates to cover their increased deposit costs.1

Regulation Q of the Federal Reserve and, therefore,
are commonly referred to as Regulation Q. One of
the primary reasons for imposing ceilings on deposit
interest rates was to reduce the number of failing
banks by reducing their interest cost. Another ob
jective was to reduce the incentives for rural banks to
hold large interest-earning balances with their cor
respondents in the financial centers.2

This article presents a brief history of deposit in
terest rate ceilings in the United States and their
effects. It then describes the process established by
recent legislation for eliminating ceilings, and its
likely impact on the interest rates that borrowers will
pay. Finally, the analysis is extended to cover the
effects of the All Savers Certificate program on in
terest rates that depository institutions will charge
on loans.

Much of the concern in the early 1930s centered
on interest payments on demand deposits. Interest
payments on demand deposits were prohibited
under the Banking Acts of 1933 and 1935. The max
imum interest rate on all time and savings deposits
was initially set at 3 percent, slightly below the
average interest rate that commercial banks and
thrift institutions had been paying on time and sav
ings deposits, but abov e then-existing market yields
on high-grade short-term securities.3 The choice of
the initial ceiling rate on time and savings deposits
indicates that the purpose of these ceiling rates on
time and savings deposits was not to keep them
below yields on alternative investments, but to re
duce deposit rates slightly and thus lower the interest
costs of depository institutions.

W H Y HAS THE F E D E R A L G O V E R N
M E N T R E G U L A T E D D E P O S IT
IN T E R E ST RATES?
Federal bank regulators received the legal author
ity to regulate interest rates that commercial banks
may pay depositors in the Banking Acts of 1933 and
1935. The interest ceilings have been set under
‘The view that the elim ination of ceilin g interest rates on deposits
would cause interest rates paid by borrowers to rise appears in
D ep o sit o r y Institu tions D ere g u latio n Act o f 1979, Hearing on S.
1.347 before the Subcom m ittee on Financial Institutions, Senate
Committee on Banking, Housing, and Urban Affairs, Part II and
Part I II , 96 Cong. 1 Sess. (Governm ent Printing Office, 1979).
See comments by Ralph W. Pritchard, first vice president, Na
tional Association o f Realtors (June 27, 1979); Thomas F. Bolger,
first vice president, Independent Bankers Association (July 18,
1979); and Henry B. Schechter, director, D epartm ent of Urban
Affairs, A FL -C IO (July 18, 1979).

During the 20 years from the mid-1930s to the
mid-1950s, the ceiling rates on time and savings
deposits were above market interest rates. In 1957
and 1962, when market interest rates rose near or
above the ceiling rates on savings deposits, these
ceilings were raised (chart 1).
2Albert H. Cox, Jr., R egu lation o f I n te res t R a tes on B an k D e p o s
its, M ichigan Business Studies, vol. X V II, no. 4 (Bujreau of Business Research, University of M ichigan, 1966).
3Charlotte E. Ruebling, “The Administration of Regulation Q ,”
this R ev iew (February 1970), pp. 30-31.

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

C h art I

3 - M o n t h T re a su ry B ill R a te a n d C e ilin g R a te o n S a v i n g s D e p o s it s a t C o m m e rc ia l B a n k s

[^ Q u a r t e r ly a v e r a g e s
L a te st d a t a p lo t t e d :3 r d q u a r t e r 1981

In 1966, interest rate ceilings were imposed on
deposits of thrift institutions. Sponsors of the en
acting legislation asserted that interest rates were
being driven up by competition for deposits among
banks and thrifts, and that ceiling interest rates on
deposits at thrift institutions would stop this escala
tion. They assumed that by permitting slightly higher
ceiling rates at thrift institutions specializing in resi
dential mortgage lending, there would be an ade
quate supply of credit for residential mortgages at
reasonable mortgage interest rates.4
These controls on interest rates paid by thrift
institutions were viewed initially as temporary mea
sures to deal with “unusual circumstances.” Over
time, however, thrift institutions have come to view
the differentials between the ceiling interest rates
on their deposits and those imposed on commercial
banks as essential in attracting deposits to be used

‘T e m p o r a r y In te res t R a te Controls, Report No. 1777, House
Com m ittee on Banking and Currency, 89 Cong. 2 Sess. (GPO,
1966); and In terest R ates a n d M ortgage C r e d i t , Hearing on S.
3687, S. 3627 and S. 3529 before the Senate Com mittee on
Banking and Currency, 89 Cong. 2 Sess. (GPO, 1966).

Digitized for 4
FRASER

for residential mortgage lending. These differentials
have been considered important elements ol a pub
lic policy designed to expand the supply of mortgage
credit and increase residential construction.5
If the differentials in ceiling rates between thrifts
and commercial banks are to stimulate the flow of
deposits to thrift institutions, ceiling interest rates on
some categories of deposits at commercial banks
must be below market interest rates. If all deposit
interest rate ceilings were abov e market interest
rates, the higher ceiling rates at thrift institutions
would not induce individuals to hold their deposits
there rather than at commercial banks. This would
occur because both commercial banks and thrifts
would be paying the lower market interest rate to
depositors instead of the higher ceiling rates. Since
1966, the ceiling rate on savings deposits at com
mercial banks has been below the three-month
Treasury bill rate (a measure of market rates) except
for only a few months in 1967, 1971, 1972 and 197677 (chart 1).
5Preston Martin, “A Case for Regulation Q ,” J o u r n a l o f the
F e d e r a l H o m e L o a n B a n k B o a r d (O ctober 1970), pp. 1-6.

FEDERAL RESERVE BANK OF ST. LOUIS

T H E E F FE C T S O F D E P O S IT
IN T E R E ST RATE C E IL IN G S
If maintaining deposit interest rate ceilings below
market interest rates, with slightly higher rates al
lowed for thrift institutions, was intended to produce
a stable supply of mortgage credit available to
homebuyers at moderate interest rates, it has failed
to do so. The growth of deposits at thrift institutions
has slowed whenever market interest rates have
risen above the deposit ceiling rates.6 These fluctu
ations in the growth of deposits at thrift institutions
may have contributed to the abrupt changes in the
pace of residential construction activity in recent
decades.7
Deposit interest rate ceilings have discriminated
against the relatively less wealthy savers.8 There are
no ceiling rates on deposits in denominations of
$100,000 or more. The ceiling rate on money market
certificates (time deposits with maturities of six
months) fluctuates with market interest rates, but
those require a minimum deposit of $10,000. Debt
obligations of the U.S. Treasury, investments with
risk characteristics most similar to deposits of fed
erally insured institutions, are sold in minimum
denominations that are substantially larger than the
average time or savings deposits of individuals.
6Edward F. M cKelvey, I n te res t R a te C eilin g s a n d D is in ter
m ed iatio n , Staff Econom ic Studies 99 (Board of Governors of the
Fed eral Reserve System, 1978).
7Dwight M. Jaffee and Kenneth T. Rosen, “ Mortgage Credit
Availability and Residential Construction,” B ro o k in g s P ap e rs on
E c o n o m i c A ctivity (2: 1979), pp. 333-76; and Neil G. Berkman,
“ Mortgage Finan ce and the Housing C y cle,” N ew E n g l a n d
E c o n o m i c R ev iew (September/October 1979), pp. 54-76. Results
of some studies, however, do not support the view that changes in
the availability of mortgage credit through thrift institutions
influence residential construction. See Francisco Arcelus and
Allan H. M eltzer, “The Markets for Housing and Housing Ser
vices, " J o u r n a l o f Money, C r e d it a n d B an kin g (February 1973),
pp. 78-99; Allan H. M eltzer, “Credit Availability and Econom ic
D ecisions: Some E vid ence from the Mortgage and Housing
Markets, ’’J o u r n a l o f F in a n ce (June 1974), pp. 763-78; and Paul
D e Rosa, “ Mortgage Rationing and Residential Investm ent:
Some Results from a Brainard-Tobin Model " J o u r n a l o f Money,
C r e d it a n d B a n k in g (February 1978), pp. 75-87.
8Edward J. Kane, “ Short-Changing the Small Saver: Federal
Governm ent Discrim ination against Small Savers during the
Vietnam W ar,” J o u r n a l o f Monet/, C r e d it a n d B an kin g (No
vem ber 1970), pp. 513-22; Edward J. Kane, “Consequences of
Contemporary C eilings on Mortgage and Deposit Interest Rates
for H ouseholds in D ifferen t E cono m ic C ircu m stan ces,” in
George M. von Furstenberg, ed.. T he G o v e r n m e n t a n d C a p ita l
F o rm a tio n (Ballinger Publishing Company, 1980), pp. 401-41;
Charles C lotfelter and Charles Lieberm an, “On the D istribu
tional Impact o f Federal Interest Rate Restrictions,” J o u r n a l o f
F in a n ce (March 1978), pp. 199-213; Edward C. Law rence and
Gregory E. Elliehausen, “The Impact of Federal Interest Rate
Regulations on the Small Saver: Further E vid en ce," J o u r n a l o f
F in a n ce (June 1981), pp. 677-84.

DECEMBER 1981

Consequently, savers with less than $10,000, who
want an investment with risk and liquidity charac
teristics similar to Treasury bills, are limited to
savings deposits at federally insured institutions.
Because of the interest rate ceilings on these de
posits, the yield is generally less than that available
on Treasury bills. Several studies have estimated
that savers have “lost” several billion dollars in
earnings as a result of the Regulation Q ceilings.9

E L IM IN A T IN G R E G U L A T IO N Q
One of the most significant sections of the D e
pository Institutions Deregulation Act of 1980 calls
for the elimination of ceilings on deposit interest
rates over a six-year period. The statement of find
ings and purpose of that section of the act reads as
follows:
The Congress hereby finds that —
(1) lim itations on the interest rates which are payable on
deposits and accounts discourage persons from saving
money, create inequities for depositors, impede the ability
of depository institutions to com pete for funds, and have
not achieved their purpose of providing an even flow of
funds for home mortgage lending; and
(2) all depositors, and particularly those with modest savings,
are entitled to receive a market rate o f return on their
savings as soon as it is econom ically feasible for depository
institutions to pay such rate.10

The act does not specify a timetable for elim
inating deposit interest rate ceilings, but delegates
those decisions to a newly created committee: the
Depository Institutions Deregulation Committee
(DIDC). Voting members of the DIDC include: Sec
retary of the Treasury; and chairmen of the Federal
Reserve Board, Federal Deposit Insurance Corpora
tion, Federal Home Loan Bank Board, and National
Credit Union Administration. The Comptroller of
the Currency is a non-voting member of the DIDC.
9Bruce W. Morgan, “Ceilings on Deposit Interest Rates, the
Saving Public and Housing F in an ce,” E q u i t y f o r th e Sm all
Saver, Hearings on S.Con.Res. 5 before the Subcom m ittee on
Financial Institutions, Senate Com m ittee on Banking, Housing,
and Urban Affairs, 96 Cong. 1 Sess. (GPO, 1979), p. 175; David H.
Pyle, “T he Losses on Savings Deposits from Interest Rate Reg
ulation,” B ell J o u r n a l o f E c o n o m i c a n d M an ag e m e n t S cience
(Autumn 1974), pp. 614-22; David H. Pyle, “ In te re st Rate
C eilings and Net W orth Losses by Savers,” in K enneth E .
Boulding and Thomas Frederick Wilson, eds., R edistribution
through the F in a n cia l System (Praeger Publishers, 1978), pp.
87-101; Robert A. Taggart, Jr., “E ffects o f Deposit Rate C eilings:
The E vid ence from Massachusetts Savings Banks,” J o u r n a l o f
Money, C r e d it a n d B a n k in g (May 1978), pp. 139-57.
loD ep o sitor y Institutio ns D ere g u latio n a n d M onetary C ontrol
A cto o f 1980, S. Rept. No. 96-640, 96 Cong. 2 Sess. (GPO, 1980),
title 11, sec. 202(a).

