Full text of Review (Federal Reserve Bank of St. Louis) : January 1981

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

CO

o
GO
O
January 1981
Vol. 63, No. 1

03
QQ
CD
>
a3
CO
CD
Qd

3 Energy Prices and Short-Run
Econom ic Performance

CD
TD
CD



IS Unreal Estimates of the Real Rate
of Interest
27 O utlook for Food and A griculture in 1981

The R e v i e w is 'published. 10 times per year by the Research Department of the Federal Reserve
Bank of St. Louis. Single-copy subscriptions are available to the public free of charge. Mail requests
for subscriptions, back issues, or address changes to: Research Department, Federal Reserve Bank
of St. Louis, P.O. Box 442, St. Louis, Missouri 63166.
Articles herein may be reprinted provided the source is credited. Please provide the Bank’s Re­
search Department with a copy of reprinted material.


2


Energy Prices and Short-Run
Economic Performance
JOHN A. TATOM

T

.
HE sharp
energy price increases that have oc­
curred since late 1978 have profoundly affected the
U.S. economy. In particular, the increase in the price
of energy resources relative to the price of business
output has reduced potential output and productivity,
raised the general level of prices, and lowered the
optimal capital intensity of U.S. production which,
in turn, will temporarily slow real business invest­
ment in the early 1980s. Higher energy prices have
also had temporary effects on total spending and
employment.

The purpose of this article is to explain and assess
the magnitude of these energy price effects. Empirical
tests are conducted using a reduced-form model for
nominal GNP, the price level and the unemployment
rate. Real GNP growth is determined implicitly in
such a model as the difference between nominal GNP
growth and the rate of price increase. This model
emphasizes the link between money stock growth and
economic activity. The sample period for estimating
the relationships ends in the third quarter of 1978 to
provide an opportunity to test the stability of the re­
lationships over the past two years, when energy prices
increased sharply. Also, major changes in economic
policy have occurred since 1978 that may have affected
fundamental relationships that explain spending, in­
flation, output and unemployment. The empirical re­
sults, including simulations from the fourth quarter of
1978 to the third quarter of 1980, strongly support the
hypotheses developed below concerning energy price
effects. An assessment of the size of the effects of
recent energy price increases is obtained from the
empirical estimates.
The estimates indirectly imply that, once energy
price effects are taken into account, no significant
shift in the relationship between the money stock and



major measures of economic performance has oc­
curred over the last two years. Neither the shift in
focus toward greater emphasis on controlling mone­
tary aggregate growth announced in November 1978,
nor a shift in policy procedures in October 1979,
appear to have exerted independent impacts on the
linkages between money and the principal measures
of economic performance.

THEORETICAL CONSIDERATIONS
A simple aggregate supply and demand model will
clarify the analysis.1 In figure 1, the economy initially
is in equilibrium with price level, P0, and real GNP
level, X0, at point A. The aggregate demand curve,
AD, is constructed given levels of such other relevant
determinants of demand as current and past monetary
and fiscal actions. The aggregate supply curve, SS, is
constructed given such other determinants of supply
as expected nominal wages, the size of the labor
force, the existing capital stock, the relative price of
energy, and technology. The price of energy (instead
of a quantity of energy) enters the model indicating
that the economy in figure 1 is “open;” energy re­
sources can be imported or exported at prices set in
a world market.2 The aggregate supply curve is con­
structed with increasing slope to show that at some
real output level, it becomes difficult to increase real
1For a more detailed discussion o f the theoretical foundation
used here, as well as a discussion of alternative macroeco­
nomic approaches and empirical evidence from several nations
supporting the theory, see Robert H. Rasche and John A.
Tatom, “ Energy Price Shocks, Aggregate Supply and Mone­
tary Policy: The Theory and International Evidence,” forth­
coming in the Carnegie-Rochester Conference Series on Pub­
lic Policy, Volume 14, 1981.
2It is important to note that the effects of a higher relative
price of energy due to exogenous energy market developments
do not depend upon the net trade status o f the economy.

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

output despite increases in the general level of prices.
At this output level, the economy achieves full em­
ployment, utilizing available capital and labor re­
sources. Suppose that such full-employment condi­
tions occur at the initial equilibrium, point A.
When the relative price of energy resources in­
creases, the aggregate supply curve shifts to S'S'. The
employment of existing labor and capital with a given
nominal wage rate requires a higher general price
for output, if sufficient amounts of the higher-cost
energy resources are to be used. Of particular interest,
however, is the level of output and price level associ­
ated with full employment of existing labor and capi­
tal. This point is indicated in figure 1 at point B.
Given the same supply of labor services and existing
plant and equipment, the output associated with full
employment declines as producers reduce their use of
relatively more expensive energy resources and as
plant and equipment become economically obsolete.
The productivity of existing capital and labor re­
sources is reduced so that potential real output de­
clines to Xi. In addition, the same rate of labor em­
ployment occurs only if real wages decline sufficiently
to match the decline in productivity. This, in turn,
happens only if the general level of prices rises suffi­
ciently (P i), given the nominal wage rate.3
The new equilibrium for the economy occurs at
point B. For aggregate demand to equal X 1 at price
level Pj, the aggregate demand curve must be unitelastic with respect to the price level. In the context
of the equation of exchange, M V =Y (where M is the
money stock, V is its velocity and Y is nominal GNP),
this means that velocity is unaffected by a rise in the
price level, a standard long-run proposition in mone­
tary theory.4
The economy may not adjust instantaneously to
point B, even if point B is the new equilibrium. For
example, price rigidities due to costly information or
other transactions costs can keep nominal prices from
adjusting quickly. The immediate incentive to cut
production and employment indicated by the leftward
shift in the aggregate supply curve need not be ac:iThe percentage rise in the price level (percentage decline in
the real w age) will equal the decline in productivity, given
employment, if the marginal productivity of labor is propor­
tional to its average productivity. This proportion? lity holds
for a Cobb-Douglas production function. The general case is
derived by Rasche and Tatom, “ Energy Price-Shocks,” Ap­
pendix 1.
4The results when some o f the assumptions used here are
relaxed, especially the short-run invariance of nominal spend­
ing to changes in the price level, are discussed by Rasche and
Tatom, “ Energy Price Shocks.”


4


JA N U A R Y

F ig u re

1981

1

The E f f e c t o f a H i g h e r R e l a t i v e Pr i c e of
E n e r g y on O u t p u t a n d t h e Pr i c e Lev el

companied immediately by the price level adjustment
sufficient to ensure the maintenance of full employ­
ment. In this event, disequilibrium GNP will be dom­
inated by the reduction in output before the equi­
librium B (and full employment) is achieved. Conse­
quently, output and prices can move along an adjust­
ment path such as that indicated by the arrow in
figure 1. The evidence below is consistent with this
adjustment process and the hypothesis that GNP is
independent of energy price changes, once the ad­
justment is completed.

EVIDENCE ON POTENTIAL OUTPUT
AND PRODUCTIVITY
The theory and existing evidence on which this
article draws deals with isolating the permanent im­
pact of a higher relative price of energy on potential
output, productivity, the desired capital-labor ratio
and the price level. Before analyzing the dynamics of
the short-run adjustment process, it is useful to re­
view the evidence from a production function ap­
proach. Assume that output in the private business
sector (Q t) is a function of hours of employment (ht),
the utilized capital stock (kt), technological change
and the relative price of energy (p?)- The production
function can be written as:
( 1 ) In Q , =

(30 + Pi In h t + |32 In k, + (3a In p? +

|3,t

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JA N U A R Y

1981

C h a rt 1

Impact of Energy Price Changes (1 /1 9 7 0 —111/1980)
on Potential O utput G row th in the Private Business Sector^
Percent

1970

Percent

1971

1972

1973

1974

1975

1976

1977

1978

1979

1980

Source: Eq u a tio n 1
[X P e rc e n ta g e ch an g e s are m e a su re d in the lo g arithm of the le v el of p o te n tia l output.
Latest data plotted: 4th q u a rte r

where t is a time trend.5 When this equation is esti­
mated for the private business sector over the period
I/1955-III/1978, the result is:
(2 ) In Q, = 1.464 + 0.705 In h, + 0.295 In k t
(14.14) (18.10)
(7.59)
- 0.093 In pi + 0.004t.
(-5 .0 6 )
(13.04)
R- = 0.97

S.E. = 0.007

D.W . = 2.03

p = 0.81

This estimate is virtually identical to those reported
for earlier periods.6
5Rasche and Tatom, “ Energy Price Shocks,” and “ Energy Re­
sources and Potential GNP,” this Review (June 1977), pp.
10-24 derive equation 1 assuming that the production func­
tion is Cobb-Douglas and explain the interpretation of the (3
coefficients in terms of output elasticities of inputs. They also
describe tests for breaks in the time trend ana for the CobbDouglas restrictions.
BFor example, see Rasche and Tatom “ Energy Resources and
Potential GNP.” The sample period conforms to that used
for the equations estimated Delow.




Chart 1 shows the direct impact on the annual
growth rate of potential output from 1/1970 to III/
1980 using the energy price coefficient in equation
2. The relative price of energy measure is calculated
by deflating the producer price index for fuels and
related products and power by the price deflator for
private business sector output. Equation 2 indicates
that a 40 percent (Ain) change in the relative price
of energy, as occurred from III/1973 to III/1974 or
from IV/1978 to 11/1980, will permanently reduce
potential output and productivity in the private busi­
ness sector by 3.7 percent.7
7Although tests conducted to detect statistical biases in esti­
mates such as equation 2 have failed to find any, it is possible
that quarterly estimates are affected by lagged responses of in­
puts to output that would result in downward biased estimates
of the coefficient on the relative price of energy (in absolute
size). For example, the estimate of this coefficient using
annual data for the period 1949-75 is 11.3 percent, implying
a 4.5 percent reduction in potential output when energy
prices rise 40 percent.

FE D E R A L. R E S E R V E B A N K O F ST. LOUIS

A rise in the relative price of energy will also re­
duce the desired capital-labor ratio, temporarily re­
ducing business investment. The theoretical under­
pinnings and magnitude of the effect of the 1973-74
energy price increases on investment are discussed
elsewhere.8 Based on that methodology and the coef­
ficient estimates in equation 2, the capital-labor ratio
can be expected to decline by 5.3 percent due to
energy price increases that occurred from IV/1978
to III/1980.9 Productivity growth will tend to be
slower than it would have been during the years of
adjustment to this decline.
The production function estimates provide evidence
that the permanent aggregate supply effects of energy
price changes occur quickly. A broader model en­
compassing aggregate demand considerations is re­
quired, however, to assess actual quarter-to-quarter
adjustments in spending, output and prices.

THE EFFECT OF ENERGY PRICES
ON THE MONEY-GNP LINK
To examine the temporary adjustments of nominal
GNP to changes in the relative price of energy, a
variant of the Andersen-Jordan equation from the
St. Louis model is used.10 This reduced-form equa­
tion relates GNP to money stock and high-employment federal expenditure variables. It is usually ex­
pressed as:
(3) GNP = |30 + Pi £ w,° , M,-, + P. £ w,1-) E,-j
i-0

j= 0

8See John A. Tatom, “Energy Prices and Capital Formation:
1972-1977,” this Review (M ay 1979), pp. 2-11.
^Assuming that the price of, capital goods relative to business
output is unaffected by a rise in energy prices, the elasticity
o f the desired capital-labor ratio with respect to the relative
■y

price of energy is ( - — ), where y and a are the output elas­
ticities of energy and labor, respectively. Given the estimates
in equation 2, y = 8.5 percent and a = 64.5 percent. Thus,
the estimated capital-labor ratio elasticity is 13.2 percent. This
figure merely suggests the magnitude, however. When the sam­
ple period is lengthened or annual data is used, the estimate
is over 15 percent, not significantly different in a statistical
sense, but larger nonetheless. Moreover, it is likely that higher
energy prices raise the relative price of goods, further depress­
ing the capital-labor ratio.
10See Leonall C. Andersen and Jerry L. Jordan, “ Monetary
and Fiscal Actions: A Test o f their Relative Importance in
Economic Stabilization,” this Review (November 1968), pp.
11-24; Leonall C. Andersen and Keith M. Carlson, “A Mone­
tarist Model for Economic Stabilization,” this Review (April
1979), pp. 7-25; and Keith M. Carlson “ Does the St. Louis
Equation Now Believe in Fiscal Policy?” this Review (F eb ­
ruary 1978), pp. 13-19.


6


JA N U A R Y

1981

where GNP, M and E are annual growth rates
(400-Ain) of GNP, the money stock (M ) and highemployment federal expenditures (E ). The coeffi­
cients on current and lagged M and E variables are
estimated using Almon polynomials. The polynomial
degree, lag length and constraints for M and E co­
efficients are those used in the model — fourth de­
gree polynomials with five lags and head and tail
constraints.
Since major strikes temporarily reduce and subse­
quently increase GNP growth, a variable is included
to capture these temporary influences.11 This variable,
St, is the change in the quarterly average of “days
lost due to strikes,” deflated by the civilian labor
force.
Monetary aggregates have been revised to reflect
the existence of transactions balances not held either
as currency or demand deposits at commercial banks.
The new measure of the money stock that can be used
directly for transactions purposes is M1B, but data on
this measure exist only since 1959. The difference in
this measure and the old measure, Ml, is very small
in 1959. More important, the growth rates of both M l
and M1B are roughly the same until the early 1970s.
Consequently, the growth of the money stock M l is
used in the estimation of equation 3 for quarters in the
sample period prior to 1959. This practice is further
supported by the fact that the properties and coe­
fficients of the estimated equation are virtually iden­
tical to old estimates using M l for sample periods
prior to the rapid growth of transactions balances in
savings accounts with the automatic transfer service.
The GNP equation, estimated for the period 1/
1955 to III/1978 is:
(4 ) GNPt = 2.662 + 1.103 2 w?-i M ,-i
(3.35)
(7 .5 0 )“ °
+ 0.003 Z w,1-, E ,-j -0.471 St.
(0.04) i"°
(-3 .6 4 )
R2 = 0.46

S.E. = 3.18

D.W . = 1.88

The equation has the usual properties that the sum
of the coefficients on money stock growth is not sig­
nificantly different from one, and that the sum of
expenditure effects is not significantly different from
zero. The strike variable is significant and has the
right sign; the mean value of the strike variable is
n See Leonall C. Andersen, “ A Monetary Model of Nominal
Income Determination,” this Review (June 1975), pp. 9-19,
for an example of using strike dummies in such a GNP
equation.

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

0.017, so that the mean strike effect is only -0.008
percent.
To examine the impact of the relative price of
energy on GNP, current and lagged values of the an­
nual growth rate of the relative price of energy are
added to equation 4. A search was conducted for the
optimal lag length using F-tests for each additional
lagged value of the growth of the relative price of
energy and for additional groups of lagged values
(up to five at a time). The criterion for including
lags is the 5 percent significance level. Up to 16
lagged values were examined. The same examination
was conducted using polynomial distributed lags up
to the fourth degree, with and without end-point con­
straints. The results are virtually identical to those
reported below and the polynomial restriction is
unimportant. The polynomial distributed lag results
are discussed in the appendix.
The optimal lag length includes the current and six
past values of the growth in the relative price of
energy. The equation estimate with the unrestricted
distributed lag for energy prices is:

JA N U A R Y

1981

GNP. An effect on GNP continues, according to equa­
tion 5, for six more quarters, even though the change
in the relative price of energy is zero in subsequent
quarters. The pattern of coefficients on the energy
price terms indicates that a current-quarter rise in
the relative price of energy tends to reduce nominal
GNP for six quarters, then increases it.
In order to test the hypothesis that a change in the
relative price of energy has no lasting effect on nomi­
nal GNP, equation 5 is estimated with the sum of the
energy price coefficients constrained to zero. The
F-statistic for the addition of the freely estimated
coefficient in equation 5 is F,,so = 0.13, which is not
significant at the 1 percent level. The constraint that
the sum of the relative price of energy effects on GNP
is zero cannot be rejected. The constrained equation

(6 ) GNP, = 2.567 + 1,147 Z w,°-, M,_, + 0.004 X w?-j E, ,
(3.32) (7 .7 0 )“ °
(0 .0 5 )J=°
- 0.444 S, - 0.054 p 1 + 0.049 p?_,
(-3 .5 7 )
(-1 .4 9 )
(1.10)
- 0.031 p?_2 - 0.025 pi-3 -0.050 p f ,
(-0 .7 3 )
(-0 .5 8 ) , (-1 .1 6 )

(5 ) GNP, = 2.677 + 1.138 £ w,°-> M ,-, - 0.009 2 w,1., E,-,
(3.20) (7 .4 9 ),=0
(-0.11 ) J' “
- 0.443 S, - 0.050 p '
(-3 .5 4 )
(-1 .3 2 )

R- -

+ 0.050 pt_, - 0.029 p'_2 -0.022 p U - 0.048 p?_„
(1.12)
(-0 .6 6 )
(-0 .5 1 )
(-1 .0 9 )
+ 0.012 p ‘ -5 + 0.106 p'-„.
(0.28)
(2.83)
R* = 0.52

S.E. = 2.97

D.W . = 1.91

An F-test (5 percent significance level) of adding
the energy price terms to equation 4 rejects the hypo­
thesis that each of the energy price coefficients is
zero (F 7,8o = 2.63). The coefficients on the variables
in equation 4 are not changed significantly in esti­
mating equation 5.
The coefficients on the relative price of energy can
be used to determine the effect on nominal spending
of an increase in the growth rate of energy prices or
of a once-and-for-all rise in energy prices. The sum of
the coefficients on the rate of increase in energy prices
indicates the long-run effect on the growth of nominal
GNP of a 1 percentage-point increase in the annual
rate of energy price increases. This sum also indicates
the effect on the level of GNP of a once-and-for-all
rise in the relative price of energy. Consider an x
percent rise in the relative price of energy in the cur­
rent quarter. Such a rise affects GNP in the current
quarter and results in a difference in the logarithm of



+ 0.010 p,-5 + 0.101 pL,.
(0.22)
(2.87)
0.53

S.E. = 2.95

D.W . = 1.91

The F-statistic for the addition of the six independ­
ently estimated variables in equation 6 to equation 4
is F6i81 =: 3.08, which exceeds the critical F-statistic
at the 1 percent significance level, so that the hypo­
thesis that each of the relative price of energy coe­
fficients is zero is again rejected. None of the coeffi­
cients in equation 6 is significantly different from
those in equation 5. Equation 6 not only supports the
hypothesis that there is no permanent effect of the
relative price of energy on GNP, it also provides evi­
dence on the adjustment process with price rigidities.
Initially, nominal GNP is reduced by an increase in
the relative price of energy, as nominal GNP is domi­
nated by the real output effect discussed above. Only
later do price level effects reverse this nominal GNP
development. After six quarters, the transitory move­
ments in GNP have washed out. The theoretical prop­
osition that the shift in aggregate supply due to energy
price changes leaves nominal demand unchanged is
supported by the estimated equation.
Energy price changes have been substantial
the end of the sample period for equations
Moreover, the growth of the money stock has
erratic since the end of the third quarter of

since
4 - 6.
been
1978,
7

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JA N U A R Y

Table 1

Table 2

Simulation of Equation 6

Simulation of Equation 4

One-quarter
period ending

Actual GNP

Simulated
GNP

IV/1978

14.6%

14.6%

1/1979

11.9

10.9

Error1
0.0%
-1 .0

One-quarter
period ending:

Simulated GNP

Error1

IV/1978

12.6%

-1.9%

1/1979

10.7

-1.2

9.7

3.9

11/1979

5.8

8.3

2.5

111/1979

11.5

11.5

-0.1

111/1979

12.8

1.2

IV /1979

8.5

12.3

3.9

IV /1979

12.6

4.1

1/1980

11.9

9.0

-2 .9

1/1980

11.6

-0.3

11/1980

-1.1

3.4

4.5

11/1980

5.8

6.8

111/1980

11.2

5.6

-5.5

111/1980

7.7

-3 .4

Mean error =

fig u r e s may not add exactly due to rounding.

especially in 1980. Thus, the ability of equation 6 to
simulate the post-sample experience is a strong test.
Using actual data for money stock, federal expendi­
tures, and relative price of energy growth rates for the
period IV/1978-III/1980 results in the predicted
growth rates of nominal GNP shown in the second
column of table 1. Column 1 shows the actual GNP
growth rates. The third column shows the simulation
errors (simulated growth minus actual growth).
Equation 6 tracks extremely well in the eightquarter post-sample period. The errors in the last two
quarters, however, suggest that the credit control pro­
gram in the second quarter and its removal in the
third quarter had an impact. Over the eight quarters,
the mean error is 0.05 percent and the root-meansquared error (RMSE) is 3.2 percent, only slightly
larger than the standard error of the equation. For the
first six quarters of the simulation, the mean error is
0.33 percent and the RMSE is 2.23 percent, less than
the standard error in equation 6.
The importance of the temporary energy price ef­
fects emerges from the same simulation experiment
using equation 4, which ignores energy prices. The
simulated GNP growth rates and residuals are shown
in table 2. Ignoring temporary energy price effects
leads to over-estimates of GNP growth. The mean
error is 1.2 percent for the eight quarters and 1.0 per­
cent for the first six quarters, much larger than in
table 1. The size of each of the residuals in table 2
is generally larger than in table 1. The RMSE is larger

8


11/1979

0.17

Root-mean-squared error = 3.18%

1981

Mean error =

1.2

Root-mean-squared error = 3.50%
1Figures may not add exactly due to rounding.

than the standard error in equation 4 and larger than
in table 1.
Despite the quality of the simulation results for
equation 6, it must be noted that the economy has
seldom been forced to adjust to large changes in the
relative price of energy. Thus, the estimates in equa­
tions 5 and 6 may be heavily influenced by the par­
ticular events surrounding 1973-75 developments. To
examine this possibility, the sample period for equa­
tions 4 - 6 is extended to III/1980. A search for the
optimal lag structure was conducted again, using the
criterion and selection procedure described above. The
optimal lag structure is the same, the current and six
lagged values of the growth of the relative price of
energy. The sign pattern, magnitude and significance
of all the coefficients, including the relative price of
energy terms, are essentially unchanged when the
sample period is extended. The equations have about
the same adjusted R2 and standard error when the
sample period is extended. Estimated over the longer
sample period, equation 6 is:
(6 ') GNP, == 2.708 + 1.165 Z w l , M ,., - 0.002 Z w?., E,
(3.31) (8.05)
(-0 .0 3 ) J-°
- 0.453 S, -0.062 p?
(-3 .7 1 ) (-1 .9 3 )
+ 0.032 p?-, - 0.003 pf-2 - 0.045 p ”-3 -0.032 p? ,
(0.77 )
(-0 .0 8 )
(-1 .0 8 )
(-0 .7 6 )
+ 0.010 p ts + .099 pt-6.
(0.24)
(2.93)
= 0.54

S.E. = 2.96

D.W . = 1.99

JA N U A R Y

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

1981

C h a rt 2

C ontrib utio n of Energy Price C hanges (1 /1 9 7 0 —111/1980)
to G N P G ro w th a

1970

1971

197 2

197 3

197 4

1975

197 6

197 7

197 8

1979

1980

1981

1982

S o u rc e : E q u a tio n 6
|_L P e r c e n t a g e c h a n g e s a r e m e a s u r e d b y c h a n g e s in t h e l o g a r i t h m

o f th e le v e l o f t h e

gro ss

n a t io n a l P ro d u c t.

L a t e s t d a t a p lo t t e d : 4 th q u a r t e r

When equation 4 is estimated over the same sample
period (I/1955-III/1980), it too does not change sig­
nificantly (the standard error is 3.18 percent). The
F-statistic for the addition of the relative price of
energy terms, Fe>89 = 3.47, is significant at the 1 per­
cent level. The lag structure, size and significance of
the energy price effects in equation 6 do not appear
to be artifacts of the 1973-75 experience.
To provide a longer perspective on the relative
price of energy’s impact on GNP, as well as a more
balanced perspective on recent developments, chart
2 provides estimates of the impact of actual energy
price developments on GNP growth for each quarter
from 1/1970 to 11/1982 using the coefficients in equa­
tion 6. These estimates span three diverse periods
from a statistical view: the period I/1970-III/1978 is
within the sample period for equation 6; the period
IV/1978-III/1980 is that of the post-sample simula­



tion of equation 6; and the estimates for IV/1980-II/
1982 are based on the assumption that the relative
price of energy does not change in IV/1980-II/1982.
The chart shows that current and past energy price
changes exerted large negative impacts on GNP
growth from 1/1974 to 1/1975 and from 11/1979 to
III/1980. In the first instance, these changes were
offset by the subsequent positive effects of past
energy price increases in III/1975-I/1976. It remains
to be seen whether the large offsetting reactions of
GNP growth to past energy price changes shown from
IV/1980-IV/1981 will materialize.12
12An important caveat is necessary. The assumption that the
relative price of energy remains unchanged after III/1980
is included to illustrate the presence, size and pattern of
lagged effects of past energy prices on future GNP growth.
It is well known that the relative price o f energy will rise
over the year III/1980-III/1981 due to U.S. energy policy.
The quarterly timing of this increase, however, is not known
with a high degree of certainty.

F E D E R A L. R E S E R V E B A N K O F ST. LOUIS

JAN U ARY

ENERGY PRICES, THE MONEY-PRICE
LINK AND REAL GNP DEVELOPMENTS
The effect of a change in the relative price of
energy on the general level of prices can be ex­
amined in the context of a simple reduced-form
equation that focuses on the link between money and
prices. In particular, the rate of increase in the GNP
implicit price deflator is primarily determined by
growth in the stock of money. Prior evidence indi­
cates that the growth of the money stock over the
past 20 quarters (five years) is a significant determi­
nant of the rate of increase in prices.13 The period of
wage-price controls, which falls within the sample pe­
riod, had a significant impact on prices. Controls tem­
porarily reduced price increases, then temporarily
raised the rate of increase. Dummy variables are in­
cluded in the price equations estimated here to
account for these effects.14
To investigate the effect of changes in the relative
price of energy on the price level, current and lagged
values of the rate of change in the relative price of
energy are added to the reduced-form relationship
between money growth and rate of price increase Pt.
The basic price equation, without energy price vari­
ables, for the period I/1955-III/1978 is:
(7 ) P, = 1.020 Z W
(27.57)1=0
R2 = 0.75
13See Denis

t-i

M,-i -2.045 D1 + 2.625 D2.
(-3.99)
(5.30)

S.E. = 1.21

S. Karnosky, “ T he

D.W. = 1.66
Link Between

M oney and

Prices: 1970-76,” this Review (June 1976), pp. 17-23.
Karnosky shows the permanent impact of a higher relative
price of energy on the price level, and the absence of a
permanent wage and price control effect. The approach
below differs slightly. The relative price of energy is used
in the price equation instead of a dummy variable for the
energy price effect, and the timing of wage and price con­
trol effects is different. Also, Keith M. Carlson, “ The Lag
from Money to Prices,” this Review (O ctober 1980), pp.
3-10, argues that since 1970 the length of the lag for past
money growth has shortened to 12 quarters. This result does
not hold for equation 8 below. The optimal lag length for
the period I/1970-III/1978 for this equation is 22 quarters,
virtually the same as used here.
14For the control period, III/1971-I/1973, the dummy variable
D1 has a value of unity, and zero in other periods. The
dummy variable D2 has a value of unity in I/1973-I/1975
and zero otherwise, to capture the effects of the ending of
price controls. The choice of the periods for control and
decontrol effects is largely motivated by the findings reported
by Alan S. Blinder and William J. Newton, “ The 1971-1974
Controls Program and the Price Level: An Econometric
Post-Mortem, National Bureau of Economic Research, Inc.,
Working Paper No. 279 (September 1978). Their results,
for the monthly consumer price index, support the view that
the retarding effects of controls on inflation ended in early
1973 and that these effects were offset by “ catch-up” infla­
tion that began at that time and continued until the first
quarter of 1975. Earlier experiments with varying the tim­
ing of this specification resulted in higher standard errors
for the price equations 7 and 8.