FEDERAL RESERVE BANK OF ST. LOUIS

The act directs the D ID C to provide for the
orderly phase-out and ultimate elimination of maxi
mum interest rates that may be paid on time and
savings deposits as rapidly as economic conditions
warrant. A primary consideration in determining
when conditions warrant raising or eliminating
these ceilings is the effect of such changes on the
safety and soundness of depository institutions. The
act lists the following methods the DIDC may use in
phasing out ceiling interest rates on deposits:
T he phase-out of' such limitations may be achieved by the
D eregulation Com m ittee by the gradual increase in such
lim itations applicable to all existing categories of accounts, the
com plete elim ination of the limitations applicable to par
ticular categories of accounts, the creation of new categories of
accounts not subject to limitations or with lim itations set at
current market rates, any com bination of the above methods, or
any other method.11

One limitation imposed on the DIDC is that it may
not raise interest rate ceilings on all deposit cate
gories above market interest rates before March
1986.
The DIDC has taken limited actions to raise or
eliminate ceilings on deposit interest rates (see table
1). The first significant action was to lift caps on
ceiling rates for time deposits with maturities of 2 ‘/2
years, which was effective August 1, 1981. The
DID C has also created a new category of IRA/Keogh
account (with minimum maturity of IV2 years) that
will have no regulated interest rate ceiling as of
January 1, 1982.

THE E F FE C T S O F E L IM IN A T IN G
R E G U L A T IO N Q O N IN T E R E ST
RATES P A ID RY R O R R O W E R S
The effects of eliminating ceiling rates on deposits
cannot be determined by examining the effects of
actions already taken by the DIDC, since few actions
to eliminate the ceiling rates have been taken so far.
Effects of eliminating deposit ceiling rates on the
interest rates paid by borrowers must, therefore, be
analyzed by considering the effects of eliminating
Regulation Q in the context of a theory that describes
how interest rates are determined.

The Mark-up Theory vs. the Competitive
Market Theory
There are several competing theories of how
depository institutions determine the interest rates
they charge borrowers. The two theories discussed
“ Ibid., title II, sec. 204(a).

DECEMBER 1981

in this section have different implications for the
impact of eliminating the ceiling rates on time and
savings deposits specified under Regulation Q.
The M ark-up Theory — Those who assert that
borrowers will be charged higher interest rates due
to the elimination of Regulation Q are generally
using a mark-up theory: Depository institutions are
presumed to determine the interest rates they charge
borrowers as a mark-up over the a verage interest rate
they pay on deposits. The average interest rate on
deposits will rise as Regulation Q is phased out,
unless market interest rates should fortuitously fall
below the Regulation Q ceilings currently in effect.
The mark-up theory, therefore, predicts that bor
rowers will pay higher interest rates as a conse
quence of the elimination of Regulation Q.
The C om petitive M arket Theory — Under this
theory, the interaction of several factors influencing
both supply and demand determine a market inter
est rate, which all lenders charge on loans with
similar characteristics. Lenders can make few loans
at interest rates above the market rate, since bor
rowers will search for the lowest rate available.
Since lenders can make all the loans they wish at the
market rate, they have no incentive to lend at in
terest rates below the market rate.
To describe this theory in more detail, consider
the determinants of the market interest rate on a
particular category of credit — residential mortgage
loans. Demand for residential mortgage credit is
determined by personal income and the preferences
of individuals for housing and for home ownership.
Several factors influence the supply of residential
mortgage credit. One factor is the interest rates on in
vestments other than residential mortgages. If, for
instance, yields rise on U.S. Treasury securities with
maturities similar to those of residential mortgages,
depository institutions and other suppliers of resi
dential mortage credit will supply less mortgage
credit at each level of the mortgage interest rate.
Another important determinant of supply is the
interest rate on deposits not subject to Regulation Q
ceilings. For example, depository institutions may
pay whatever interest rate they wish on time de
posits in denominations of $100,000 or more. In the
competitive market, depository institutions will bid
up the interest rates they are willing to pay on
deposits fr e e o f Regulation Q ceilings until these
rates are sufficiently close to their lending rates to
eliminate the incentives to make additional loans.
Consequently, it is the interest rate that depository

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS

Table 1

DIDC Changes in Interest Rate Ceilings on Deposits
Date of Meeting
May 29, 1980

Effective
Date of Change
June 2, 1980

Type of Deposit

Nature of Change

Small savers certificates
(time deposits with
m aturities of 30 months
or more, no m inimum
denom ination)

P rio r ceilings: Com mercial banks were perm itted to
pay the yield on 21/2-year Treasury securities less 75
basis points, and th rift institutions were perm itted to
pay 25 basis points more than com m ercial banks. The
maximum interest rates perm issible, however, were
11.75 percent at com m ercial banks and 12 percent at
th rift institutions.
Changes: Ceiling rates relative to yield on 21/2-year
Treasury securities raised 50 basis points. Ceiling rates
w ill not fall below 9.25 percent at com m ercial banks
or 9.50 percent at th rift in stitu tion s. The maximum
ceiling rates of 11.75 and 12 percent were retained.

May 29, 1980

June 5, 1980

Money market certificates
(time deposits in denom
inations of $10,000 or
more with m aturities of
six months)

Raised the ceiling rate from the discount yield on sixmonth Treasury bills established at the most recent
auction to that rate plus 25 basis points at both com
m ercial banks and th rift in stitu tion s.1

O ctober 9, 1980

December 31, 1980

NOW accounts

Set the ceiling rate on NOW accounts at 5.25 percent
fo r com m e rcial banks, m utual savings banks, and
savings and loan associations. The ceilin g rate on
interest-bearing checkable deposits was 5 percent
until December 31, 1980.

June 25, 1981

August 1, 1981

Small savers certificates

Eliminated caps on these ceiling rates of 11.75 percent
at com m ercial banks and 12 percent at th rift institu
tions. W ith the caps lifted, th rift institutions may pay
the yield on 21/2-year Treasury securities, and com
mercial banks may pay 25 basis points less.

September 3, 1981

O ctober 1, 1981

All Savers Certificates

Adopted rules fo r All Savers Certificates specified in
the Econom ic Recovery Act of 1981.

September 22,1981

November 1, 1981

Money market certificates

Depository institutions are now perm itted to pay the
higher of the discount rate on six-m onth Treasury bills
at the most recent auction, plus 25 basis points, or
the average auction rate in the past four weeks, plus
25 basis points.

S ep tem be r22 ,1981

January 1, 1982

IRA/Keogh accounts

Created a new category of IRA/Keogh account with
m inim um m aturity of 11/2 years, no regulated interest
rate ceiling, and no m inim um denom ination.

’ Other changes in the ceiling rate on money m arket certificates are relevant when the yield on six-m onth Treasury bills falls below
8.75 percent.

institutions pay on deposits u n co n stra in ed by
Regulation Q that influences the interest rates they
charge on loans.

rates for borrowers, if individuals are induced to save
more of their income in response to the higher inter
est rates available on deposits.

Under the competitive market theory, a change in
Regulation Q ceilings will affect interest rates on
residential mortgages only if it affects interest rates
on unregulated deposits or on alternative invest
ments. One implication of this theory is that elim
inating Regulation Q ceilings might reduce interest

The effects of eliminating Regulation Q under the
competitive market theory are in sharp contrast to
the effects under the mark-up theory. The mark-up
theory predicts that the elimination of Regulation Q
would cause interest rates paid by borrowers to rise,
while the competitive market theory suggests that

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

C h a rt 2

Comparison of Mortgage Interest Rate with Cost of Funds
to S&Ls and 10-Year Treasury Bond Yield 11
Per c ent

P e r c e nt

[1 S e m ia n n u a l a v e ra g e s
Latest d a ta plotted: 1st h a lf 1981

interest rates on loans would either be unaffected or
would decline.
W hat’s the E viden ce? — Chart 2 presents some
evidence on whether U.S. interest rates on residen
tial mortgages are determined according to the mark
up or competitive market theory. The average cost of
funds to savings and loan associations (S&Ls) over
Digitized for 8
FRASER

six-month intervals since 1966 is shown in conjunc
tion with the average interest rate on conventional
residential mortgages and the average yield on U.S.
Treasury securities with maturities of 10 years over
the same six-month periods.
Chart 2 clearly indicates that there is no fixed
mark-up between the average cost of funds to S&Ls

FEDERAL RESERVE BANK OF ST. LOUIS

and the average interest rate on residential mort
gages. The difference between the average mort
gage interest rate and the average cost of funds to
S&Ls has varied widely, from 165 basis points in the
first half of 1966 to 386 basis points in the first half of
1980.
Chart 2 shows that there is a much closer rela
tionship between the average mortgage interest rate
and the average yield on U.S. Treasury securities
with maturities of 10 years than the relationship
between the mortgage interest rate and the average
cost of funds.12 The difference between the mort
gage interest rate and the yield on 10-year Treasury
bonds has a standard deviation of 27 basis points,
compared with a standard deviation of 59 basis
points for the difference between the mortgage
interest rate and the average cost of funds to S&Ls.
These comparisons provide evidence that interest
rates on residential mortgages are determined in a
competitive credit market. Homebuyers must pay
interest rates on mortgages that are competitive with
yields on alternative investments in order to receive
credit.
Chart 3 presents additional evidence on whether
interest rates are determined according to the mark
up or the competitive market theory. The difference
between the prime loan rate charged by commercial
banks and the average interest rate they pay their
depositors on total time and savings deposits is
highly variable, ranging from 49 basis points in 1972
to 461 basis points in 1980. Thus, once again, there
appears to be no fixed mark-up between the prime
rate and the average interest rate paid on time and
savings deposits.
There is a much closer relationship, however,
between the prime loan rate and the rate that com
mercial banks pay on their three-month certificates
of deposit, which are free of Regulation Q ceilings.
The differential between the prime rate and the
three-month certificate of deposit yield has a
standard deviation of 73 basis points, compared with
a standard deviation of 144 basis points for the
differential between the prime rate and the average
interest rate paid on time and savings deposits.
Again, the interest rate relationships presented in
12T he conclusion that mortgage interest rates are more closely
related to the yield on U.S. Treasury securities wit!) maturities
of 10 years than to the average cost of funds to S& Ls has been
confirmed using regression analysis. See Thomas Mayer and
Harold Nathan, “ Mortgage Rates and Regulation Q ,” Working
Paper S eries No. 171 (Departm ent of Econom ics, University o f
California at Davis, Ju ly 1981).