10


1981

Table 3

Simulation of Equation 8
One-quarter
period ending

Actual P

Simulated P

Error1

IV/1978

9.3%

6.1%

-3.2%

1/1979

8.1

6.2

-1 .9

11/1979

7.5

6.5

-1.0

111/1979

7.5

7.5

0.0

IV /1979

7.8

8.6

0.7

1/1980

8.9

9.2

0.3

11/1980

9.4

8.4

-0.9

111/1980

8.8

9.6

0.8

Mean error =

-0.7

Root-mean-squared error = 1.48%
•Figures may not add exactly due to rounding.

Since a constant is not significant in any of the price
equations estimated, it is omitted. The sum of money
growth coefficients is not significantly different from
unity; the price control dummy variables are signifi­
cant and have the correct sign. A test of the hypothe­
sis that price controls had no permanent impact on
the price level could not be rejected at the 5 percent
significance level, although that constraint is not im­
posed here. Twenty lagged money growth rates were
included because, for a variety of sample periods
examined previously, this lag length is optimal ( mini­
mum standard error). A third-degree polynomial dis­
tributed lag with a tail constraint is used to estimate
the current and lagged money growth coefficients.
Up to 16 lagged values of the rate of change in the
relative price of energy were examined using an
unrestricted distributed lag. An F-test (5 percent sig­
nificance level) was used for the significance of addi­
tional lagged values and sets of lagged values. In
no case is the current energy price variable signifi­
cant (generally its t-statistic is less than one-half
in absolute value and usually has a negative sign),
so it is dropped. The optimal lag structure includes
four lagged values of the rate of increase in the rela­
tive price of energy. This equation is:
20

(8 ) P, - 0.990 Z Wt-1 M ,-, - 1.895 D1 + 1.388 D2
( 2 7 . 5 0 ) ( - 3 . 8 9 )
(2.28)
+ 0.014 p?_, + 0.044 pt_2 - 0.012 p^_3 + 0.029 pi-,.
(0.90)
(2.62)
(-0 .7 2 )
(2.07)
R2 = 0.78

S.E. = 1.15

D.W . = 1.74

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JA N U A R Y

1981

C h a rt 3

Contribution of Energy Price Changes (1 /1 9 7 0 —111/1980)
to the Rate of Increase of Prices ll

Source: E q u a tio n 8
|_1_ P e rc e n ta g e c h a n g e s a re m e a s u re d by c h a n g e s in the lo g a rith m o f the le v e l o f th e g ro s s n a tio n a l p ro d u c t d e fla to r.
Latest d a ta plotted: 4th q u a rte r

The F-statistic for the addition of the four lagged
energy price terms to equation 7, F4>86 = 3.63, is sig­
nificant at the 5 percent level.16

decline in potential output is not rejected. This rein­
forces the earlier result that a rise in the relative price
of energy has no permanent effect on nominal GNP.

The sum of the energy price effects on the level of
the GNP deflator in equation 8 is 0.075 ( S.E. =
.0235). For the sample period I/1955-III/1978, the
elasticity of potential private business sector output
with respect to the relative price of energy is -0.093,
according to equation 2. The price level elasticity of
the relative price of energy in equation 8 is not sig­
nificantly different from this estimate. Thus, the hypo­
thesis that the price level effect is the same as the

The results of a post-sample simulation of equation
8 are shown in table 3. The rate of price increase is
underestimated during late 1978 and early 1979. Be­
ginning in 11/1979, however, the errors are quite
small. The average error for the last six quarters
in the post-sample period is -0.01 percent. For the
eight-quarter period, the average error is -0.7 per­
cent. The RMSE of 1.5 percent is not significantly
larger than the standard error during the sample
period.18 These results contrast sharply with a simu­
lation of equation 7, which omits energy price

15Since the GNP and the price estimates are reduced-form
equations, the exogenous variables in each are potentially
the same. When the wage and price control dummies and
the additional lagged money terms included in equation 8
are added to the GNP equation 6, none o f the coefficients
is significant individually or as a group at the 5 percent
significance level. Thus, these variables are not included in
equation 6. Also, when the strike variable and expenditure
variables included in equation 6 are added to the price
equation 8, they too are insignificant (all t-statistics are less
than 0.4 in absolute value), so they are omitted. It can be
concluded that equations 6 and 8 are drawn from the same
model with a common set o f exogenous variables.




16When equations 7 and 8 are reestimated through the third
quarter of 1980, there are no important changes in the opti­
mum lag length, the coefficient estimates or the fit of the
equations. The standard error of equation 8 rises to 1.169
and the adjusted R- rises to 0.80. The pattern of energy
price coefficients remains the same and the sum effect for a
rise in the relative price of energy is 0.066, essentially the
same as above. The sum of the money growth coefficients
(1.015) remains essentially unity. The F-statistic for the
addition of the energy price coefficients is F4,M = 4.46, which
is significant at the 1 percent level.

F E D E R A L R E S E R V E B A N K O F ST. LOU IS

JA N U A R Y

1981

C ontribution of Energy Price C hanges (1 /1 9 7 0 —111/1980)
to Real G N P G ro w th i

1970

1971

1972

1973

1974

1975

1976

1977

1978

1979

1980

1981

1982

S o u rc e : E q u a tio n s 6 a n d t
[ P e r c e n t a g e c h a n g e s a r e m e a s u r e d b y c h a n g e s in t h e l o g a r i t h m

o f t h e le v e l o f r e a l g r o s s n a t i o n a l p r o d u c t .

Latest d a t a plotted: 4th q u a rte r

changes. For the same eight-quarter period, the simu­
lation of equation 7 underestimates inflation in every
quarter by an average of 1.8 percent (RMSE =
1.93). The differences are particularly large begin­
ning in the third quarter of 1979 when the simulation
error for equation 7 is -0.8 percent; thereafter, the
error is -0.9 percent, -1.9 percent, -2.6 percent and
-1.7 percent, respectively.
The impact of energy price changes on prices and
observed real output can be found from equations 8
and 6. Chart 3 shows the contribution of changes in
the relative price of energy to the rate of price in­
crease from I/1970-IV/1981 under the assumptions
used above for the effects on GNP growth. Changes
in the relative price of energy have had negligible
impacts on the GNP deflator except following the two
periods of sharp increases. In the first instance, the
rate of increase in the GNP deflator was raised on
average by over 2 percentage points during the four

12


quarters from 1/1974 to 1/1975. The same result
occurred from III/1979 to III/1980. On an annual
basis, the price level impact exceeded 0.6 percentage
points in only three years: 1974 and 1980, when the
impact was an additional 2.1 percentage points, and
1975, when it was 1.1 percentage points.
The GNP and price level effects are combined in
chart 4 to obtain real GNP effects. In general, chart
4 shows the negative permanent impact of the sharp
increase in relative energy costs in 1973-74 and 197980. This effect, however, is mixed with the transitory
impact associated with the dynamic adjustments of
output and prices due to the supply shock.
In table 4 the cumulative impact of a 40 percent
increase in the relative price of energy in the current
quarter is indicated for GNP, prices, and the difference,
real output.17 Note that after six quarters, there is no
effect on GNP, the price level is 3.0 percent higher
and real output is 3.0 percent lower. These develop­

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

ments illustrate the permanent effects of the energy
price increase. During the transition, however, GNP
is relatively lower and prices are affected somewhat
less than their permanent changes. Real output does
not fall as much as its permanent decline until two
quarters after the energy price rise. Subsequently,
real output overshoots its ultimate decline, then re­
turns to the level of the permanent decline. If the
permanent effect on real output is taken to be an
estimate of the immediate potential output effect, the
gap between potential output and actual real GNP
initially narrows so that the unemployment rate for
the labor force declines. Subsequently, actual out­
put is reduced relatively more than its permanent
decline so that the unemployment rate will tempo­
rarily rise. After six quarters, the decline in real GNP
is the permanent change. According to the theory,
the permanent decline arises because of a fall in
potential output and productivity. Consequently, the
unemployment rate would not be expected to change
beyond the period of transition.

ENERGY PRICES, THE MONETARY
GROWTH-UNEMPLOYMENT RATE LINK
Transitional unemployment can be examined using
a reduced-form equation similar to those above. The
general theoretical considerations that are useful here
are (1) that the economy tends to full-employment
equilibrium unless disturbed by shocks such as policyinduced fluctuations in aggregate demand or supply,
and (2 ) that demand-stimulus, especially through
changing the rate of money stock growth, can tempo­
rarily reduce the unemployment rate. These consider­
ations have been explored to a limited extent in a
reduced-form framework.18 The hypothesis that the
unemployment rate equals the full-employment un­
employment rate plus a component that reflects the
past history of money growth that leads to temporary
departures of the economy from full-employment
could not be rejected.
For the hypothesis examined here, changes in the
excess of the unemployment rate (U ) over a full17A once-and-for-all rise in the relative price o f energy o f 40
percent is equivalent to a 160 percent increase during the
current quarter, when measured at an annual rate. The GNP
effect is found by summing the energy price coefficients
times 160 in equation 6, and dividing by four to obtain
quarterly differences. The price effects are found by summing
the coefficients in equation 8, and again multiplying by
4 0 (1 6 0 /4 ).
18See John A. Tatom, “Does The Stage o f the Business Cycle
Affect the Inflation Rate?” this Review (September 1978),
pp. 7-15.




JA N U A R Y

1981

Table 4

The Effects of a 40 Percent Increase
in the Relative Price of Energy
Quarter

GNP

0

-2.14%

0

1

-0.20

0.54

-0.74

2

-1.45

2.31

-2.32

3

-2.44

1.82

-4.26

Prices
%

Real output
-2.14%

4

-4.45

3.00

-7.45

5

-4.06

3.00

-7.06

6

0

3.00

-3.00

7

0

3.00

-3.00

employment unemployment rate (Up) are taken as
the dependent variable, AUN, where UN = (U -U F).
The full-employment unemployment rate is that de­
veloped by Clark (1977).19 Changes in excess unem­
ployment are potentially a function of the exogenous
variables considered above.
An examination of such a relationship yields the
following results. First, the federal expenditure
growth variables and strike variable that enter the
GNP equation 6 are not significant in any of the
equations estimated. While the coefficient estimates
for current and past federal expenditure growth
variables have the expected sign pattern — initially
negative, then positive — none of the t-statistics for
the individual coefficients or sum coefficients is larger
than 0.4 in absolute value. In addition, the F-statistic
for the set of federal expenditure variables is less
than 0.1, so they are omitted below. Also, the strike
variable in equation 6 and the wage and price con­
trol dummy variables in the price equation 8 do not
have t-values in excess of one in any of the un­
employment equation estimates, so they too are
omitted.20 Finally, a constant term was not significant
in any of the estimated equations, so it is omitted.
19See Peter K. Clark, “ Potential GNP in the United States,
1948-80,” U.S. Productive Capacity: Estimating the Utiliza­
tion Gap, (St. Louis: Center for the Study o f American
Business, Washington University, 1977), pp. 21-66.
20This is in sharp contrast to the view that controls distorted
the observed relation of unemployment to output growth
expressed by Michael R. Darby, “ Price and W age Controls:
The First Two Years,” “ Price and W age Controls: Further
Evidence” in Karl Brunner and Allan H . Meltzer, eds.. The
Economics of Price and W age Controls, Camegie-Rochester
Conference on Public Policy Series, supplement to the
Journal of Monetary Economics, Volume 2 (19 76 ).

13

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

A search for the optimum lag structure for energy
price changes and monetary growth was conducted.
The criterion for the optimum lag for energy prices was
an F-test at the 5 percent significance level for the ad­
dition of past energy price changes. This test was con­
ducted for several specifications of the lag length (6
to 30 quarters) for current and past money growth
effects. In every case, the optimum structure includes
the past six quarters of relative energy price changes.
Since the current-quarter effect never has a t-value as
large as 0.5 in absolute value, it is omitted. The cri­
terion for selecting the optimum lag structure for a
third degree polynomial lag of current and past
money growth is to minimize the standard error of
the equation estimated with the six past energy price
terms, with and without the other variables discussed
above. The optimum lag structure in every case in­
cludes the current and nine past money growth rates.

JAN U AR Y

1981

Table 5

The Unemployment Equation
(1/1955-III /1978)
Dependent variable: A (U t Independent
variable

U f ,,)

Coefficient

t-statistic

M,

-0.021

-2.02

M.-,

-0.026

-4.65

M .-.

-0.023

-4.18

-0.015

-2.71

M .-,

-0.005

-0.98

M .-s

0.007

1.80

M .-6

0.017

3.95

M.-,

0.024

4.23

M«-»

0.025

4.06

M.-,

0.017

3.87

M ,-.

-0.001

-0.07

.

-0.006

-1.95

p?-2

0.005

1.49

P' 3

0.005

1.46

p t.

0.010

3.19

The choice of the 10-quarter period for money
growth effects is highly suspect, but fortunately it
does not affect the energy price estimates. In particu­
lar, changes in the unemployment rate are expected
to be a function of changes in the "GNP gap” in an
Okun’s Law framework. Changes in the GNP gap, in
turn, are a function of the growth rate of potential
output and the growth rate of actual output. Accord­
ing to the GNP and price results above, the growth
rate of actual output is affected by money stock
growth for about five years, so changes in the excess
unemployment rate would be expected to have the
same lag structure. In searching the lag space, equa­
tions with 22 lagged money growth rates had a local
minimum standard error for lags from 10 to 30 quar­
ters, and this standard error is 0.8 percent higher
(0.264 for equation 9) than with nine lagged terms.
None of the properties of equation 9 are altered when
22 lagged values of money growth are included. In
particular, the optimum lag, sign pattern, magnitude
and t-statistics for the individual energy price terms
are identical, as is the F-test for the addition of these
terms. The difference is that after the ninth lag,
money growth coefficients are initially small and posi­
tive, then small and negative with a sum that is not
significantly different from zero. Because of the cri­
terion adopted for selection of the optimum lag struc­
ture, and the independence of the energy price effects
to the lag structure choice, the shorter lag for money
growth is used here.

sequently, the excess unemployment rate is restored
to its initial level. The energy price terms add signifi­
cantly to the equation, while the sum effect is not
significantly different from zero, as hypothesized
above; the F-statistic for the addition of the lagged
values of the change in the relative price of energy
is F6,s6 = 6.25, which is significant at the 1 percent
level. Finally, as hypothesized above, a once-and-forall rise in the relative price of energy initially reduces
the unemployment rate. According to equation 9, the
sum of the coefficients for period t-1 and t-2 is nega­
tive; thereafter, the cumulative sum is positive until
period t-6, when the sum is positive but not signifi­
cantly greater than zero.

The unemployment rate equation 9 is presented in
table 5. Note that an increase in the rate of money
growth has a transitory effect, leading to reductions in
the excess unemployment rate for five quarters. Sub­

Equation 9 was also estimated with the sum of the
energy price coefficients set equal to zero. The Fstatistic for this constraint is F i)86 = 1.85 which is not
significant at the 5 percent level. Thus, the hypothesis

Digitized for14
FRASER


i
i= 0

P ?

0.005

1.57

-0.010

-3.10

0.009

1.36

P t-5
P<

«

£

pi,

R-

=

J-l

0.59

S.E.

=

0.262

D.W. = 1.82

p =

0.41

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JA N U A R Y

1981

C hart 5

Portion of the Change in the U nem ploym ent Rate
Due to Energy Price Changes (1 /1 9 7 0 —111/1980)
Percent

Percent

Source: E q u a tio n
La te s t d a t a p l o t t e d : 4 t h

9 , w i t h t h e su m o f t h e e n e r g y

price co efficie n ts

c o n s t r a i n e d to z e r o

q u a rte r

A post-sample simulation of equation 9 tracks
changes in the excess unemployment rate from IV /
1978 to III/1980 very well. The mean error for the
eight quarters is -0.015 percentage points. The RMSE
is 0.323, which is large relative to the standard error
of equation 9. However, in 11/1980 and III/1980
there are relatively large errors reflecting unusually

slower, then faster, GNP growth and so an unusually
larger, then smaller, rise in the unemployment rate.
For the first six quarters of the simulation, the RMSE
is only 0.177 percentage points, which is much smaller
than the standard error of equation 9. The mean error
for the first six quarters is the same as for the eight
quarters. This fit is also supported by extending the
sample period for equation 9 through the third quar­
ter of 1980. The adjusted R2 is 0.58 and the standard
error is 0.267. The sum statistics, and the pattern,
magnitude and t-statistics for the individual coeffi­
cients are virtually the same as for the earlier period.
The same results apply to the constrained version of
equation 9.

21The estimates and tests for equation 9 were also conducted
using an Almon polynomial to estimate the impact of energy
price changes. A second degree polynomial with no end­
point constraints proved superior to higher order polynom­
ials (third and fourth) for the energy price effect. The op­
timal lag is again six quarters for energy prices, and 10
quarters for money growth. The standard error of the equa­
tion is slightly lower, 0.261. Only the significance of the
energy price coefficients are noticeably changed by such an
estimation procedure. These coefficients from t-1 to t-6,
with t-statistics, are -0.006(^2.40), 0.004(2.86), 0.009(5.82),
0.009(5.60), 0.003(2.12) and -0 .0 0 3 (-3 .2 8 ). The adjusted
R2 for this equation is 0.59. A local minimum standard error
occurs with 21 lagged values of money growth (S.E. =
0.263) for lags up to 30 quarters.

Chart 5 shows the impact of actual increases in
energy prices on the change in the excess unemploy­
ment rate since the first quarter of 1970. The coeffi­
cients from the constrained version of equation 9 are
used to compute these effects. Generally the effects
are trivial, except in 1974-75 and in 1979-81. During
the first three quarters of 1974, the cumulative impact
of the energy price increase was to reduce the unem­
ployment rate by 0.5 percentage points. During the
next three quarters, the excess unemployment rate
rose 1.7 percentage points, and the difference was

that the significant effects of a rise in the relative
price of energy are temporary cannot be rejected. The
individual coefficient estimates, with t-statistics, for
the past six quarters are: -0.008(-2.77), 0.004(1.13),
0.003(1.05), 0.009(2.89), 0.004(1.24) and -0.011
(-3.81).21




15

FE D E R A L. R E S E R V E B A N K O F ST. LOUIS

largely offset in the last two quarters of 1975. On
average, the unemployment rate was 0.3 percentage
points lower in 1974 and 0.7 percentage points higher
in 1975 due to the 1973-74 energy price increases. For
the recent round of energy price increases, the esti­
mates indicate that the unemployment rate was low­
ered by about 0.3 percentage points in 1979, was
unaffected on average in 1980 and will be 0.3 points
higher in 1981 due to the economy’s dynamic adjust­
ment to higher energy prices.
Note that if the 1973-74 episode is dated from the
first quarter of 1974 to the first quarter of 1976, the
positive cumulative impact of the sharp increase in
energy prices occurs only in the four quarters of 1975
when it is 0.6, 1.2, 0.8 and 0.2 percentage points, re­
spectively. This period begins at the trough quarter
of the recession. In the second instance, if the impact
is summed beginning in the first quarter of 1979, the
cumulative impact is not positive until the third quar­
ter of 1980, when it is 0.3 percentage points, and in
the next three quarters, when it is about 0.5 percent­
age points. After mid-1981, the temporarily higher
unemployment rate is quickly eliminated by the dy­
namic functioning of the product and labor markets.
In each instance, the temporary increase in the
unemployment rate does not occur until the worst part
of the output reduction is complete and the economy
is apparently recovering on its own. Second, in each
case, when the unemployment rate is temporarily
high, the energy-price-induced component is a rela­
tively small part of the total. Finally, in each case
the highest levels of positive cumulative unemploy­
ment impacts associated with energy price develop­
ments have been quickly reversed. Of course, these
conclusions provide no support for exercising mone­
tary restraint in the face of sharp energy price and
price level surges. On the other hand, they do not
warrant even temporary demand stimulus.

SUMMARY AND CONCLUSION
The sharp increase in energy prices in 1979 and
1980 reduced both potential output and productivity,
and temporarily increased the inflation rate in the
same way, and to the same extent, as in 1974-75. In
addition, the absence of perfect price flexibility can
give rise to a transition to short-run equilibrium dur­
ing which total spending, actual output and the un­
employment rate are affected. These effects are
strongly supported by the empirical estimates for the
period ending in the third quarter of 1978 or in the

16


JAN U ARY

1981

third quarter of 1980. The results support the claim
that these effects are transitory.
The equation estimates indicate that a rise in the
relative price of energy reduces potential output im­
mediately but that the price level effect of this reduc­
tion occurs more slowly (over the subsequent year).
Initially, total spending is dominated by reduced out­
put with little change in prices; subsequently, prices
are increased. There are strong positive output and
GNP effects associated with these price increases
toward the end of a six-quarter adjustment period.
The output reduction due to an energy price increase
initially is smaller than the decline in potential out­
put, then overshoots it, before returning to the size
of the permanent decline. The pattern of unemploy­
ment rate developments matches this outcome: Ini­
tially, the unemployment rate declines, then rises to
higher levels before falling sufficiently so that, after
six-quarters, an energy price increase has no effect
on the unemployment rate.
The magnitude of the transitional effects on GNP
prices, output and the unemployment rate have been
estimated for the two sharp increases in the relative
price of energy in 1973-74 and 1979-80. In 1973-74
and early 1975 there were relatively large reductions
followed by relatively large increases in spending
growth associated with a rise in the relative price of
energy. On average, GNP growth was lowered 0.6
percentage points in 1973, 1.5 percentage points in
1974 and raised 1.5 percentage points in 1975. Due to
the 1979-80 episode, GNP growth is estimated to have
been 0.8 percentage points lower in 1979 and 2.0 per­
centage points lower in 1980. These effects are esti­
mated to be offset by faster GNP growth in 1981. The
extent of temporary inflation rate effects is estimated
to be largest in 1974 and 1980 when energy price de­
velopments temporarily added 2.1 percentage points to
measured inflation rates.
The temporary effects on real output growth are
reflected in unemployment rate developments. The
estimates show that energy price developments re­
duced the unemployment rate by 0.5 percentage
points during the first three quarters of 1974, then
raised it over the next three quarters, so that at the
peak of the unemployment rate in 11/1975, 1.2 per­
centage points were associated with energy price
increases. This transitional increase was eliminated
quickly. In 1979-80, the peak positive impact of energy
price increases is about 0.5 percentage points late in
1980 and early 1981; this impact is estimated to be
eliminated by the end of 1981.

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JAN U AR Y

The empirical investigation is conducted so that
the energy price effects are estimated using data from
the period prior to the recent episode of price in­
creases. Aside from providing a stronger test of the
hypotheses using the 1979-80 increases, the approach
provides an opportunity to examine the impact of
the increased emphasis on money growth reductions
announced in November 1978 and reinforced by the
announcement of procedural changes in October 1979.

1981

The simulations for GNP, inflation and unemployment
conducted from IV/1978 to III/1980 indicate no
change in the basic reduced-form relationships and
no independent impact of these announcements or any
actions intended to implement the slowing of money
growth. Instead, the reduced-form relationships ap­
pear to explain spending, price, output and unemploy­
ment rate developments as well as they did previously.

Appendix 1
The GNP Results Using an Almon Lag
for Energy Price Changes
The purpose of this appendix is to provide comparable
estimates to equation 5 using an Almon polynomial dis­
tributed lag rather than an ordinary distributed lag. When
equation 5 is estimated for the period I/1955-III/1978
using both third and fourth degree polynomials, with and
without end-point constraints, for up to 16 lagged terms
for the growth in the relative price of energy, the “best”
equation is found using the third degree polynomial with
six lagged terms without end-point constraints. The speci­
fication for the other variables is the same as in equation
5 in the text. The estimated equation for the period
I/1955-III/1978 is:
(1 .1 ) GNP. = 2.681 + 1.124 Z w,°-i M,_, - 0.003 Z w?., E,-j
(3.25)
( 7 . 5 9 ) ( - 0 . 0 4 ) J*°
- 0.475 S, + 0.024 1
(-3 .8 3 )
(0 .4 4 )k-°
R- = 0.53

wiU p,'-*

S.E. = 2.93

D.W . = 1.96

where the actual coefficients on pf-» are:
current

-0.028 (-0 .9 6 )

t-1

-0.0 1 0 (-0 .6 5 )

t-4

-0.041 (-2 .3 4 )

t-2

-0.001 (-0 .0 7 )

t-5

-0.004 (-0 .2 2 )

t-3

-0 .0 2 9 (-2 .1 8 )

t 6

0.116( 4.01)

This estimate is similar to equation 5 in the text. The sum




of the money growth coefficients is not significantly differ­
ent from one, and the sums of the expenditure growth
variables and energy price change variables are each not
significantly different from zero. The pattern of energy
price effects and magnitude are the same as in equation
5. The fit of the equation is essentially the same as for
equation 5. An F-test of the three additional coefficients
estimated in equation 5 indicates they do not add signifi­
cantly to the explanatory power of equation 1.1. The
F-statistic for the addition of the energy price variables
to equation 4 in the text is F«,83 = 4.43, which is signifi­
cant at the 1 percent level. When equation 1.1 is used to
simulate GNP in the eight-quarter post-sample period, the
results are essentially the same as the results in table 1 in
the text.
When equation 1.1 is estimated over the sample period
ending in III/1980, the optimal lag length and polynomial
degree remain the same, as do the other properties de­
scribed above. The F-statistic for the addition of the en­
ergy price variables is F«.»i = 5.14, which is significant at
the 1 percent level. The coefficients on the changes in the
relative price of energy (from current to t-6) with t-statistics are: -0.041(-1.56), 0.008(0.56), 0.003(0.17),
-0.023(-1.96), -0.036(-2.14), 0.003(0.15) and 0.118
(4.17). The sum of the energy price coefficients is 0.024
(0.44). The adjusted R2 of the equation is 0.54 and the
standard error is 2.93 percent. The Durbin-Watson sta­
tistic is 2.00.