DECEMBER 1981

chart 3 are more consistent with the competitive
market theory than with the mark-up theory.

Is the Mortgage Market
Separate From O ther Credit Markets?
Despite the above evidence suggesting that in
terest rates charged borrowers are more closely
related to market interest rates uncontrolled by
Regulation Q than to the average interest rates paid
on deposits, the possibility that the elimination of
Regulation Q would increase the interest rates paid
by one class of borrowers — homebuyers — has not
been ruled out. This possibility, produced by certain
regulations and tax incentives affecting thrift insti
tutions, is discussed in this section.
Since 1966, the existence of higher ceilings on
their deposit interest rates have given thrift insti
tutions an advantage over commercial banks in at
tracting deposits. At the same time, however, thrift
institutions are faced with regulations that limit their
investments in types of assets other than mortgages.
In addition to these regulations, thrift institutions
are also given tax incentives to specialize in resi
dential mortgage lending: The deductions from
gross income allocated to bad debt reserves, which
are, therefore, not subject to income tax, are larger for
institutions that invest more of their assets in
mortgages.
As a result of the higher ceiling interest rates
allowable (which attract deposits) and the regula
tions and tax incentives that favor mortgage lending,
thrift institutions might charge residential mortgage
lending rates that are below market interest rates (on
securities with characteristics similar to residential
mortgages). Eliminating Regulation Q would re
move the advantage that thrift institutions have in
attracting deposits. As a result, the share of credit
channeled to residential mortgages would decline
and interest rates on residential mortgages would
rise relative to other interest rates.
This result is unlikely for several reasons. First,
the reactions by other suppliers of credit would tend
to offset these effects, as long as non-thrift institu
tions are making residential mortgage loans as well. If
thrift institutions increase the amount of mortgage
credit they offer at prevailing interest rates, other
lenders will simply reduce the quantity of resi
dential mortgage credit they supply, shifting their
investments to other sectors of the credit market.
The net result might be no change in mortgage in
terest rates, but an increase in the proportion of
9

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS

C h art 3

Relationship B etw een the Commercial Bank Prime Rate
and Selected Deposit Interest Rates
Percent

Commercial bank prime ra

Market yield on
3-month C D s /

Average interest rate on ^
time and savings de positsii

1972

1973

l i A t com m ercial b a n k s
Latest d a ta plotted: 1 9 80

residential mortgage loans made by thrift institu
tions relative to non-thrift institutions.
Of course, it is possible that the increase in the
supply of mortgage credit by thrifts may not be fu lly
offset by reductions in supply by other lenders.
Digitized for 10
FRASER

Again, however, an increase in the net supply of
residential mortgage credit would not necessarily
depress mortgage interest rates relative to yields on
alternative investments. The reason is that pre
dictable adjustments in the demand for credit would

FEDERAL RESERVE BANK OF ST. LOUIS

tend to offset the effects of this shift in the supply on
mortgage interest rates. Suppose that, initially,
interest rates on residential mortgages are decreased
relative to other market interest rates due to an
increase in the supply of deposits and mortgage
loans at thrift institutions. This triggers increases in
the quantity of mortgage credit demanded at pre
vailing mortgage interest rates until these rates are,
once again, in line with other interest rates. There
are a variety of reactions by individuals that would
cause an increase in demand for mortgage credit. For
example, those seeking to borrow to invest in
business firms would take out second mortgages on
their homes rather than seek business loans at
commercial banks. Also, individuals buying homes
would obtain mortgages with smaller percentage
downpayments, and invest their wealth instead at
interest rates higher than the rates they pay on
mortgages.
There is a simple method to test whether the
residential mortgage market is truly separate from
other credit markets. We can determine this by ex
amining the correlation between the difference of
the average mortgage interest rate and the yield on
10-year Treasury bonds with the rate of growth in
time and savings deposits at mutual savings banks
and savings and loan associations. If the correlation
is significantly negative — if the spread between the
mortgage interest rate and the 10-year bond rate
tends to narrow when time and savings deposits at
thrift institutions grow at a faster rate — the resi
dential mortgage market is, to some extent, sepa
rated from other credit markets. When their deposits
increase rapidly, thrift institutions reduce the
mortgage interest rate relative to other interest rates
in order to acquire enough residential mortgages to
retain the tax advantages from specializing in
mortgage lending.
In fact, the correlation between the interest rate
spread and the growth rate of time and savings
deposits at thrift institutions is positive. Using
monthly observations from January 1968 through
July 1981, the correlation coefficient is 0.234, which
is statistically significant at the one percent level.
Using quarterly averages for 1/1968 through 11/1981,
the correlation coefficient is 0.262, which is not sta
tistically significant at the five percent level.
This result confirms the conclusion reached in the
previous section. The competitive market theory is
consistent with the actual behavior of interest rates.

DECEMBER 1981

Therefore, eliminating Regulation Q would not
affect mortgage interest rates adversely.

IM P L IC A T IO N S F O R IN TEREST
RATES O F ALL
SAVERS C E R T IF IC A T E S
The analysis presented above has implications for
the effects of the All Savers Certificate (ASC) pro
gram on interest rates paid by borrowers at depos
itory institutions. ASCs are special time deposits
with maturities of one year. The ceiling rate on ASCs
is equal to 70 percent of the average yield set in the
most recent auction of one-year Treasury securities.13
Individuals may declare up to $1,000 in interest on
ASCs tax free (up to $2,000 on joint returns).
Depository institutions issuing ASCs are receiv
ing deposits at interest rates below market rates. For
individuals subject to relatively high marginal tax
rates, the tax-free yield on ASCs is greater than the
after-tax return on many alternative investments.
Depository institutions are required to invest 75
percent of the funds raised by issuing ASCs in
housing and agricultural loans. Details of the legis
lation and the regulations issued to implement the
program provide depository institutions with a great
deal of flexibility in meeting these investment re
quirements. The objectives for establishing the ASC
program, however, included increasing the amount
of credit available to the housing and agricultural
sectors of the credit market.
The structure of regulations under the ASC pro
gram is similar to that for promoting mortgage
lending by thrift institutions. Differentials in Reg
ulation Q ceilings have given thrift institutions
advantages in attracting deposits, and thrifts have
been given tax incentives to specialize in mortgage
lending. All depository institutions that take ad
vantage of the ASC program to attract deposits at
interest rates below market rates are required to
allocate increases in their assets to certain sectors
of the credit market.
The analysis developed earlier indicates that the
inflow of deposits at institutions given inducements
to specialize in mortgage lending has not lowered
the level of mortgage interest rates relative to other

13Treasury secrities with maturities of one year are generally
auctioned every four w eeks on a Thursday. T h e average yield
on a Thursday auction determines the new ceilin g rate on All
Savers Certificates beginning the following Monday.

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS

rates. The ASC program is therefore unlikely to have
any appreciable impact on the interest rates charged
on housing and agricultural loans relative to other
interest rates. The ASC program may have some
effect on the quantity of housing and agricultural
loans, as depository institutions and their borrowers
develop methods of classifying loans in the cate
gories that will meet the investment requirements
of the ASC program. The primary effects of the ASC
program will be to reduce the interest costs of depository institutions and the income tax of investors.

been the opening of branch offices. Since depository
institutions have not been allowed to compete di
rectly on the basis of interest rates they offer to pay
on deposits, they have been competing indirectly by
offering convenient locations for depository ser
vices.16 Many branches that were profitable when
Regulation Q ceilings were below market interest
rates will become unprofitable when deposit inter
est rate ceilings are lifted.

THE E F FE C T S O F E L IM IN A T IN G
R E G U L A T IO N Q O N PROFIT S O F
D E P O S IT O R Y INSTITUTION S

Under the directives of the Depository Institu
tions Deregulation Act of 1980, the Depository In
stitutions Deregulation Committee is in the process
of lifting interest rate ceilings on time and savings
deposits. That committee has taken some steps to
raise the ceilings, but the most significant actions to
eliminate the ceilings on deposit interest rates are
yet to come.

Eliminating Regulation Q will raise the interest
rates paid to depositors relative to market interest
rates. The evidence cited above indicates that bor
rowers at depository institutions will not pay higher
interest rates due to the elimination of Regulation Q.
Decontrol of interest rates paid on deposits, there
fore, will tend to reduce the net income of depository
institutions.
Several studies indicate, however, that the net
income of depository institutions will not decline by
the full amount of the increase in interest paid on
deposits. Because interest rate ceilings on deposits
have been below market interest rates, depository
institutions have increased expenditures to attract
deposits by means other than increasing interest
payments on deposits.14 These non-interest expen
ditures to attract deposits are estimated at between
40 and 50 percent of the direct interest expense
depository institutions saved by paying only the
ceiling interest rates on deposits rather than market
interest rates.15
Although depository institutions can quickly
eliminate some types of non-interest expenditures
made to attract deposits, such as gifts of merchandise
for depositors who open or add to accounts, they will
incur losses in eliminating other expenditures. One
major expenditure intended to attract deposits has
14Thomas E ric K ilcollin and Gerald A. Hanweck, “ Regulation Q
and C om m ercial Bank P ro fitab ility ,” R e search Papers in
Banking and Financial Econom ics (Board ot Governors of the
Fed eral Reserve System, 1981).
15Taggart, “ E ffects o f D eposit Rate C eilin g s,” and Lew is J.
Spellman, “ D eposit Ceilings and the Efficiency of Financial
Interm ediation,” J o u r n a l o f Fin an ce (March 1980), pp. 129-36.