17

Unreal Estimates
of the Real Rate of Interest
W. W. BROWN and G. J. SANTONI

J n the nearly five decades since the publication of
Irving Fisher’s The Theory of Interest,1 economists
have engaged in numerous attempts to measure the
ex ante real rate of interest. The effort devoted to ob­
taining these estimates reflects the fact that the ex
ante real interest rate conveys information about
some fundamental economic relationships. The ex ante
real interest rate is the expected net rate of increase
in wealth arising from additional investment. Alterna­
tively, it can be viewed as the value of present con­
sumption in terms of future income and, consequently,
is implicit in the relative price of present consumption
in terms of capital goods. Each of these is reconciled
with the others by the profit-seeking market activity
of individuals.2
Like other relative prices, the ex ante real interest
rate enters the optimizing calculus of individuals and
ultimately affects resource allocation. Each decision
an individual makes, to save or invest or to change
current consumption relative to either of these, is a
choice which, implicitly at least, involves considera­
tion of the ex ante real interest rate.
Changes in the ex ante real interest rate transmit
information about changes in the relative values of
resources employed in alternative uses and eventually
result in a reallocation of resources to higher valued
The authors are associate professors of economics at California
State University, Northridge. Santoni is a Visiting Scholar at
the Federal Reserve Bank of St. Louis.
1Irving Fisher, The Theory of Interest and Capital (N ew York:
Augustus M. Kelley, 1965).
2For a more complete discussion see Armen Alchian and W il­
liam Allen, Exchange and Production: Competition, Coordina­
tion and Control (Belmont, California: Wadsworth, 1977), pp.
435-36.

Digitized for18
FRASER


uses. Changes in this interest rate reflect changes in
the net demand for present consumption goods rela­
tive to future consumption goods. The allocation of
present resources to the production of these goods
will be redirected in response to the change in their
relative values.
Since all goods are more or less durable (i.e., they
yield consumption streams which persist over vary­
ing lengths of time), the reallocation of present re­
sources resulting from a change in the ex ante real
interest rate will pervade all markets. In the absence
of information about the movement of the ex ante
real interest rate, it is difficult to distinguish “disturb­
ances” (resource reallocation) induced by shifts in
the demand for present consumption goods relative
to future consumption goods from those caused by
shifts in aggregate demand for both present and fu­
ture goods. From the point of view of the policy­
maker, the distinction is crucial. If the disturbance is
the result of a shift in relative demands, resources
will be reallocated to higher-valued uses and com­
munity net wealth will rise. If the disturbance is the
result of a shift in aggregate demand, any temporary
reallocation of resources occurring during the disturb­
ance must be to lower-valued uses causing community
net wealth to fall. Policymakers might wish to elimi­
nate the latter result but should not attempt to re­
tard the former.
While information about changes in the ex ante
real interest rate is valuable to the policymaker, it is
difficult to obtain. The ex ante real interest rate re­
flects the expectations of individuals regarding future
events. As such it can not be directly observed. It is,
of course, possible (and inexpensive) to observe the

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

consequences of decisions that are made on the basis
of these expectations. The wealth consequences asso­
ciated with any economic decision can always be cal­
culated after the fact. However, this ex post real rate
of return does not bear on economic decisions since
it is only known after these decisions have been made.
Unlike the ex ante real rate of interest, the ex post
real rate of return is irrelevant to the process of re­
source allocation.
Since the ex ante real interest rate can not be
observed directly, individuals interested in estimating
its magnitude have been led to employ the simple
Fisherian relationship that the nominal (market) rate
of interest is equal to the sum of the ex ante real rate
of interest and the anticipated rate of inflation in the
general level of prices. The relationship implies that
empirical estimates of the ex ante real interest rate
can be obtained by subtracting some measure of the
anticipated rate of inflation in the general level of
prices from the nominal rate of interest. As a result,
previous estimates of the ex ante real interest rate
have turned on the complicated problem of measur­
ing the anticipated rate of inflation.
Virtually all previous studies have dealt with this
problem by modeling the anticipated rate of inflation
in the general level of prices as some function of past
changes in the consumer price index (CPI) or GNP
deflator.3 If the real rate of interest is not changing,
this method may produce “reasonably” accurate esti­
mates of the anticipated rate of inflation in the gen­
eral level of prices. Unfortunately, if the real rate of
interest is itself changing, these commonly used price
indices will produce biased estimates of actual changes
in the general level of prices. Consequently, use of
these indices to proxy expected future price level
3Recent examples include Albert E. Burger, “ An Explanation of
Movements in Short-Term Interest Rates,” this Review (July
1976), pp. 10-22; John A. Carlson, “ Short-Term Interest
Rates as Predictors of Inflation: Comment,” American Economic
Review (June 1977), pp. 469-75; Michael Echols and Jan
Walter Elliot, “ Rational Expectations in a Disequilibrium
M odel of the Term Structure,” American Economic Review
(M arch 1976), pp. 28-44; Jan Walter Elliot, “ Measuring the
Expected Real Rate of Interest: An Exploration of Macroeco­
nomic Alternatives,” American Economic Review (June 1977),
pp. 429-44; Eugene F. Fama, “ Short-Term Interest Rates as
Predictors of Inflation,” American Economic Review (June
1975), pp. 269-82; Eugene F. Fama, “ Inflation Uncertainty
and Expected Returns on Treasury Bills,” Journal of Political
Economy (June 1976), pp. 427-48; Martin Feldstein and Otto
Eckstein, “ The Fundamental Determinants of the Interest
Rate,” The Review of Economics and Statistics (November
1970), pp. 363-75; P. J. Hess and J. L. Bicksler, “ Capital
Asset Prices Versus Time Series Models as Predictors of Infla­
tion,” Journal of Financial Economics (D ecem ber 1975), pp.
341-60; William P. Yohe and Denis S. Karnosky, “ Interest
Rates and Price Level Changes, 1952-1969,” this Review
(Decem ber 1969), pp. 18-38.




JAN U AR Y

1981

changes in Fisher’s equation will prejudice measure­
ment of both the level and direction of movement of
the real rate of interest.4
This particular problem arises in a number of recent
articles dealing with the inflationary period since the
late 1960s which have reported sharply declining
and negative ex ante real rates in 1974 and 1975.5
The theoretical possibility of a negative ex ante real
rate of interest is not at issue here.6 Casual observa­
tion suggests that the preconditions for a negative ex
ante real interest rate do not now exist, nor did they
exist in 1974 and 1975.7 More importantly, however,
sharply declining ex ante real rates imply specific
kinds of economic adjustments which were contrary
to those that actually occurred during this period.
The purpose of this article is to demonstrate that
the estimates of the ex ante real rate obtained by
these previous studies are spurious. Following Alchian
and Klein,8 it is first demonstrated that, when real
rates of interest are rising, commonly used price
indices will overstate changes in the general level of
prices. This introduces a downward bias into esti­
mates of the real rate of interest when the estimates
depend on measured changes in these price indices.
Secondly, evidence is presented which indicates that
the ex ante real rate of interest increased during

4This bias exists apart from the tax and uncertainty effects
noted by others. See, for example, James E. Pesando and L.
Smith, “ Tax Effects, Price Expectations and the Nominal Rate
of Interest,” Economic Inquiry (June 1976), pp. 259-69;
Michael Darby, “ The Financial and Tax Effects of Monetary
Policy on Interest Rates,” Economic Inquiry (June 1975), pp.
226-76; Y. Amihud and A. Bamea, “ A Note on Fisher Hypo­
thesis and Price Level Uncertainty,” Journal of Financial and
Quantitative Analysis (September 1977), pp. 525-29.
5See for example Elliot, “ Measuring the Expected Real Rate
of Interest: An Exploration of Macroeconomic Alternatives;”
Fama, “ Interest Rates as Predictors of Inflation;” Hess and
Bicksler, “ Capital Asset Prices Versus Time Series Models as
Predictors of Inflation;” Pesando, “ On the Efficiency of the
Bond Market: Some Canadian Evidence,” Journal of Political
Economy (D ecem ber 1978), pp. 1057-76.
6Like Fisher, who discusses negative rates in the context of
shipwrecked sailors whose store of figs is deteriorating, we
think that “ The fact we seldom see an example of zero or
negative interest rates is because of the accident that we
happen to live in an environment so entirely different . . .”
(Fisher, The Theory of Interest and Capital, p. 192).
7Such preconditions would imply “ . . . a world in which the
only provisioning for the future consisted in carrying over
initial stocks of perishable food, clothing and so forth and if
every unit so carried over into the future were predestined
to melt away . .
(Fisher, The Theory of Interest and
Capital, p. 91).
8Armen Alchian and Benjamin Klein, “ On a Correct Measure of
Inflation,” Journal of Money, Credit and Banking (February
1973), pp. 173-91.

19

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

1974-1975. These results suggest that the previously
reported falling and/or negative estimates of the ex
ante real rate are statistical artifacts. To put it di­
rectly, they are nothing more than the predictably
spurious consequences of the method used to generate
them.

MEASUREMENT OF THE REAL RATE
The methodology commonly used in measuring the
real rate of interest is represented by the following
three equations:
(1 )

r = i - P.

(2 )

Pe = f ( C ) , f' > 0

( 3)

J = i - P.,

Equation 1 states the familiar theoretical relation­
ship developed by Fisher between the ex ante real
rate of interest (r), the observed nominal rate of
interest (i) and the anticipated future rate of infla­
tion (P,,), assuming continuous compounding. Equa­
tion 2 characterizes the methodology commonly em­
ployed in estimating the anticipated rate of inflation.
It indicates that ^estimates of the anticipated future
rate of inflation ( Pc) are obtained from observation of
past changes in some price index (C ).lJ
Finally, equation 3 states that estimates of^the ex
ante real rate (r) are derived by subtracting P,, from
the observed nominal rate of interest.
Since neither r nor P«. is directly observable, the
validity of this process for accurately estimating the
9The index most frequently used is the CPI. See Burger, “ An
Explanation of Movements in Short-Term Interest Rates;”
Elliot, “ Measuring the Expected Real Rate of Interest: An
Exploration of Macroeconomic Alternatives;” Fama, “ Infla­
tion Uncertainty and Expected Returns on Treasury Bills;”
Hess and Bicksler, “ Capital Asset Price Versus Time Series
Models as Predictors of Inflation;” Yohe and Karnosky,
“ Interest Rates and Price Level Changes, 1952-1969.” The
GNP deflator has been used less frequently. See Feldstein
and Eckstein, “ The Fundamental Determinants of the Interest
Rate.” The procedure used to estimate expected inflation for
period t from the observation of past levels of some price
index is, roughly, the following: An estimate of the period t
price level is made in period t-1. This estimate is a weighted
average of past price levels. That is,

C, = I W iC,;
t-i

i= t -i

where the left-hand term is the estimate and the W are the
weights assigned to past price levels. The estimated change
in the price level is obtained by subtracting the price level
in period t-1 from the estimate for period t as follows

ACt ~ ,-iC, —Ct-i.
Last, the estimated change in the price level is defined to be
the estimate of expected inflation for period t,

ACt

- Pet •


20


JA N U A R Y

1981

ex ante real rate depends crucially ^on whether P(, is
a reliable proxy for P(.. Typically, P,. is regarded as
“good” or “bad” depending on how well it predicts
the actual contemporaneous rate of change in the
particular price index being used. The implicit assump­
tion is, of course, that contemporaneous changes in
the index reflect true changes in the general level
of prices.
Fama’s justification of his use of the CPI is fairly
typical. He comments:
The Bureau of Labor Statistics Consumer Price In­
dex (CPI) is used to estimate AP, the rate of change
in the purchasing power of money from the end of
month t-1 to the end of month t. The use of any
index to measure the level of prices of consumption
goods can be questioned. There is, however, no need
to speculate about the effects of shortcomings of the
data on the tests. If the results of the tests seem
meaningful, the data are probably adequate.10
Several authors have questioned whether functions
of past rates of change in the CPI, or GNP deflator,
serve as reliable predictors of expectations regarding
future price level change.11 Others have commented
on how measurement errors in the indices must be
taken into account when estimating real interest
rates.12 None, however, have tried to confirm the
validity of the estimates by observing economic rela­
tionships known to depend on the real rate of interest.
Alchian and Klein have noted a significant difficulty
in using changes in common price indices as measures
of changes in the general level of prices, or “purchas­
ing power of money.” In particular, they argue that
changes in the purchasing power of money are deter­
mined by changes in the prices of both present con­
sumption goods and long-lived assets, not just changes
in the prices of present consumption goods alone.
They comment:
The analysis . . . bases a price index on the Fish­
erian tradition of a proper definition of intertemporal
consumption and leads to the conclusion that a price
10Fama, “ Short-Term Interest Rates as Predictors of Inflation,”
p. 247.
11See Carlson “ Short-Term Interest Rates as Predictors of In­
flation,” Edward J. Kane and Burton G. Malkiel, “ Autore­
gressive and Nonautoregressive Elements in Cross-Section
Forecasts of Inflation,” Econometrica (January 1976), pp.
1-16.
1;!See Fama, “ Inflation Uncertainty and Expected Returns on
Treasury Bills;” Feldstein and Eckstein, “ The Fundamental
Determinants of the Interest Rate;” Kane and Malkiel, “ Auto­
regressive and Nonautoregressive Elements in Cross-Section
Forecasts of Inflation;” C. Nelson and G. Schwart, “ ShortTerm Interest Rates as Predictors of Inflation: On Testing
the Hypothesis that the Real Rate of Interest is Constant, ’
American Economic Review (June 1977), pp. 478-86.