Digitized for 12
FRASER

C O N C L U S IO N S

Some supporters of ceilings on deposit interest
rates claim that eliminating the ceilings will cause
depository institutions to raise the interest rates diey
charge borrowers. An analysis of interest rates does
not support this view. Interest rates paid by bor
rowers are determined by market rates that are
exempt from Regulation Q ceilings. Consequently,
elimination of Regulation Q ceilings will not cause
loan rates to rise, but may cause them to decline if
depositors save more with higher deposit interest
rates. Profits of depository institutions will not
decline by the full amount of the increase in interest
expense resulting from eliminating Regulation Q,
since these institutions will eliminate some non
interest costs that were incurred to attract deposits
when Regulation Q ceilings were binding.
Similar implications also hold for the effects of the
All Savers Certificate program on interest rates paid
by borrowers. Although depository institutions are
required to invest at least 75 percent of funds raised
by issuing All Savers Certificates in housing and
agricultural loans, that requirement is unlikely to
result in lower interest rates on such loans relative to
other market interest rates.
16Law rence J. W hite, “ P rice Regulation and Quality Rivalry in a
Profit-maximizing M odel: The Case o f Bank Branching,” J o u r
nal o f M oney, C r e d it a n d B a n k i n g (February 1976), pp. 97-106;
and W illiam M. Peterson, “ The E ffects o fln te re st Rate C eilings
on the Number o f Banking Offices in the United States,” R e
search Paper No. 8103 (Federal Reserve Bank o f New York,
1981).

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

A Comparison of the St. Louis Model and
Two Variations: Predictive Performance and
Policy Implications
LAURENCE H. MEYER and CHRIS VARVARES

T
-M. HE St. Louis Model was first published in the
Federal Reserve Bank of St. Louis Review in April
1970.1 This model, with modifications, has been
used for years at the St. Louis Fed to provide alter
nate scenarios for the response of inflation, output
and the unemployment rate under different mone
tary policy assumptions. In addition, it continues to
be identified by those outside the St. Louis Federal
Reserve as the model underlying the Bank’s policy
prescriptions.

performance and policy implications of two varia
tions of the St. Louis Model — one incorporating a
Phillips Curve, the other a monetarist reduced-form
for inflation. Both versions outperform the St. Louis
Model’s inflation predictions, and both yield nearly
identical predictions and policy multipliers.
This article is organized as follows: The first sec
tion reviews the current version of the St. Louis
Model. The second section introduces two alterna
tive versions of the St. Louis-type model. The first
version substitutes a simplified Phillips Curve for
the St. Louis Model’s price-change equation; the
second version introduces a simple reduced-form
equation for inflation in place of the Phillips Curve.
The third section compares the predictive perfor
mance and policy implications of these three
models. The final section summarizes our findings.

This article has three basic themes. First, the struc
ture of the St. Louis Model can be simplified and its
predictive performance improved. Second, the St.
Louis Model’s specification of the demand slack
variable in its Phillips Curve may bias the equation’s
estimate of inflation’s response to demand slack and,
therefore, could yield an overly optimistic assess
ment of the cost of reducing inflation in terms of the
higher unemployment during the transition to a
lower rate of inflation. Third, a monetarist reduced- THE C U R R E N T V E R S IO N O F THE
form equation for inflation, in which inflation depends ST. L O U IS M O D E L
directly on current and past monetary growth, is not
The St. Louis Model consists of five estimated
inconsistent with the existence of a Phillips Curve.
This is demonstrated by comparing the predictive equations and a number of identities. The key equa
tions are the Andersen-Jordan or St. Louis nominal
The research reported here was begun w hile Laurence H.
income reduced-form equation and the equation for
M eyer was a Visiting Scholar at the Federal Reserve Bank of
the
change in the price level. There are also equa
St. Louis. The results do not necessarily reflect the views of the
St. Louis Federal Reserve Bank staff, nor should the models
tions for the unemployment rate, the long- and short
presented be view ed as new versions o f the St. Louis Model.
term interest rates, the anticipated change in the
The authors gratefully acknowledge financial support from the
price level and the change in output. The only sig
In stitu te o f Banking and F in an cial M arkets at W ashington
University.
nificant change since the model was introduced has
been the substitution of a rate of change (or dot)
1LeonalI C. Andersen and Keith M. Carlson, “ A Monetarist Model
for Econom ic Stabilization,” this R ev iew (April 1970), pp. 7-25.
version of the Andersen-Jordan nominal income

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

reduced-form equation for the original first differ
ence (or delta) version.2

The Andersen-Jordan Equation
The Andersen-Jordan equation is currently speci
fied in rate-of-change or dot form; compound annual
rates of change are used for nominal income (Y), the
money supply (M1B is the definition of money cur
rently used with the St. Louis Model, M), and the
high-employment level of government expenditures
(G). Dots over a variable indicate compound annual
rates of change.
4
(STL-1) V, = a„ +

an M „ +
i= ()

4
i

a2i G ,,

i= ()

The parameter estimates (t-values in parentheses)
for the equation estimated from 1/1955 through IV/
1980 are as follows:3
an
a io
an
a 12
a i3
an
-an

=
=
=
=
=
=
=

2.87
.46
.45
.24
.026
-.0 7 1
1.12
.44

(3.26)
(4.32)
(6.49)
(2.51)
( .398)
(-.1 2 )
(7.44)

a 20
»21
a22
a23
a2j
- « 2i

=
=
=
=
=
=

.061
.048
-.0 0 1
-.0 5
-.0 6
-.0 0 3

(
(

1.61)
1.66)
(-.0 3 4 )
(-1 .9 4 )
(-1 .7 8 )
(-0 .0 3 8 )

SE = 3.6 DW = 2.04

The coefficients on the M variables approxi
mately sum to unity while the coefficients on G
sum approximately to zero. Thus, the estimated co2The Andersen-Jordan equation was initially reported in Leonall
C. Andersen and Jerry L. Jordan, “ Monetary and Fiscal Actions:
A T est o f Their Relative Importance in Econom ic Stabilization,”
this Rev iew (November 1969), pp. 11-24. In that version, changes
in the money supply had a strong and persistent influence on
nominal income, while government expenditures had a weak
initial impact that eroded to no effect at all after a single year,
and tax changes had no effect at all. Benjam in Friedman noted
subsequently that, when data were included through m id-1976,
fiscal policy variables entered the reduced-form equation with
larger, persistent effects. (Benjam in M. Friedman, “Even the
St. Louis Model Now Believes In Fiscal Policy,” J o u r n a l o f
M oney, C re d it, a n d B an kin g (May 1977), pp. 365-67.) In a reply,
Carlson noted that the delta version o f the St. Louis equation,
when estimated with data through m id-1976, suffered from
heteroscedastieity. (Keith M. Carlson, “Does the St. Louis Equa
tion Now Believe in Fiscal Policy?” this R eview (February
1978), pp. 13-19.) He therefore reestimated the equation in dot
form, a standard approach to elim inating this problem. The dot
version produced policy effects similar to the original delta
version over the earlier time period: strong, persist monetary
effects and weak, transitory fiscal effects.
3The equation is estimated using an Almon polynomial distrib
uted lag (PD L) procedure with a fourth degree polynomial and
with coefficients o f the lag distributions restricted to zero at both
ends o f the lag distribution.

efficients support the general conclusions associated
with a monetarist viewpoint: Monetary change is the
key variable explaining nominal income movements
while fiscal variables have at best a minor and transi
tory effect.4

The Inflation Sector
The inflation sector of the St. Louis Model includes
three equations: a price-change version of a Phillips
Curve, an identity defining the anticipated change
in the price level and an equation for the long-term
interest rate. The weights in the distributed lag of
inflation in the long-term interest rate equation are
used to construct the anticipated-price-change vari
able; this variable, in turn, is included as an argu
ment in the price-change equation. This structure is
unnecessarily complicated. The predictive perfor
mance of the model with respect to inflation can be
improved with a simpler and more conventional
specification of the Phillips Curve in which the
weights on the distributed lag on past inflation are
estimated as part of the estimation of the Phillips
Curve.5

4Although the Andersen-Jordan equation has been controversial
since it was first introduced, attempts to develop more eclectic
versions allowing for a permanent effect o f fiscal variables on
nominal income have generally been unsuccessful. For a survey
o f em pirical evidence on the Andersen-Jordan equation, see
Laurence H. Meyer and Robert H. Rasche, “ Em pirical E v i
dence on Stabilization Policies,” in Sta b iliz a tio n Policies:
L esson s f r o m the 1970s a n d I m p lic a t io n s f o r the 19H()s,” Pro
ceedings of a Conference sponsored by the Center for the Study
of American Business and the Federal Reserve Bank o f St. Louis,
1980, pp. 41-102. There is, on the other hand, considerable evi
dence suggesting that simple reduced forms may yield unrelia
ble estimates o f policy multipliers. See, for example, Franco
Modigliani and Albert Ando, “ Im pact o f Fiscal Actions on
Aggregate Incom e and the M onetarist Controversy,” in Jerom e
L. Stein, ed., M on etar ism (Amsterdam, North Holland; 1976),
pp. l? -4 2 ; and Stephen M. Goldfeld and Alan S. Blinder, “ Some
Implications o f Endogenous Stabilization Policy',” B r o o k in g s
P apers on E c o n o m i c A ctivity (3:1972), pp. 585-640.
5Many of the criticism s o f the St. Louis inflation sector discussed
in this section were initially raised in comments by Nordhaus
and Gordon at the time the St. Louis Model was presented at an
N B E R conference on price determination in 1970. See Otto E ck
stein, ed., T he E c o n o m e t r ic s o f P rice D eter m in ation , Proceed
ings of a Conference sponsored by the Board o f Governors o f
the Federal Reserve System and the Social Science Research
Council, Federal Reserve System, June 1972. The St. Louis
Model was described in the volume in Leonall C. Andersen and
Keith M. Carlson, “An Econom etric Analysis o f the Relation o f
Monetary Variables to the Behavior of Prices and Unemploy
m ent,” pp. 166-183. Comments on the St. Louis Model’s model
ling of inflation appear in W illiam D. Nordhaus, “Recent Devel
opments of Price Dynam ics,” pp. 16-49; and in Robert J.
Gordon’s discussion of the Andersen and Carlson paper, pp.
202 - 12.

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

The price-change equation — The price-change
equation in the St. Louis Model is:

produce an upward bias in the model’s response of
inflation to monetary change.