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

index u sed to m easure inflation must include asset
p rices (italics a d d e d ). A correct measure o f changes
in the nom inal m on ey cost o f a given utility level is
a p rice index for wealth. I f m onetary im pulses are
transmitted to the real sector o f the econ om y b y p ro­
ducing transient changes in the relative prices o f
service flows and assets, (i.e., b y produ cin g short-run
changes in ‘the’ real rate o f interest), then the co m ­
m only used, incom plete, current flow price indices
provide biased short-run measures o f changes in the
‘purchasing p ow er o f m on ey.’ 13

The CPI and GNP deflator largely exclude the
prices of long-lived goods and existing capital assets.14
Consequently, changes in these price indices will de­
pend on changes in the real rate of interest because
of the well-known difference in the interest elasticities
of the market prices of short- and long-lived goods.

JAN U A R Y

1981

output is also unchanged, there will be no change in
the general level of money prices or the level reflected
in a Fisherian price index (i.e., one which includes
asset prices). However, since the prices of short­
lived goods rise relative to the prices of long-lived
goods when the real interest rate rises, the money
prices of short-lived goods (long-lived goods) will
rise (fall) relative to the general level of money
prices. Thus, when the real interest rate is rising,
commonly used price indices, in which the prices of
short-lived goods receive a relatively heavy weight,
will rise introducing a systematic upward bias into
the estimation of changes in the general level of
prices. The reverse holds when the real interest rate
falls.

Our criticism of the methodology currently used
to measure the ex ante real rate of interest rests on
two interrelated points. First, the quantity weights
used in calculating the CPI and GNP deflators do
not accurately reflect the mix of goods actually avail­
able to individuals. As a result, changes in these
commonly used price indices produce biased esti­
mates of actual changes in the general level of prices
when the real interest rate is changing. Second, given
that it is the expectation of market participants con­
cerning the future rate of inflation in the general
level of prices that is relevant in Fisher’s theory of
the nominal rate of interest, estimates of the real in­
terest rate that employ past changes in a commonly
used price index as a proxy for expected inflation will
be biased when the real rate is changing. Each of
these points is demonstrated below.

If an increase in the real interest rate produces an
increase in the general level of money prices through
a once-and-for-all rise in velocity, the resulting in­
crease in commonly used price indices will contain
two components: 1) an increase due to the rise in
the general level of prices and 2) an increase due to
the bias introduced by capturing only part of the
price changes that have occurred. However, wealthmaximizing market participants will ignore both of
these components in forming their expectation re­
garding the future rate of inflation in the general
level of prices. They will ignore the first component
because it represents a once-and-for-all change which
leaves the future rate of inflation unaffected. They
will ignore the second component because its effect
is to overstate the true change in the general price
level. On the other hand, estimates of price expecta­
tions that employ the common methodology (the
ability to reproduce actual changes in the CPI) will
include both.

Point 1: Changes in the General Level of
Prices versus Changes in Commonly
Used Price Indices

This argument can be presented more formally.
Assume there are two kinds of goods — short-lived,
Qs, and long-lived, QL— and money. Suppose, in the
base period, the real rate of interest is r0. Then,

THE MEASUREMENT PROBLEM

Assume initially that an increase in the real rate of
interest occurs and that both the quantity of money
and its velocity are unchanged.15 If the quantity of
13Alchian and Klein, “ On a Correct Measure of Inflation,”
p. 173.
14Durable goods have a weight of 18.75 percent in the CPI.
Nondurable goods and services have weights of 47.19 and
34.03 percent, respectively. See Bureau of Labor Statistics,
Handbook of Methods, Bulletin 1910, 1976. The GNP de­
flator includes the prices of currently produced capital
goods but it excludes the prices of existing capital assets.
15Economic theory suggests that velocity will rise with an
increase in r. This is discussed below.




(4 )

M. • V„ = P„s • Q„s + P»L- Q„l

where M0 is the money supply, V0 is velocity, and Po
and PJ are the prices of short- and long-lived goods,
respectively.
If the interest rate increases to r1; velocity will rise
as relative prices change.16 Let
16Quantities will eventually adjust as well but that is ignored
here. In any case, the quantity adjustment which takes
place makes no difference for the measurement of the
change in a fixed weight index.
21

JA N U A R Y

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

(5)

p f - q.s + p,l - q,:

Fi

PoS • QoS + PoL • Qo

represent the level of a Fisherian price index in the
current period. If the change in the interest rate was
the only change that affected the index between the
base and current period, the change in the Fisherian
price index is
(6 )

AF = Fi - 1.

Let
(7 )

C, =

Pf • Qo

AC = C, - 1.

It is a simple matter to show that an increase in
the real rate of interest will have a greater effect on
the commonly used price index than on the Fisherian
price index. We know that
(9 )

P?/P,L > PoS/P„L

because a rise in the real rate of interest increases
the price of short-lived goods relative to long-lived
goods. Now consider the Fisherian index which can
be written as
F> —

Point 2: Biased Estimates of the
Real Interest Rate
If r remains unchanged, changes in commonly used
price indices accurately reflect changes in a Fisherian
index of prices. Consequently, the methodology sum­
marized in equations 1-3 will yield accurate esti­
mates of r for such periods. However, during periods
in which r is changing, bias in the common price
indices introduces, through equations 2 and 3, bias
into any estimate of the real interest rate that em­
ploys these indices.

P? • Qo

represent the level of a commonly used price index
in the current period. It differs from the Fisherian
index in that it excludes prices of long-lived goods.
The change in this price index, due to the change in
r occurring between the base period and the current
period, is
(8 )

1981

P? , , rQ ,? + (P,L/Pf)Q„L

X L

P? ' ' LQoS +

(P o
L/P o
S)Q.l1'

That is,
QoS + (P,L/Pf)QoL
LQos +

( p 0l/ p ;s)Q o
l

The term in the brackets is less than one since, from

(9),
pjyp? < PoL/Pos

To demonstrate this second point, ignore other fac­
tors that affect common price indices (e.g., a change
in the monetary growth rate) and express C as a
function of the real rate of interest. That is,
(10)

C = 0( r ), 0' > 0.

The error generated in estimating the real interest
rate by the method employed in the studies refer­
enced earlier is given by
( 11)

f - r = P. - f ( 0 ( r ) ).

The error in estimated changes in the real rate is
obtained by differentiating equation 11 with respect to
r. In doing so, note that the price expectations (Pe) of
market participants are based upon the anticipated
future rate of change in the general level of prices
in the sense of Fisher’s theory and not upon onceand-for-all changes produced by changes in r. Hence,
price expectations will be unaffected by changes in r
while the estimate of price expectations will vary
positively with such changes. That is,
dr

d£ d0

dr

o0 dr

(12) — = 1 - — — .
The term df . 3 0 is always positive. Estimates of
dfi

dr

changes in the ex ante real rate of interest will always
understate any actual change that occurs.

and thus
Qos + (P,VPIs)QoL< Q«s + (P„L/Pos)QoL.

It follows that Fi < C t and AF < AC.
In general, when the real interest rate is increasing,
use of price indices that are based primarily on short­
lived goods will introduce a systematic upward bias
into estimation of changes in the general level of
prices (in the Fisherian sense). The reverse is true
during periods of decline in the real interest rate.17
17Interestingly, Alchian and Klein commented on this source
of inherent measurement error in the CPI and CNP de-


22


Even worse, the procedure employed in previous
work can err in assessing the direction of change in
the real rate. If the effect of a change in the interest
rate on the commonly used price index described in
flator, but did not pursue its implications for estimating the
real rate of interest. They remark: “ It should be noted that
although our discussion emphasizes that movements in asset
and service prices differ largely because of differing rates of
adjustment to cyclical monetary disturbances there may also
be a significant secular bias due to changing equilibrium
real asset yields. ( The apparent increase in real rates of
interest over the years is ignored in our discussion.)” Alchian
and Klein, “ On a Correct Measure of Inflation,” p. 180.

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JA N U A R Y

1981

Table 1

Selected Estimates of the Real Rate of Interest1
Year

Elliot
short-term

Carlson
T-bill rate

St. Louis Fed
yield on high
grade corp. bonds

Ex post
short-term
yield

1970

0.57%

2.38%

2.86%

2.58%

1971

1.69

1.CS5

2.18

2.02

1972

2.13

1.28

2.72

2.52

1973

1.07

2.35

2.84

2.10

1974

-0.41

0.40

1.78

0.28

1975

—

0.07

0.05

-2.25

'T h e interest rate we report is the annual average of the various subperiods. In the case of
Elliot, we report his neo-Keynesian monetaiy estimate which he accepts as most accurate.
The Federal Reserve Bank of St. Louis discontinued publishing estimates prior to the end of
1975. The estimate we attribute to them for 1975 is one that we calculate using their method
of estimation.

equation 12 is sufficiently large, dr will be negative.
dr
Hence, even though the change in the real rate is
positive, the estimated change could be negative. This
may explain the declining estimated real rates re­
ported for the mid-1970s.

EVIDENCE ON CHANGES IN THE
REAL RATE
Table 1 presents some previously reported esti­
mates of the ex ante real rate of interest from 1970
to 1975. Additionally, it presents the difference be­
tween current short-term market rates and contem­
poraneous rates of change in the CPI. The latter
would represent the “true” ex post yield if changes
in the CPI measured changes in the general level of
prices without error.
All of these estimates show dramatic declines in
1974 and 1975, years in which substantial increases
were recorded in the CPI. Elliot’s reaction to his
results is perhaps typical. He asserts:
. . . som e relationship appears to exist betw een the
tem poral pattern o f the real rate and the current rate
o f inflation. . . . T h e negative and statistically signifi­
cant nature o f this relationship suggest that expected
real rates are systematically low ered when the most
current realized rate o f inflation is increasing.18
18Elliot, “ Measuring the Expected Real Rate of Interest: An
Exploration of Macroeconomic Alternatives,” p. 442. For
similar statements see Carlson, "Short-Term Interest Rates
as Predictors of Inflation: Comment,” p. 472; Feldstein and




However, before concluding that changes in the
CPI affect the real rate of interest, it seems appropri­
ate to determine whether other evidence is consistent
with this hypothesis. Changes in the ex ante real rate
of interest imply specific behavior in the prices of
long-lived assets relative to the prices of short-lived
assets. Falling real rates of interest in 1974 and 1975
should have been accompanied by a rise in the pres­
ent prices of long-lived assets (which produce future
consumption services) relative to the prices of short­
lived goods. Evidence indicates, however, that the
relative price of long-lived assets fell during 1974
and 1975. This evidence is inconsistent with the con­
tention that the ex ante real rate of interest declined
precipitously during this period.

SOME EVIDENCE FROM
INDIVIDUAL MARKETS
The movement of relative prices in various markets
is examined below. As noted earlier, a change in the
ex ante real rate of interest shows up as a change in
the relative price of less durable (present) goods in
terms of more durable (capital) goods. An increase
in the ex ante real rate of interest reflects an increase
in the demand for present goods relative to capital
goods. Consequently, the price of present goods in
terms of capital goods will rise. This adjustment in
relative prices mirrors the change in the ex ante real
interest rate.
Eckstein, “ The Fundamental Determinants of the Interest
Rate,” p. 366; Yohe and Karnosky, “ Interest Rates and Price
Level Changes, 1952-1969,” p. 24 and p. 26.

23

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

By its nature, this type of evidence requires exami­
nation of price movements in individual markets. This
procedure of examining relative price movements is
always open to the charge that any observed relative
price change in an individual market may be due to
circumstances unrelated to a change in the ex ante
real interest rate. As was noted previously, however, a
change in the ex ante real interest rate pervades all
markets. If an examination of a number of markets
reveals that the price of the less durable good has
consistently moved in the same direction relative to
the price of the more durable good, the contention
that the observed change in relative price is due to
the impact of special circumstances in each of these
markets loses much of its force.
Since the ex ante real interest rate can not be di­
rectly observed, any evidence about its magnitude or
direction of change will always be circumstantial. The
evidence presented below is no exception. However,
as Thoreau has noted, “ (s)om e circumstantial evi­
dence is very strong, as when you find a trout in the
milk.”
The evidence presented below is reasonably con­
sistent across the various markets for the 1968-1975
period. Moreover, changes in the price ratios examined
correspond perfectly across markets for the 1972-1975
period. However, the direction of change in the ex
ante real interest rate implied by these price ratio
changes occurring during the later period contradicts
that reported in previous studies. This contradiction
is perhaps not surprising. We have shown that past
increases in the real rate will introduce a downward
bias into estimates of the present change in the ex
ante real interest rate. Examination of changes in the
price ratios occurring in all four markets indicates an
increase in the ex ante real interest rate in the two
years immediately preceding 1974. Three of the four
markets indicate an increase in the real rate in the
three years immediately preceding 1974. The above
contradiction is the “trout” whose presence is verified
by this evidence.

JA N U A R Y

1981

Table 2

Spot and Futures Prices
1924-1926 = 100
Year

Index of
spot prices

Index of
futures prices

Ratio of spot
prices to
futures prices

1960

141.80

141.22

1.004

1961

149.85

148.44

1.009

1962

149.85

143.90

1.041

1963

159.83

154.49

1.034

1964

142.99

136.82

1.045

1965

142.47

139.31

1.022

1966

139.44

136.71

1.019

1967

142.88

141.79

1.007

1968

144.45

143.26

1.008

1969

144.90

139.10

1.041

1970

145.07

144.81

1.001

1971

144.35

146.30

.986

1972

189.49

184.58

1.026

1973

340.51

320.50

1.062

1974

384.53

357.26

1.076

1975

296.33

287.88

1.029

SOURCE: The Dow Jones Commodities Handbook, Dow
Jones Company, New York 1977, pp. 178-179.

the Dow Jones index of spot prices to the Dow Jones
index of futures prices was 1.019, with a standard
deviation of .018 (see table 2). Between 1973 and
1975 this ratio averaged 1.057. In 1974, when previous
studies report a precipitous decline in the real rate
(see table 1), the ratio reached its highest level
(1.076) in the entire 16-year period. Relative price
behavior in the commodities markets is inconsistent
with a falling ex ante real rate of interest in 1974
and 1975.

2. Durable and Nondurable Goods: Durable goods,
by definition, embody a longer-lived stream of future
1.
The Commodity Markets: Changes in the real services than do nondurable goods. Therefore, falling
rate of interest will be reflected in changes in spot
real rates of interest imply a decrease in the price of
relative to futures prices. The spot price of a good is
nondurable goods relative to the price of durable
today’s price for delivery today while the futures price
goods.
is today’s price for delivery in the future. A decrease
From 1960 to 1972 the average ratio of the U.S.
in the real rate must be reflected in a decrease in the
Bureau of Labor Statistics’ index of nondurable goods
value of present (spot) goods relative to future goods.
prices to its index of durable goods prices was .976
Spot prices will fall relative to futures prices when the
(table 3), with a standard deviation of .040. Between
ex ante real rate of interest falls.
1973 and 1975 it averaged 1.122. In 1974 it was 1.156.
Between 1960 and 1972 the average annual ratio of
Again, this relative price behavior is inconsistent with

24FRASER
Digitized for


F E D E R A L R E S E R V E B A N K O F ST

LOUIS

JA N U A R Y

Table 3

Table 4

Nondurable and Durable Goods Prices

Ratios of Earnings/Stock Prices
and Price of Nondurable Goods/
Stock Prices

Year

Index of
nondurable
goods prices

1960

89.4

Index of
durable
goods prices
96.7

Ratio of
nondurable
goods prices
to durable
goods prices

Year

Standard and
Poor’s Stock
Price Index1

Earnings/Price
ratio X 100

Ratio of
nondurable
goods prices
to stock
prices

.924

1961

90.2

96.6

1962

90.9

97.6

.924

1960

55.8

5.90

1.61

1963

92.0

97.9

.939

1961

66.2

4.62

1.36

1964

93.0

98.8

.941

1962

62.4

5.82

1.45

1965

94.6

98.4

.961

1963

69.9

5.50

1.31

.933

1966

98.1

98.5

.995

1964

81.4

5.32

1.14

1967

100.0

100.0

1.000

1965

88.2

5.59

1.07

1968

103.9

103.1

1.007

1966

85.3

6.63

1.15

1969

108.9

107.0

1.017

1967

92.0

5.73

1.08

1970

114.0

111.8

1.019

1968

98.7

5.67

1.05

1971

117.7

116.5

1.010

1969

97.8

6.08

1.11

1972

121.7

118.9

1.023

1970

83.2

6.46

1.37

1973

132.8

121.9

1.089

1971

98.3

5.41

1.19

1974

151.0

130.6

1.156

1972

109.2

5.50

1.11

1975

163.2

145.5

1.121

1973

107.4

7.12

1.23

1976

169.2

154.3

1.097

1974

82.8

11.60

1.82

1977

178.9

163.2

1.096

1975

86.2

9.12

1.89

1978

192.0

173.9

1.105

1976

102.0

8.90

1.66

1977

98.2

10.80

1.82

1978

96.0

12.05

2.00

SOURCE: Department of Labor, Bureau of Labor Statistics,
Consumer Price Index, Special Indexes.

the dramatic decline in the real rate suggested by the
estimates in table 1.
Furthermore, the estimates in table 1 do not appear
to be appropriately related to relative prices over
extended periods. If estimates generated by the stand­
ard method track the real rate, they should be posi­
tively correlated with the relative price ratios. This is
not the case, however, between 1960 and 1975. The
correlation between Elliot’s estimates and the ratio of
nondurable prices to durable prices is -.625. Between
his estimates and the ratio of spot and futures prices,
the correlation is -.484. The corresponding coefficients
for Carlson’s estimates are -.459 and -.073. Those for
the St. Louis Fed are -.692 (significant at the 5 per­
cent level) and -.121.
None of these estimates of the ex ante real rate of
interest generated by the standard method moved in
the direction implied by movements in these relative



1981

'Standard and Poor’s Statistical Service, Security Price Index
Record, Standard and Poor’s Corporation, New York, N.Y.

prices during the 1969-1975 period. The correlations
suggest that the effect 9f . 90 described in equation
90 9r
12 may be sufficiently large to make dr negative.
dr
3.
The Stock Market: The stock market provides
further evidence on this issue. Because stock prices
represent the present value of expected future earn­
ings, a decrease in the ex ante real rate of interest
will be reflected by a rise in the price of shares rela­
tive to current earnings and a fall in the earnings to
price ratio. During the period 1960-1972, earnings to
price ratios averaged 5.709 (table 4) with a standard
deviation of .511. In 1974 and 1975, earnings to price
ratios reached levels of 11.60 and 9.12, respectively.
25

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JAN U ARY

1981

In addition, a decrease in the rate of interest will
he reflected by a fall in the price of nondurable pres­
ent consumption goods relative to stock prices. Be­
tween 1960 and 1972 the ratio of the Index of Non­
durable Good Prices to the Standard and Poor’s Stock
Price Index averaged 1.234, with a standard deviation
of .177. In 1974 and 1975 it rose to 1.82 and 1.89,
respectively. Again, this relative price behavior is
clearly inconsistent with the contention that the ex
ante real rate of interest fell in 1974 and 1975.

because these price indices give substantial weight
to the prices of current consumption goods, as op­
posed to the prices of assets productive of future
consumption (capital goods), but also because they
reflect the impact of once-and-for-all changes in prices
produced by changes in the real interest rate. There­
fore, it is impossible when using this estimation pro­
cedure to separate changes in the real interest rate
from changes in the rate of inflation. As a result, the
method produces biased estimates of changes in the
ex ante real rate of interest.

CONCLUSIONS

Furthermore, the direction of this error is predict­
able. In particular, when the real rate of interest rises,
as in 1974 and 1975, the current method of estimation
will understate the change in the real rate. Evidence
from the mid-1970s suggests that estimates of the
real rate based on the CPI failed to detect the direc­
tion of change in the real rate.

The method currently used to estimate the ex ante
real rate of interest can lead to serious error. The
error arises because this method requires the in­
vestigator to measure the expectations of market
participants regarding the future rate of inflation.
Unfortunately, since these expectations are never
directly observed, the accuracy of the measurement
is questionable.
Price expectations have typically been approximated
by observing past rates of change in either the CPI
or the GNP deflator. This method of approximation
assumes, first, that expectations about the future rate
of inflation depend largely on the past rate of inflation
and, second, that the past rate of inflation is accu­
rately reflected by the past rate of change in these
price indices. This article has put aside the first issue
and argues that past rates of change in the CPI and
the GNP deflator may not accurately reflect the past
rate of inflation.
We have shown that real interest rate changes
themselves affect these indices. This occurs not only


26


Because estimates of the real rate employing mea­
sures of anticipated inflation based on common price
indices are suspect unless real rates are unchanging,
their value is severely limited for use in formulating
economic policy. Estimates of the ex ante real rate
of interest are important to policymakers if they aid
in distinguishing shifts in relative demands from
shifts in aggregate demand (i.e., are able to actually
detect changes in the real interest rate). However, the
widely employed method of estimation breaks down
precisely during periods in which the ex ante real in­
terest rate changes. Consequently, estimated changes
in the ex ante real rate of interest should be checked
against the behavior of the relative prices known to
depend upon the real rate prior to employing these
estimates for economic policy purposes.

Outlook for Food and Agriculture in 1981
NEIL A. STEVENS

RODUCTION of a number of major food prod­
ucts in the United States is predicted to decline
slightly from 1980 levels. Continued increases in de­
mand for food, and thus faster increases in food prices,
are in prospect. Also, a sizable increase in net farm
income is expected. These were among the conclu­
sions presented by U.S. Department of Agriculture
(USDA) analysts at the 1981 Food and Agricultural
Outlook Conference in Washington, D.C., last No­
vember and are summarized in this article.
Factors Underlying the 1981 Forecasts
Food price developments result from the interac­
tion of demand and supply forces. The major factors
affecting the demand for food include per capita real
income, population growth and the general rate of
inflation. In 1980 economic growth slowed or stopped
in most countries. At the outlook conference, USDA
analysts expected only slight economic growth in the
United States during the first half of this year, with
somewhat more rapid growth in the second half. More
recent developments, however, suggest that economic
growth in the first half of the year may be more
sluggish than USDA analysts expected. Although real
income growth will not be a major factor increasing
the demand for food, world population growth will
continue to increase and will contribute to an increase
in export demand for U.S.-grown foodstuffs.
Supply factors dominated the outlook for food prices
and farm income in 1981. The bulk of food output in
the first half of 1981 will be derived largely from
livestock production already under way and 1980 crop
production.
World production of grains, which includes wheat,
rice, and feed grains, in 1980 was roughly equal to
that in 1979. However, world stocks at the beginning
of the year were lower so that available supplies in
the 1980/81 marketing year are down about 2 per­
cent. Among grains, however, supplies of food grains
(wheat and rice) have increased relative to feed
grains (corn, sorghum, oats and barley). World food
grain production in 1980/81 rose 3.5 percent over
1979/80; feed grain production fell about 3 percent,
largely as a result of a severe U.S. drought.



Reduced feed grain supplies will affect U.S. retail
food prices primarily through lower livestock produc­
tion and higher prices of livestock products. Total
meat output in 1981, including beef, veal, mutton and
poultry, is expected to decline 1 to 3 percent below
the record levels in 1980. In contrast, total meat out­
put rose about 3 percent last year.
Most of the decline in meat output will reflect re­
duced pork production. USDA analysts expected hog
producers to reduce the June-November pig crop by
about 10 percent as a result of feeding losses in the
spring and summer of 1980. Consequently, pork pro­
duction in the first half of 1981 was expected to
drop about 11 percent from a year earlier. Recent
information, however, indicates that the June-November pig crop is down only 5 percent from the year
before; as a result, pork supplies may decline only 6
percent in the first half of 1981. Production in the
second half is more uncertain, but pork supplies are
anticipated to be 5 to 10 percent below levels of a
year earlier.
In contrast, production of beef, broilers and tur­
keys is expected to increase somewhat. Reef supplies
will be relatively large in the first quarter of 1981
due to increased placements in feedlots during last
summer’s drought. However, production in the sec­
ond quarter is expected to fall below that of a
year earlier. Broiler output in the first half of 1981
is projected to be up slightly from 1980. Expansion
in the second half of the year may increase output
by 3 percent above the 1980 level. Turkey produc­
tion, given relatively high prices, may increase about
6 percent. Egg output in 1981 is expected to be about
1 percent less than in 1980. Most of this decline is
anticipated in the first quarter; production in the rest
of the year will be about the same as last year. Milk
output, on the other hand, is expected to rise 1 to 3
percent.
Crop-related food supplies, as a whole, are ex­
pected to expand slightly in 1981. The supply of
cereal crops provides a substantial base for the pro­
duction of cereal and bakery goods. However, the
prices of some ingredients — in particular, oils and
sugar — are predicted to rise. Reduced production of
27

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JAN U AR Y

orange crop, is expected to be down about 20 percent.
Large stocks of frozen orange juice, however, will
tend to moderate the price impact of the freeze.