A second unusual feature is that it uses a different
demand slack variable than that used in most empiri
cal Phillips Curves. Generally, either the unem
ployment rate or the (percentage) GNP gap (poten
tial or full-employment output minus actual output)
is used as the measure of demand slack. The St.
Louis demand slack variable (DSL), on the other
hand, is defined as

(STL-2) A P * t =

bo +

i
hi, DSLt-i +
i= 0

b2 AP-\

where A is the first difference operator, DSL is the
demand slack variable (defined below), APA is the
anticipated change in the price level (also defined
below), and AP*t, the change in the price level, is
specified as
(STL-2a) A P * t = AP , • X M,

where X is the level of real GNP. The explanation for
this form of the price change variable will be given
below.
The parameter estimates when the equation is
estimated over the period I/1955-IV/1980 are as
follows:

1)2

=
.65
(.77)
=
.012 (.53)
= .028 (3.02)
= .036 (5.59)
=
.038 (3.73)
=
.033 (2.97)
=
.019 (2.60)
=
.166 (5.93)
= 1.29 (25.50)

bo
bio

bn
bi2
1)1.3

bu
1)15

Sb„

.88 SE =

Although the price-change equation is, in essence,
a Phillips Curve equation, it has several unusual
features. First, it explains the first difference in the
price level (the implicit GNP deflator), while Phillips
Curves are typically specified in terms of the inflation
rate or the rate of change in nominal wages.6 Its delta
form reflects the now-abandoned delta specification
of the Andersen-Jordan equation; it made the pricechange equation dimensionally compatible with the
income-change equation, allowing the change in
output to be solved for via a simple “identity.” Since
the Andersen-Jordan equation is now used in dot
form, the retention of the delta form for the pricechange equation is unnecessary. Moreover, the delta
specification, due to the possibility of heteroscedasticity, could produce an upward bias on the coeffi
cients in that equation, including the coefficients on
both the demand slack variable and on the anticipated-price-change variable.7 These impacts would
sMore precisely, the price change variable is the change in the
price level times lagged real output. See equation STL-2a above.
7This possible source o f bias in the St. Louis price-change equa
tion was noted by Gordon in his comments on the St. Louis

(STL-2b) DSL, = AY, -

(POTRT, -

X M),

where POTRT is the level of potential output as
measured by the Rasche-Tatom series.8 This specifi
cation of the demand slack variable may seriously
bias upward the equation’s estimate of the response
of inflation to demand slack, inasmuch as it allows
changes in nominal income associated with changes
in the price level to “explain” changes in the price
level.9
The sum of the coefficients on the demand slack
variable determines the degree to which decelera
tions in monetary growth are initially reflected in
declines in the rate of growth of output and hence
increases in the unemployment rate. Meyer and
Rasche report simulations of the St. Louis Model
with different values of this parameter (its value
based on a sample through 1/1975 and its value
based on a sample through IV/1979 where the sum
is three times larger) and demonstrate the dramatic
differences in the implied responses of output and
the unemployment rate to monetary decelerations.10
M odel in the E ckstein volume. Gordon argued that the results
of the price-change equation “are plagued by heteroscedastieity”
(p. 209). In response to the presence o f heteroscedasticity, the
nominal incom e was changed from a delta to a dot specification,
although the price-change equation, where heteroscedasticity
may have been more of a problem,was left in first difference form.
8S ee Robert H. Rasche and John A. Tatom, “ Energy Resources
and Potential Output,” this R ev iew (June 1977), pp. 10-24, for a
discussion of this series.
9T he possibility that the St. Louis specification of the price-change
equation yields an upward biased estim ate of the response of
inflation to demand slack was suggested by the remarkable
behavior o f the sum o f the estim ated coefficients on the demand
slack variable as additional years o f data were added to the
sample period during the 1970s. After 1975, the estimated
coefficient begins to rise as more data is included; by the end of
1979, the coefficient is almost three times its value for the sample
period ending before 1975. This pattern is consistent with what
would be expected if the specification yielded biased estimates
for the reason suggested above. This bias would be expected to
becom e more serious during a period w here changes in nominal
incom e w ere dominated by changes in the price level.
10See Laurence H. M eyer and Robert H. Rasche, “On the Costs
and Benefits o f Anti-Inflation P olicies,” this R eview (February
1980), pp. 3-14.

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

The long-term interest rate equation —The equa
tion for the long-term interest rate (RL) is:
16

(STL-.3) RLt — Co + ci \lt + c'2 Zt +

— c.ii

Xt-i

i=0

+ i! C4i (P,.i/(U^UFt.i)),
i= 0

where X is the rate of change in real GNP, Z is a
dummy variable, allowing for a shift in the constant
term over the period, U is the unemployment rate,
and U F is a measure of the rate of unemployment at
“full employment.” The parameter estimates for the
long-term interest rate equation are as follows:
C'O
Cl
C'2
-C3i
-C ji

= .82
=
.02
=
.82
= .29
= 1.04

(1.42)
(.65)
(2.77)
(1.69)
(14.23)

.89

The measure of expected inflation in the above
equation is a distributed lag on past inflation adjusted
for the level of demand pressure as proxied by the
ratio of the unemployment rate to the full-employment
rate of unemployment (UF) where the latter is
measured by series developed at the Council of
Economic Advisors. This equation not only provides
predictions of the long-term interest rate, it also
provides the weights, the C4i coefficients, used in the
anticipated-price-change equation.
There are a number of questionable features of
this long-term interest rate equation, particularly
related to its role in providing the weights for an
expected price-change variable. First, the weighted
sum of current and past inflation rates can be viewed
as a measure of the expected inflation rate only if we
assume that a one percentage point increase in the
expected inflation rate increases the long-term interest
rate by one percentage point. We cannot, however,
separate out the weights that convert current and
past inflation rates into the expected rate of inflation
and the coefficient that translates an increase in the
expected inflation rate into an increase in the long
term interest rate. Recent work on the implications
of specific tax structures for interest rate behavior in
inflationary periods indicates that the simple Fisher
ian view that a percentage point increase in the
expected inflation rate raises the long-term interest
rate by a percentage point is no longer so obvious.11
“ See, for exam ple, Martin Feld stein, “ Inflation, Incom e Taxes
and the Rate o f Interest: A Theoretical Analysis,” A m erican

Digitized for 16
FRASER

Second, the expected-price-change variable that
is derived from a long-term bond equation is likely
to relate to a much longer horizon (the average term
to maturity of long-term bonds) than is relevant to
the formation of price expectations in the context of
the Phillips Curve (the current period or at most an
average of price change expected over the average
length of contracts, implicit and explicit). This differ
ence in horizon may affect the number of relevant
lags and the weighting applied to past inflation.
There is one additional question about the specifi
cation of the long-term interest rate equation. One
can derive a somewhat similar equation by begin
ning with a money demand equation in which
the demand for money depends on the long-term
interest rate and current and past rates of inflation
and by solving that equation for the long-term
interest rate as a function of the level of real money
balances, the level of real output, and current and
past rates of inflation. However, the long-term rate
would depend on the level of real money balances
rather than the rate of change in nominal money
balances and on the level of real income rather than
the rate of change in real income.
Finally, the Durbin-Watson statistic is very low,
suggesting serious serial correlation ofthe residuals.
Reestimating the equation using the CochraneOrcutt technique to correct for first-order serial cor
relation yields quite different parameter estimates
for the money and output variables and an unimpres
sive equation in terms of the significance of key
parameter values.
The anticipated-price-change equation — The
equation for anticipated price change (APA) is an
■identity, given by
17
(S T L -4 ) APA, = Y m ( [ 1 c 4i (P,-i/((Jt.i/UFt.,) ) .01
i= l
+ 1 ] 25 - 1 ).

This seemingly complicated equation transforms the
weighted distributed lag on current and past inflation
into the first difference of the price variable used in
the St. Louis Model. This price change variable is
not the first difference ofthe implicit price deflator;
it is, instead, the first difference in the implicit de
flator multiplied by the lagged value of the level of
real output. This particular form of the price change
variable is necessary because ofthe way that output
is determined in the model.
E c o n o m i c R ev iew (D ecem ber 1976;, pp. 809-20; and John A.
Tatom and Jam es E. Turley, “Inflation and Taxes: D isincentives
for Capital Form ation,” this R e v ie w (January 1978), pp. 2-8.

FEDERAL RESERVE BANK OF ST. LOUIS

To determine the dynamics associated with the
price-change equation and, in particular, whether the
St. Louis price-change equation implies a long-run
trade-off or a vertical Phillips Curve, one must solve
for the implied sum of the coefficients on lagged
price changes. The equation for the change in the
price level can be expressed directly as a function
of demand slack and a distributed lag on past price
changes:
5

A P *, = ho + 1 hi, DSL,., +

i =0

b2i' AP,.i.

i= l

The sum of the coefficients on past price changes
can in turn be related to the parameter on the antieipated-priee-change variable in the price-change
equation, STL-2, and the coefficients on past inflation
in the long-term interest rate equation.
V

l) 2i = h i ■ {

1 C4i / M ean (U /U F) }

The 2b£i term equals 1.13, based on the St. Louis
Model’s estimates for 1)2 and the C4i parameters. The
fact that the sum of coefficients on past price changes
exceeds unity results in a dynamic instability in
long-run simulations with the Model and a more
rapid response of inflation to monetary change than
if this sum were constrained to unity. This feature of
the price-change equation reinforces the influence
of the upward bias in the coefficient on the demand
slack variable.

The Unemployment Gap Equation
The unemployment rate (U) is determined by the
following equation:
(S T L - 5 )

ITC A P, = do G A P , + d i G A IV i,

where UGAP = LI — UF and GAP is the percentage
gap between potential output and actual output
(GAP = ((POTRT - X)/POTRT)* 100). The unemployment rate is then calculated from the identity,
(S T L -6 )

U, =

U F, +

U G A P,.

The parameter estimates for STL-5, based on the
sample period I/1955-IV/1980 are:
do = .0 2 4

(.3 8 )

ill = .4 5

(7.2)

R2 = .7 8

= .5 5

= .3 8

The pattern of coefficients on the gap variables
in this equation are different from what might have
been expected. The coefficient on the GAP variable
in the contemporaneous period is essentially zero,

DECEMBER 1981

implying that a change in the level of output relative
to potential output has no impact on the unemploy
ment rate in the same quarter. In addition, the
Durbin-Watson statistic is low, suggesting the
possible omission of other important explanatory
variables.

The Short-Term Interest Rate Equation
We ignore the remaining equation in the St. Louis
Model, the equation for the short-term interest rate
(4- to 6-month commercial paper rate). This variable
does not appear elsewhere in the model and we are
not interested in the model’s predictions for interest
rates.

The Output Identity
The change in output is determined in the St.
Louis Model via an “identity.” Using first differ
ences, AY can be expressed as
AY, = P,_i AX, + X,.i AP, + AX, AP,.

The price change variable in the St. Louis Model is
thus not AP, but rather X-iAP, the dollar change in
total spending due to price changes (ignoring the
interaction term, AX AP). The “change-in-output”
variable in the St. Louis Model is then determined
by an approximation to the actual identity since the
interaction term is excluded. Thus the change in
output in the St. Louis Model, P-i AX, is defined by
(STL-7) P,.| AX, = AY, - X,.i AP,.

RE FR A IN S O N A ST. L O U IS M O D E L
T H E M E : P H IL L IP S C U R V E A N D
M O N E T A R IST R E D U C E D - F O R M
A PPRO A CH ES TO
T H E IN F L A T IO N RATE
In this section, we present two variants of the
St. Louis Model. The two versions differ from each
other only in the equation used to explain the infla
tion rate. The first version includes a fairly conven
tional Phillips Curve and the second utilizes a
monetarist reduced form instead. Thus, the inflation
sector of the St. Louis Model is collapsed to a single
equation in each of these two alternative models.
Each of the revised versions includes the AndersenJordan equation and an unemployment equation. To
avoid the appearance that either of these variants
are intended to or actually have superceded the St.
Louis Model previously presented, the two versions
are labeled UCITYPC and UCITYRF, designating
17

FEDERAL RESERVE BANK OF ST. LOUIS

that they were developed in University City (alias
UCITY), a suburb immediately west of the city of
St. Louis and adjacent to Washington University.
This is intended to remind the reader that these
versions are close to the St. Louis Model, but not
identical o it. Of course, the PC and RF refer to the
different] ting feature of the two versions, the
Phillips C irve (PC) or the monetarist reduced-form
(RF) equation used to explain the rate of inflation.
First, we present the equations that the two
versions have in common: the reduced-form equa
tion for nominal income, the equation for the unem
ployment gap, and the identity that converts pre
dicted values for nominal income and price level
into predictions for output. Then, we will detail
the two alternative specifications of the inflation
equation.

The Identity Relating Nominal Income,
Output and the Price Level
The relation between nominal income (Y), output
(X) and the price level (P) can be expressed by the
identity,
Y =

FX.

We wish to avoid the use of an approximation to
solve for output, as the current version of the St.
Louis Model does. The model will yield solutions
for the rate of change in both nominal income and the
price level. However, the equation,
Y = X + P,

is only an approximation when the dot variables are
measured by compound annual rates of change, an
approximation that becomes poorer as the size of the
rates of change increases. To solve this problem,
the rates of change are defined as changes in the logs
of Y, X, and P (delta log specification). Taking logs
and then first differences of the equation, Y = P X,
yields the identity,
(U C IT Y -7 ) A in Y = A in X + A in F .

The Andersen-Jordan and inflation equations both
will be specified in terms of delta logs. The identity
above will then be used to determine the change in
the log of output.

The Andersen-Jordan Equation
There is, of course, little difference between the
dot and delta log specifications of an equation. The
delta log specification is given by,
Digitized for18
FRASER

DECEMBER 1981

4
( l !C I T Y - l )

A i n Y, = a<> +

a, ,

Ain

M ti

i =0
4
+

a 2 i Al l ! Gt-i.

i =0
The parameter estimates for the sample period
I/1955-IV/1980 are:12
an

ai<>

=
=
=
=
=

a i2
a 13
a i4
—a i j
R2

2 .69
(3.26)
(4.34)
.45
.44
(6.59)
(2.6)
.24
.032
(.50)
- .0 6 6 (-.5 9 )
(7.19)
1.10
.45

SE = 3.3 3

=
=
=
=
=
=

a 2 <>
1*21
a22
<*23

a24
—a 2 i

.062 (1.59)
.054 (1.81)
.004
(.116)
—,0 52( - 2 . 0 2 )
- . 0 6 9 ( - 2 .0 1 )
~ .0 0 1 (-- 0 .0 1 )
2.04

The Unemployment-Gap Equation
The specification of the unemployment-gap
equation is unchanged from the St. Louis Model
(STL-5). The only modification is that it is estimated
with a correction for second order autocorrelation.
The estimates for this equation are:
do
di

=
=

in
[i2

=
=

.26
.17
1.17
-.3 8
.74

(14.3)
(9.5)

.19

DW =

2.04

where pi and p2 are the values of the rho coefficients
on the first and second lagged values of the residual.
Note the dramatic decline in the standard error of
this equation, compared with the one in the St. Louis
Model.13
The two revised versions include equation
(STL-5), as reestimated above, and the identity
(STL-6).

The Level o f Output
Because the level of output is used in the GAP
variable in the Phillips Curve, we must also include

12The change in log variables are all m ultiplied by 400 prior to
estimation so that they approximate annual rates of change.
13The param eter estim ates o f the revised equation are quite close
to those presented in John A. Tatom, “ Econom ic Growth and
Unemployment: A Reappraisal o f the Conventional View ,” this
R ev iew (O ctober 1978), pp. 16-22. Tatom corrects the levels
equation for first-order serial correlation and also presents a
first difference equation, also with a correction for first-order
serial correlation.

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

an identity to determine the level of output fiom the
predicted values of the change in the log of output
and last period’s level of output.
(UCITY-8) X, = exp (In X,., + Ain Xt)

The Phillips C urve
The Phillips Curve equation uses a delta log speci
fication for the rate of change in the price level,
measures the demand slack in the economy with the
GNP gap variable, and proxies expected inflation
with a distributed lag on past rates of inflation. The
latter distributed lag can also be interpreted as
capturing an element of inertia due, for example, to
the existence of implicit or explicit contracts.
The Phillips Curve also includes the differential
in the rate of increase in the producers’ price of
energy relative to the rate of increase in the implicit
price deflator for GNP. This variable, labeled
ENERGY, is intended to capture a major source of
“supply shocks” that dramatically have affected
the inflation rate over a couple of periods in the data
sample, in particular, during the latter part of 1973,
throughout 1974 and, more recently, in 1979 and
early 1980. This variable is lagged two periods,
reflecting some experimentation with other simple
lag patterns.
The Phillips Curve also includes a dummy vari
able to capture the influence of the price controls
during the period from 111/1971 through 1975. The
variable, labeled CONTROLS, allows for a negative
impact during the first part of the period and an off
setting positive influence associated with “catch-up”
effects during the period after which controls were
relaxed and then removed. The sum of the values
the dummy variable takes on over this period is con
strained so that the net price control effect on infla
tion is zero. Specifically, CONTROLS is 0 up to II/
1971, 1 from III/1971-IV/1972, .2222 in 1/1973,
—.7778 from II/1973-I/1975 and 0 thereafter.14
The estimated Phillips Curve equation is
( l ’ClTY-2)

Ain

P, - #>

GAPn

+ /S2 CONTROLS,
+

p3 ENERGY,.,
20

/Si, Ain P,.,

i=l

14This specification of the controls variable was borrowed
from John A. Tatom at the Federal Reserve Bank o f St. Louis.

The distributed lag on inflation is estimated using
a third degree polynomial with no end-point restric
tions. We have employed the lagged GAP in this
equation, preserving the simple recursive structure
of the St. Louis Model. The empirical estimates with
the contemporaneous GAP were almost indistin
guishable from this equation.
The parameter estimates for the period 1/1955IV/1980 are:
.85
-.2 2
-1 .0 8 5
.044

0»
0i
02

.19
.14
.097
.066
.043
.028
.020
.017
.019
.023

011
042
043
044
045

/Sifi
047
04S
049

Hi w
R2

.827

(3.14)
(-4 .3 )
(-2 .6 5 )
(3.82)
.030
(4.32)
/Sin
.037
(5.19)
0412
.044
(6.39)
0413
(5.72)
.045
0414
(3.24)
.051
0415
.050
(1.83)
0416
(1.2.3)
.043
0117
.030
(1.1)
04 IS
(1.32)
.010
/Si 19
-.0 1 9
(1.87)
0420
.96 9
(14.04)
i-04, =
SE

1.166

DW =

(2.63)
(3.24)
(3.54)
(3.66)
(3.77)
(3.9)
(3.8)
(2.43)
(.51)
(-.5 6 )

2.10

Note that the GAP variable is highly significant,
that the sum of coefficients on the past inflation rates
is not significantly different from unity, and that
both the controls dummy and the energy differential
variables are significant.

The Inflation Reduced-Form Equation
The inflation reduced-form equation explains the
inflation rate in terms of current and lagged values
of monetary growth and the energy inflation dif
ferential and controls dummy variables discussed
above.15 This equation is also specified in delta log
form:
' “M onetarist reduced-form equations for inflation have been em
ployed for some tim e at the Federal R eserve Bank o f St. Louis.
The equation was initially reported in 1976 in D enis S. Karnosky, “T he Link Betw een M oney and Prices: 1971-76,” this
R e v ie w (June 1976), pp. 17-23. Som e refinements have been
presented in Keith M. Carlson, “T h e Lag from M oney to
P rices,” this R ev iew (O ctober 1980), pp. 3-10, and in John A.
Tatom, “ Energy Prices and Short-Run Econom ic Perfor
m ance,” this R ev iew (January 1981), pp. 3-17. A sim ple annual
version o f a monetarist reduced-form inflation equation was
presented in Jerom e L. Stein, “Inflation, Em ploym ent and Stag
flation, " J o u r n a l o f M on etary E c o n o m i e s (April 1978), pp. 193228.

3. The St. Louis Phillips Curve uses an unusual
demand slack variable, the change in nominal in
come minus the lagged real GNP gap; the UCITY
Phillips Curve uses the GNP gap.

19
( UC IT Y -2') A i n P,

yi.

A l n Mf_.

i=0
+ y2 C O N T RO LS ,
+ y3 E N E R G Y ,.2 .

This distributed lag in monetary change is estimated
using a third degree polynomial with no end-point
restrictions. The parameter estimates for the sample
period I/1955-IV/1980 are:
yio
yu
yi2
yi3
yi-t
yis
yi6
yi?
yis
yi9

.039
.047
.054
.058
.061
.063
.064
.063
.062
.060

y2
y.3
R2

yiio
ym
yii2
yii3
ym
yii5
yii6
yin
yn«
yiw

(1.34)
(2.45)
(4.02)
(5.19)
(5.38)
(5.27)
(5.25)
(5.36)
(5.50)
(5.52)

Vyii

= l.,04

.057
.055
.052
.050
.046
.044
.042
.041
.041
.042

(5.28)
(4.81)
(4.30)
(3.89)
(3.66)
(3.59)
(3.12)
(2.22)
(1.11)
(.35)

.822

1.173

1.62

The parameter estimates on the controls dummy
and the energy inflation variable are both significant,
and the coefficients on the monetary change variable
sum to unity. The two inflation equations, LJCITY-2
and UCITY-2', perform quite similarly with respect
to in-sample error, with a very slight edge to the
Phillips Curve.

S ummary o f D ifferences o f U C I T Y Models
and the St. Louis Model
A summary of the St. Louis and UCITY models
is given in table 1. The differences between the St.
Louis Model and the UCITY models can be sum
marized as follows:
1. The nominal income and inflation equations are
both specified symmetrically in delta log form in the
UCITY models, allowing the change in the log of
output to be solved for via an identity. In the St.
Louis Model, the nominal income equation is in a
rate-of-change specification, the price equation in
first difference, and the change in output is solved
for via an approximation.
2. The St. Louis Model employs a three-equation
inflation structure. The UCITY models employ alter
native single equations for inflation.
Digitized for20
FRASER

5. One of the UCITY models substitutes a mone
tarist reduced-form equation for inflation for the
Phillips Curve. The St. Louis inflation sector is built
around a price-ehange version of a Phillips Curve.
6. The unemployment equation is estimated
using a correction for second-order autocorrelation
in the UCITY models. It is estimated using ordinary
least squares in the St. Louis Model.

(33.5)

- 1 .9 4 ( -4 .7 2 )
.045 (4.12)

4. The weights on past inflation in the St. Louis
Phillips Curve are derived from an equation for the
long-term interest rate. In the UCITY model, the
weights are estimated directly during estimation of
the Phillips Curve.

C O M P A R IN G T H E T H R E E M O D E L S :
P R E D IC T IV E P E R F O R M A N C E A N D
P O L IC Y IM P L IC A T IO N S
This section compares the predictive performance
and the policy implications of the three models. The
results reported here bear directly on the three
themes outlined at the beginning of the paper. First,
in-sample and out-of-sample static simulations are
used to compare the predictive performances of the
St. Louis Model and the two UCITY models. Second,
the responses of output, unemployment and inflation
in the models to a deceleration in monetary growth
are compared. Third, the two UCITY models are
compared to determine whether any differences
exist in their predictive performance or policy
multipliers.

Predictive P erform ance o f the
Three Models
Because the two UCITY models include two
significant variables not included in the St. Louis
Model — the controls dummy and the energy infla
tion differential — it would not be surprising if they
perform better than the St. Louis model. In order to
determine the degree to which differences in predic
tive performance were due to the addition of these
variables, two additional versions of each UCITY
model were estimated: one without the controls
dummy, the other without the controls dummy and
the energy-inflation differential.

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS

Table 1

Summary of St. Louis and UCITY Models
St. Louis Model

UCITY Models
P hillips Curve

(1) Yt = ao +

4
ai, Mt-j
i=0

2 a2i Gt-i
i=0

A in Yt =

«o +

a ii A in Mt-i

Same as UCITYPC model

i =0
+

(2) A P *, = b0 +

M onetarist Reduced Form

4
2
i= 0

« 2j A in Gt-i

5
£ bii DSLh + b2 APA,
i=0

(2a) AP*, = X,.t AP,
(2b) DSLt = AY, -

A in Pt = /3o + /3i GAPm
(POTRTt -

Xm )

(3) RLt = Co + c i Mt + C2 Zt
16
16
+ 2 c3i Xm + S c4j (Pt-i/(U,-i/UF,-,))
i=0
i=0

A in Pt =

19
£ y ii A in Mt-i
i=0

+ /82 CONTROLSt

+ 72 CONTROLSt

+ (3s ENERGY,-2

+ y3 ENERGYt-2

20
2
i= 1

4i A in Pt-i

17
(4) A P At -

Y,-i( [ 2 c4i( Pt-i/{U,-,/UFt.,)).01 + 1]25 _ -,)
i =1

(5) UGAPt = do GAPt + d i GAP m

Same as St. Louis Model but
corrected for 2nd-order auto
correlation

Same as UCITYPC model

(6) U, = UF, + UGAP,

Same as St. Louis Model

A in Y = A in X + A in P

Same as UCITYPC model

Xt = exp(ln Xm + A in Xt)

Same as UCITY PC model

(7) Pm A X t = A Y t (8) X, -

Xm +

XM AP,

p t-i A x t
Pt-1

In-sam ple static sim ulation results — The insample root-mean-square errors (RMSEs) for inflation
(P), rate of change in nominal GNP (Y), rate of change
in real GNP (X), level of real GNP (X), GNP Gap
(GAP), and unemployment rate (U) for the various
versions of the UCITY models and for the St. Louis
Model are presented in table 2. Table 3 reports the
percentage declines in RMSEs in the two UCITY
models compared with the St. Louis Model. The two
UCITY models uniformly predict more accurately
than the St. Louis Model (the sole exception being
the rate of change in nominal GNP for which the
equations and hence predictions are virtually
identical).

large and, surprisingly, is accounted for to only a
minor degree by the addition of the two new vari
ables, although each does marginally improve the
inflation predictions. The inflation RMSEs for the
St. Louis Model, UCITYPC, and UCITYRF were
2.11, 1.12 and 1.14, respectively. This translates into
a reduction in the RMSE for inflation of 47 percent and
46 percent in the UCITYPC and UCITYRF models,
relative to the St. Louis Model. When hath the con
trols dummy and energy inflation differential vari
ables were excluded, the RMSEs in the UCITYPC
model increased to 1.29 and UCITYRF to 1.42, still
dramatically below the RMSE in the St. Louis
Model.

The improvement in the inflation forecast is quite

These results indicate that: (1) the inflation predie-

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

Table 2

Full In-Sample Static Stimulations, I/1955-IV/1980 (root-mean-square errors)

St. Louis

UCITYPC

UCITYPC
w/o ENERGY

UCITYPC
w/o CONTROLS
w/o ENERGY

UCITYRF

UCITYRF
w/o ENERGY

UCITVRF
w/o CONTROLS
w/o ENERGY

2.11

1.12

1.20

1.29

1.14

1.23

1.42

3.23

3.22

3.24

2.98

3.06

3.08

3.07

3.13

3.14

8.80

8.09

8.29

8.62

8.32

8.54

8.58

GAP

.79

.72

.74

.75

.76

.55

.26

.27

.26

.27

tion in the St. Louis Model can be improved by
substituting either a more traditional Phillips Curve
or a monetarist reduced-form for the St. Louis
Model’s price-change equation; and (2) inflation
predictions with the two versions of the UCITY
model are very close, not surprising given the small
differences in the standard errors in the two inflation
equations.
The UCITY models also outperformed the St.
Louis Model for the rate of change in output, the
level of output, the GNP gap, and the unemployment
rate, although the degree of improvement is smaller
for the two output variables and GAP than for the
inflation rate and the unemployment rate. For the
rate of change in output, the RMSE in the St. Louis
Model was 3.24, compared with 2.98 in the UCITYPC
model and 3.07 in the UCITYRF model. This repre
sents an improvement in the RMSEs of 5.5 percent
to 8 percent in the two UCITY models, a much small
er improvement than might have been expected
given the margin of improvement for inflation.
As in the case ofthe inflation predictions, eliminating
the controls dummy and inflation differential vari
ables results in a small deterioration in the quality
ofthe predictions from the UCITY models, but still
leaves those predictions superior to those from the
St. Louis Model. Interestingly, the improvement in
the predictions for the unemployment rate was about
as great as for the inflation rate, surprising in compari
son with the much smaller improvement in predic
tions ofthe GAP, but less surprising in light ofthe
particularly poor statistical quality o fth e St. Louis
unemployment equation.
O ut-of-sam ple static foreca sts — The three
models were re-estimated over the shorter period,
1/1955-1V/1976, and static forecasts were made for
Digitized for22
FRASER

Table 3

Percent Reduction in In-Sample UCITY
RMSEs (relative to St. Louis Model)
UCITYPC

UCITYRF
46.0%

46.9%

8.0

5.2

8.1

5.5

GAP
U

8.9

5.1

49.1

the period I/1977-IV71980. The results of the outof-sample static forecasts were consistent with the
in-sample results. The two UCITY models again
outperformed the St. Louis Model for all variables
(except nominal income, of course). The improve
ment for inflation was somewhat smaller (33 percent
and 17 percent for UCITYPC and UCITYRF, respec
tively) while the improvement for the output and
GAP variables was somewhat larger (9 percent to
15 percent in the UCITY models) than in the case of
the in-sample results. Once again, the unemploy
ment rate predictions for the St. Louis Model were
poor compared with the UCITY results. The out-ofsample RMSEs for the various variables are reported
in table 4, and the percent improvement in RMSEs
in the UCITY models is given in table 5.

The Response o f Output, Unemployment
and Inflation to Monetary C hange in
the T hree Models
The UCITY models were developed, in part, to
improve the predictions of inflation, output and

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

Table 4

Out-of-Sample Static Simulations: 1/1977-IV/1980 (root mean square errors)

St. Louis

UCITYPC

UCITYPC
w/o CONTROLS

UCITYPC
w/o CONTROLS
w/o ENERGY

UCITYRF

UCITYRF
w/o CONTROLS

UCITYRF
w/o CONTROLS
w/o ENERGY

2.12

1.43

1.51

1.67

1.77

1.93

1.81

3.78

3.80

3.92

3.34

3.58

3.47

3.58

3.84

3.55

14.36

12.12

13.00

12.59

13.04

14.04

12.92

GAP

.95

.81

.87

.84

.87

.93

.86

.59

.28

.29

.30

.32

.30

Table 5

Percent Reduction in Out-of-Sample
UCITY RMSEs (relative to
St. Louis Model)
UCITYPC

UCITYRF
16.5%

32.5%

14.8

8.7

15.6

9.2

GAP

14.7

8.4

52.5

49.2

unemployment from those available using the St.
Louis Model presented here. Also of interest are the
differences in the policy simulations obtained using
the three models.
For the policy simulations, CEA projections for
potential output and for high employment govern
ment expenditures were used for the period from
I/1981-IV/1984. Two alternative monetary growth
rates were used: as the “base” series, we used a
constant rate of 5 percent per year, for the “policy”
series we used 2 percent per year. We then com
puted the differences in the rates of change of
nominal and real income, and differences in the
level of real GNP, in the GNP gap, in the unemploy
ment rate, and in inflation between the base and
policy simulations. The results are reported in tables
6, 7 and 8 for the inflation rate, the rate of change in
real output and the unemployment rate. The figures
reported in each case are the values in the policy
run minus the values in the base run.

The results confirmed our expectations about the
direction of the differences, but the magnitude of
the differences between the St. Louis and UCITY
simulations were somewhat smaller than expected.
Inflation declines more rapidly in the St. Louis
Model and, as a consequence, the decline in the rate
of growth of output and the increase in unemploy
ment are smaller in the St. Louis Model.
For the inflation rate, all three models’ projections
are close during the first year, with inflation falling
about 0.4 percentage points. By the end of the
second year, inflation has fallen 1.8 percentage points
in the St. Louis Model, compared with only 1.2 in
the two UCITY models. By the end of the fourth
year, inflation has fallen by 4.0 percentage points in
the St. Louis Model compared with 2.8 and 2.9 per
centage points in the UCITY models. Thus, while
inflation has fallen more rapidly in the St. Louis
Model, the decline by the end of the fourth year
exceeds the equilibrium response, implying a ten
dency to overshoot.
In the St. Louis Model, the rate of increase in out
put declines for the first 12 quarters, the decline
exceeding 2 percent per year for the first 6 quarters.
In the two UCITY models, the rate of increase in out
put is lower throughout the 16-quarter simulation
horizon, by 2 percent per year or more for eight
quarters.
The unemployment results indicate that the
monetary deceleration raises the unemployment
rate for 16 consecutive quarters for each model, but
that unemployment is about 0.6 percentage points
higher in the two UCITY models at the end of the
simulation horizon compared with the St. Louis
Model.
23

Table 6

Table 7

Dynamic Simulation Results: Inflation
Rate (difference between the 2 percent
and 5 percent simulations)

Dynamic Simulation Results: Rate of
Change in Real GNP (difference
between the 2 percent and 5 percent
simulations)

Date

St. Louis

UCITYPC

UCITYRF

1981 I

-.0 1

-.0 4

Date

-.0 7

-.0 6

-.1 2

St. Louis

UCITYPC

III

-.2 0

- .2 4

-.4 3

-.3 8

1982 I

-.7 5

UCITYRF

1981 I

-1 .1 7

-1 .5 7

-1 .3 3

-2.51

-2 .5 5

-2 .5 7

-.4 0

III

-3 .2 2

-3 .0 8

-3 .1 5

-.5 9

-.5 7

-3 .0 2

-3 .0 3

-3 .0 8

-1.11

-.8 0

- .7 7

1982 I

-2 .4 4

-2 .6 3

-2 .7 2

III

-1 .4 8

-1 .0 2

-.9 8

-2 .0 5

-2 .4 3

-2 .5 3

-1 .8 4

-1 .2 4

-1 .2 0

III

-1 .6 6

-2.21

-2 .3 2

-1 .2 9

-1 .9 8

-2 .1 0
-1 .8 8

1983 I

-2 .1 7

-1 .4 7

-1 .4 2

-2 .4 9

-1 .6 9

-1 .6 4

1983 I

-.9 7

-1 .7 5

III

-2 .7 9

-1 .9 0

-1 .8 6

-.6 5

-1 .5 3

-1 .6 5

-3 .0 6

—2.11

-2 .0 8

III

-.3 4

-1.31

-1 .4 4

1984 I

-3 .3 2

-2 .3 2

-2 .2 8

-.0 5

-1 .1 2

-1 .2 2

-3 .5 5

-2 .5 2

-2 .4 8

1984 I

.02

-.9 0

-1 .0 2

III

-3 .6 7

-2 .7 2

-2 .6 6

.41

-.7 0

-.8 2

-3 .9 5

-2 .9 2

-2 .8 2

III

.63

-.5 1

-.6 4

.83

-.3 1

-.4 8

C om paring the Predictive Perform ance
and Policy Implications o f the Phillips
C urve and Monetarist Reduced-Form
Inflation Equations
There is a considerable literature that views the
Phillips Curve and the monetarist reduced form for
inflation as mutually exclusive, alternative inflation
equations.16 Generally, these “competitive” alterna
tive approaches are tested by investigating the con
sequences of adding monetary change to a Phillips
Curve or introducing the unemployment rate into a
monetarist reduced form.17

l6See, for exam ple, Keith M. Carlson, “ Inflation, Unemployment,
and M oney: C om paring the E v id e n ce from Tw o Sim p le
M odels,” this R ev iew (Septem ber 1978), pp. 2-6; and John A.
Tatom, “ Does the Stage of the Business Cycle Affect the Infla
tion R ate?” this R eview (Septem ber 1978), pp. 7-15; and Stein,
“ Inflation, Em ploym ent, and Stagflation.”
17Tests o f this kind have been reported by Franco M odigliani and
Lucas Papademos, “Targets for M onetary Policy in the Coming
Years,” B r o o k in g s P apers on E c o n o m i c Activity (1:1975), pp.
141-63; George L. Perry, “ Slowing the W age-Price Spiral: The
M acroeconom ics V iew ,” B r o o k in g s P a p e rs on E c o n o m i c A ctiv
ity (2:1978), pp. 259-91; Stein, “ Inflation, Em ploym ent and Stag-

Digitized for24
FRASER

We do not view the Phillips Curve and monetarist
reduced form as mutually exclusive, alternative
models of the inflation process but rather as structural
vs. reduced form approaches to explaining inflation.
Because we view the two inflation equations as
reasonable alternative specifications, we find no
value in tests that add monetary change variables to
the Phillips Curve or unemployment rates to the
monetarist reduced form. Such experiments mix
structural and reduced-form equations. We would
not expect to be able to obtain significant coefficients
on both monetary change and unemployment rates
in an inflation equation, and, consequently, no such
experiments were conducted. Instead, we compared
the two inflation equations individually and as alter
native components of a St. Louis-type model; we
found that the two inflation equations were virtually
indistinguishable in predictive performance and
policy implications.
First, when the single equation performance of
the Phillips Curve and monetarist reduced form
flation;” and John A. Tatom, “What E v er Happened to the
Phillips Curve?” Federal Reserve Bank o f St. Louis R es ea rch
Papers, No. 81-008.

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS

Table 8

Dynamic Simulation Results:
Unemployment Rate (difference
between the 2 percent and 5 percent
simulations)
Date

St. Louis

1981 I

UCITYPC

UCITYRF

0.15

0.08

.15

0.40

0.30

III

.41

0.67

0.59

.78

0.94

0.90

1982 I

.91

1.19

1.17

1.34

1.42

1.41

III

1.54

1.63

1.70

1.81

1.82

1983 I

1.83

1.97

1.99

1.93

2.11

2.13

III

1.99

2.23

2.26

2.03

2.33

2.36
2.44

1984 I

2.04

2.41

2.02

2.48

2.50

III

2.00

2.53

2.56

1.94

2.56

2.60

were compared, the Phillips Curve and monetarist
reduced forms yielded standard errors of 1.166 and
1.173, respectively. Thus the two equations fit the
data almost equally well. Note the high level of
significance of the key variables in both equations —
the gap and the sum of the coefficients of past infla
tion in the Phillips Curve and on the sum of the
coefficients of monetary change in the monetarist
reduced form.
Second, the in-sample and out-of-sample static
forecasts of the two UCITY models were compared,
the only difference beingthatone includes a Phillips

Curve while the other includes a monetarist reduced
form. Looking at tables 2 and 4, we observe that the
performance of the two models is very close, with a
small but consistent edge to the Phillips Curve for
virtually all variables in both in- and out-of-sample
results.
Finally, policy simulations with the two UCITY
models yielded remarkably similar results. Looking
at tables 6, 7 and 8, we observe that the policy multi
pliers are nearly equal in both cases for inflation,
output and unemployment. Rounded off to the first
decimal point, they are almost identical, particularly
after the first four quarters.

C O N C L U S IO N S
In this paper, we reviewed a current version of the
St. Louis Model and presented two alternative ver
sions, referred to as UCITY models, that retain the
Andersen-Jordan nominal income reduced form but
simplify the inflation sector and improve the estima
tion of the unemployment rate. In the UCITYPC
version, we replaced the St. Louis Model’s inflation
sector with a more conventional Phillips Curve. In
the UCITYRF version, we substituted a monetarist
reduced form for inflation for the Phillips Curve.
We demonstrated that the UCITY versions yield
improved predictive performance of the major
economic variables of interest to policymakers when
compared with the St. Louis Model. In addition, the
St. Louis Model yields more rapid deceleration of
inflation and a smaller temporary rise in unemploy
ment in response to a deceleration monetary growth
than in the UCITY models. Finally, the UCITY
models yield very similar predictive performances
and virtually identical policy multipliers, suggesting
that the Phi Hips Curve and monetarist reduced form
are both reasonable, alternative equations for ex
plaining inflation and correspond to structural vs.
reduced-form approaches to modeling inflation.

FEDERAL RESERVE BANK OF ST. LOUIS

DECEMBER 1981

FEDERAL RESERVE BANK OF ST. LOUIS
REVIEW INDEX 1981
JANUARY

JUNE/JULY

John A. Tatom, “ Energy Prices and Short-Run
Economic Perform ance”

Neil A. Stevens, “ Indexation of Social Security
Benefits — A Reform in Need of Reform ”

W. \N. Brown and G. J. Santoni, “ Unreal Estimates
of the Real Rate of Interest

Scott E. Hein and James C. Lamb, Jr., “ Why the
Median-Priced Home Costs So M uch”

Neil A. Stevens, “ O utlook for Food and A griculture
in 1981”

Dallas S. Batten, “ Inflation: The Cost-Push M yth”

FEBRUARY

AUGUST/SEPTEMBER

Michael E. Trebing, “ The New Bank-Thrift Competi
tion: Will It Affect Bank Acquisition and Merger
Analysis?”

R. Alton G ilbert and Michael E. Trebing, “ The FOMC
in 1980: A Year of Reserve Targeting”

R. \N. Hafer, “ Selecting a Monetary Indicator: A Test
of the New Monetary Aggregates”

Clifton B. Luttrell, “ Grain Export Agreem ents— No
Gains, No Losses”

MARCH
Scott E. Hein, “ Deficits and Inflation”
G. J. Santoni and Courtenay C. Stone, “ Navigating
Through The Interest Rate Morass: Some Basic
Principles”
R. IV. Hafer, “ The Impact of Energy Prices and
Money Growth on Five Industrial Countries”

OCTOBER
Lawrence K. Roos, “ Lessons We Can Learn”
Dallas S. Batten, “ Money Growth Stability and Infla
tion: An International Com parison”
R. W. Hafer, “ Much Ado About M2”

Keith M. Carlson, “ Recent Revisions of GNP”
NOVEMBER
APRIL
Anatol B. Balbach, “ How C ontrollable is Money
G row th?”
Clifton B. Luttrell, “ A Bushel of Wheat for a Barrel
of Oil: Can We Offset OPEC’s Gains With a Grain
Cartel?”
Dallas S. Batten, “ Foreign Exchange Markets: The
Dollar in 1980”
MAY
Michael David Bordo, “ The Classical Gold Stan
dard: Some Lessons fo r Today”
John A. Tatom, “ We Are All Supply-Siders Now !”
Keith M. Carlson, “ Trends in Federal Revenues:
1955-86”

G. J. Santoni and Courtenay C. Stone, “ What
Really Happened to Interest Rates?: A Longer-Run
Analysis”
Keith M. Carlson, “ Trends in Federal Spending:
1955-86”
Clifton B. Luttrell, “ The Voluntary Autom obile
Import Agreement with Japan — More Protectionism”

DECEMBER
R. Alton Gilbert, “ Will the Removal of Regulation Q
Raise Mortgage Interest Rates?”
Laurence H. Meyer and Chris Varvares, “ A Com pari
son of the St. Louis Model and Two Variations:
Predictive Performance and Policy Im plications”