Table 1

Retail Food Price Changes
(from previous year)
Food category

1978

1979

All food

10.0%

10.9%

Food away from home
Food at home
Meats

1980
8.7%

19811
12.2%

9.0

11.2

10.0

10.4

10.5

10.8

8.1

13.0
17.9

18.7

17.0

3.5

Beef and veal

22.9

27.3

6.4

13.5

Pork

12.9

1.5

-2 .6

27.6

Other meats

17.8

14.7

4.1

17.5

10.3

5.0

4.1

18.0

9.5

9.8

9.2

9.6

-5.5

9.5

-3.1

16.9

6.7

11.6

10.2

10.7
11.0

Poultry
Fish and seafood
Eggs
Dairy products
Fats and oils

9.5

8.0

6.7

Fruits and vegetables

11.1

8.0

7.0

8.0

Sugar and sweets

12.2

7.8

22.4

21.5

Cereals and bakery
products

8.9

10.1

11.9

10.9

Nonalcoholic beverages

5.7

5.0

10.8

12.0

Other prepared foods

8.0

10.1

10.9

10.3

'U SD A forecast
SOURCE: Paul C. Westcott, “ 1981 Food Price Outlook”
(Presented at the 1981 Agriculture Outlook
Conference, Washington, D.C., November 19,
1980), p. 10.

oilseeds will cause the prices of fats and oils to rise
by 11 percent. World sugar production in 1980/81 is
slightly above the reduced 1979/80 crop, but begin­
ning stocks are estimated to be down for the second
consecutive year. Raw sugar prices in late 1980 were
up 67 percent from the 1979 average. At the retail
level, the price of sugar and other sweeteners advanced
22 percent in 1980; a similar advance is expected in
1981. A significant development in the sweetener mar­
ket is the sharp increase in the use of corn-derived
sweeteners. These have grown from about 16 percent
of the market in 1970 to about 33 percent in 1980 and
may reach nearly 50 percent of the market for nutri­
tive sweeteners by 1985.
Supplies of fruits, including fresh apples and most
canned and frozen fruits, are greater than a year ago.
Until the January freeze in Florida, a record citrus
crop was expected. That crop has now been reduced
significantly by the freeze. Florida orange production,
which accounts for about 75 percent of the total
Digitized for
28FRASER


1981

Supplies of processed vegetables, both canned and
frozen, are down about 6 percent in 1980/81, reflect­
ing a planned cutback in production. The fall 1980
potato crop is down about 12 percent, and the fresh
winter vegetable crop was reduced by the recent
freeze in Florida. Hence, supplies of some fresh vege­
tables will be reduced until replanted crops come to
market.
1981 Food Price Increases —
Higher than General Inflation
Given these demand and supply developments, food
prices on a yearly average basis are projected to rise
about 12.5 percent from 1980 to 1981. This increase
is expected to be somewhat greater than the antici­
pated rate of inflation as measured by the consumer
price index (CPI). In contrast, food prices in 1980
rose 8.7 percent, while the CPI increased 13 percent.
Developments in the first half of 1980, however,
were quite different compared to the second half of
the year. In the first six months, farm prices were
below the previous year, reflecting large grain sup­
plies from the 1979 harvest and record meat output.
Farm prices began to rise sharply in the second half
of 1980 as increases in livestock production slowed
and as the effects of the summer drought on crop
production led to higher grain prices. As a result,
retail food prices rose substantially in the second half
of the year.
The projected year-to-year price changes for vari­
ous food groups in 1981 are shown in table 1. Meats,
poultry, eggs and sugar are the main food groups
with the greatest projected price increases in 1981.
Large increases for these groups reflected the expected
decline in overall meat output. This decline and,
hence, the upward pressure on the prices of animal
protein foods may not be as great as USDA analysts
expected, since indications are that pork production
will not decline as much as anticipated. On the other
hand, the increase in fruit and vegetable prices is now
likely to be greater than projected due to the freeze
damage in Florida.
Farm Income to Recover Much of
Last Years Decline
Price increases of meats and other livestock prod­
ucts are expected to lead to a recovery of farm income

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

JAN U AR Y

Table 2

Farm Income (billions of dollars)
Cash receipts
Crop

1978

1979

19801

19811

$112.5

$131.5

$140

$158

53.5

62.8

71

77

Livestock

59.0

68.6

69

81

Other income

14.0

14.0

16

17

Total farm income 126.5

145.5

156

175

100.8

118.6

131

147

Net cash income

33.8

35.8

34

—

Net farm income
(before inventory
adjustment)

25.7

26.9

25

28

Net farm income
(after inventory
adjustment)

26.1

31.0

Production expenses

24
29.5
(23-25) (27-32)

HJSDA forecast
SOURCE: George H. Hoffman, “ Farm Income Situation
and Outlook” (Presented at the 1981 Agricul­
ture Outlook Conference, Washington, D.C., N o­
vember 19, 1980), pp. 6-10.

from the relatively low 1980 level. Overall, net farm
income of farm operators is forecast at about $30
billion, up 20 percent from 1980, and only slightly
below the 1979 level (table 2).
For the year 1980 as a whole, prices received by
farmers were only 2 percent above 1979, and total
cash receipts were up 6.5 percent. Total farm income
including cash receipts, other cash income, govern­
ment payments and imputed income on such items as
family dwellings, was up about 7 percent. Production
expenses, reflecting general inflation trends in the
economy, rose over 11 percent in 1980. Increases in
input prices were led by a 39 percent increase in fuel
and energy, a 23 percent rise in fertilizer and a 20
percent gain in short-term interest rates.
As a result of the faster rise in farm input prices
over farm commodity prices, 1980 net farm income
after inventory adjustment was about $24 billion,
down $7 billion from 1979. Cash flow, a measure of
the farmer’s ability to meet short-run obligations which
excludes imputed income and expenses, did not de­
cline so sharply. This measure totaled around $34 bil­
lion, down 5 percent from 1979. Much of the differ­
ence between net farm income and cash flow was due
to a reduction in inventories.



1981

In general, crop producers in 1980 fared better than
livestock producers. Crop receipts went up 13 per­
cent, while livestock cash receipts remained un­
changed. However, there were considerable income
differences among crop farmers. In some areas,
drought reduced crop yields significantly and incomes
were down sharply. Good yields and significantly
higher prices resulted in improved incomes in the
upper Midwest and eastern Com Belt. Among live­
stock producers, hog operators fared relatively poorly.
Dairy farmers, reflecting increased government price
supports, had generally profitable operations.
The projected increase in farm income in 1981 is
based largely on a 16 to 20 percent increase in live­
stock receipts. Crop receipts are expected to increase
by 6 to 10 percent.
Production expenses are expected to increase by
11.5 percent, about the same rate as in 1980. Farm
origin input costs, unlike last year, are expected to
rise sharply. Total feed expenses will rise by 15 per­
cent or more to ration the reduced supply among
competing uses. Expenses for purchased livestock are
expected to rise about 10 percent. Petroleum-based
inputs such as fuel, fertilizer and chemicals may regis­
ter significant gains. With a slower rate of overall
inflation expected, a moderation of price increases for
manufactured inputs is anticipated. Total interest
expense is not anticipated to rise significantly, even
though farm debt will be larger. Short-term interest
rates are expected to fall from 1980 levels.

OUTLOOK FOR MAJOR
FARM PRODUCTS
Feed Grains
U.S. production of feed grains in 1980 was about
198.3 million tons, 17 percent below the record harvest
in 1979. Planted acreage was 2.4 percent above 1979,
but a severe drought in some major producing areas
reduced yields to 1.95 metric tons per acre, down 16
percent from 1979’s record yields. With the relatively
large feed grain carryover from 1979/80, overall U.S.
supplies (production plus stocks) for the 1980/81
marketing year totaled 250.5 million tons, 12 percent
below 1979/80. Production of com, the major feed
grain, was 6.6 million bushels in 1980/81 compared
with 7.9 million bushels in 1979/80. Supplies (produc­
tion plus carryover) totaled 8.2 million bushels, down
from 9.2 million bushels in 1979/80.
With the quantity of feed grains essentially fixed
by the 1980 harvest plus 1979/80 carryover stocks,
29

JA N U A R Y

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

price changes in the next few months will result pri­
marily from changes in demand. Demand is expected
to remain strong, reflecting increased demand for
exports and domestic food. Domestic livestock feed­
ing, primarily hog feeding, however, will decline
somewhat from the previous year as average feed
grain prices are expected to be significantly above
those for last year. Com and sorghum prices are ex­
pected to average about $3.40 and $3.30 per bushel,
respectively, compared with $2.50 and $2.35 per
bushel, respectively, last year.
Total feed grain use may be only slightly below
that of the 1979/80 season, and carryover stocks will
be substantially reduced. U.S. stocks are expected to
fall from 52 million metric tons at the end of the
1979/80 marketing year to 21 million tons at the end
of the current year. The ratio of stocks to utilization
is expected to decline to 9.2 percent, the lowest since
1975.
While feed grain prices are expected to be higher
this year than last year, their impact on incentives to
increase production will be partially offset by a sizable
increase in prices paid by farmers for production
items. Nevertheless, higher feed grain prices, and
hence higher profits, will provide incentive for farmers
to increase feed grain acreage and production inputs
per acre. According to USDA projections, given nor­
mal weather conditions this year, com yields are likely
to rise to about 103 bushels per planted acre, up 14
percent from last year but still below the record 109
bushels per acre in 1979.
Food Grains
The supply of food grains, wheat and rice, is rela­
tively more abundant than feed grains, reflecting the
relatively large harvest in 1980. U.S. wheat produc­
tion in 1980 was a record 2.37 billion bushels, up 11
percent from a year earlier, and 1980/81 supplies
totaled a record 3.3 billion bushels. On a worldwide
basis, however, wheat production was smaller than
anticipated, which, coupled with reduced feed grain
crops, led to wheat price advances in the late summer
and fall. With projected use of U.S. wheat near the
production level, stocks at the end of the 1980/81
marketing year are expected to remain near last year’s
level of 900 million bushels. Wheat prices are expected
to follow a normal pattern of seasonal strength through
the remainder of 1980/81 and may average $4.05 per
bushel for the year, about 25 cents per bushel above
last year’s price. Fall plantings of winter wheat are
estimated to be up 11 percent from 1980 and, with
30



1981

normal yields, another large U.S. wheat crop is in
prospect for 1981.
U.S. rice production in 1980 was estimated at 145
million hundredweight (cw t.), 10 percent above the
year before. Yields were down from recent years, but
producers planted 16 percent more acres, reflecting the
relatively high profits expected from rice production.
While beginning stocks were down, total U.S. rice
supplies are up 4.4 percent and world rice sup­
plies are up about 3 percent. U.S. farm prices for
rice in the 1980/81 season are expected to average
about $11.50 per cwt., up about 10 percent from the
1979/80 average. Production costs are expected to
rise substantially, however, and a reduction in rice
acreage is likely in 1981.
Soybeans
Production of soybeans, which constitutes about 88
percent of U.S. oilseed production, declined 20 per­
cent in 1980. The impact of this unexpected shortfall
on prices and consumption, however, was blunted by
a large inventory and an increase in oilseed produc­
tion elsewhere in the world. World oilseed supplies
are down only about 3 percent. World consumption
is expected to continue expanding, and ending stocks
will be down substantially from the 1979/80 level. The
year-end world stock-to-use ratio is expected to be
around 9.4 percent, still above most recent years. U.S.
soybean supplies, however, are relatively low, and
prices in the 1980/81 season may average $7.90 per
bushel, up from $6.25 a bushel in the previous season.

While soybean prices are likely to be substantially
higher than last year, the soybean/corn price ratio
does not provide farmers with the incentive to shift
from com to soybean production. Thus, acreage is
not expected to change much from last year’s level.
However, a return to normal soybean yields in 1981
would result in a sharp recovery in U.S. soybean
production.
Cotton
U.S. cotton production of 11.1 million bales in 1980
was down 24 percent from the relatively large 1979/
80 crop. World production was also down, largely a
result of the decline in the U.S. crop. With demand
for cotton relatively strong, prices in late 1980 aver­
aged 34 percent above the previous October.
Despite relatively low supplies and high prices, cot­
ton acreage may decline by a half million acres or
more in 1981. Prices of soybeans and grain sorghum

F E D E R A L R E S E R V E B A N K O F ST. LOUIS

have increased relative to cotton, providing an incen­
tive to shift acres from cotton to these crops. In addi­
tion, the costs of producing cotton, in absolute terms,
have increased more than some competing crops. As­
suming normal yields of about 1 bale per acre, sup­
plies of cotton are expected to remain tight through­
out 1981 and into 1982.
Tobacco
U.S. tobacco production in 1980 recovered from the
relatively small crop in 1979, but because of the hot
and dry growing conditions, the quality of some
tobacco is low. Tobacco production rose 17 percent
from the very small 1979 crop as a result of a 12 per­
cent increase in acreage and a 4 percent rise in yields.
With much lower carryover stocks, however, total
supplies for the 1980/81 marketing year are down
about 2 percent. Tobacco production is heavily in­
fluenced by government price supports and marketing
quotas, and the current law mandates a 12 percent
rise in price supports for eligible tobacco.
Beef Cattle
The liquidation phase of the cattle cycle ended in
1980. The number of cattle and calves on farms as of
July 1, 1980, indicated a rapid rebuilding, with cattle
numbers up 4 percent from 1979. The 1980 calf crop
of 45.5 million head was up 6 percent from 1979.
Several factors, however, may increase costs of produc­
tion, thereby limiting future beef herd expansion.
These include a substantial increase in land converted
from pasture into cropland, and higher energy costs,
which limit fertilization of pastures.
Higher catde prices are in prospect for 1981, par­
ticularly in the second quarter, as total meat supplies
are expected to fall below levels of a year ago. Cattle
feedlot operators increased placements during the
summer as drought led to larger marketings of feeder
cattle. These cattle will come onto the slaughter mar­
ket in the first quarter and will moderate increases in
prices. Choice steer prices are expected to average
around $73 per cwt. in the first quarter.
Although cattle marketings for slaughter will rise
somewhat in the second quarter, the slaughter of non­
fed beef should fall below the 1980 level if grazing
conditions return to normal. As a result, overall beef
production will likely fall and the price of choice
steers will rise. Despite increased feeding costs, profit
margins are expected to increase in the second quar­
ter. However, cattle prices are not expected to in­
crease much further in the second half of the year,



JA N U A R Y

1981

and feeding margins may be reduced or even become
negative. The profitability of feeding operations in
the second half will depend on feed costs and there­
fore on the outlook for 1981 grain crops.
Hogs
Hog producers experienced large losses in the first
half of 1980 as large meat supplies led to prices below
$30 per cwt. in April and May. Producers reacted
by slaughtering more breeding stock and cutting back
on breeding inventory. At the outlook conference, the
June-November pig crop was anticipated to decline
about 10 percent; more recent information, however,
indicates about a 5 percent decline.
Lower production in the first half of 1981 will re­
sult in higher hog prices and upward price pressure
on all animal products. With a 10 percent decline in
pork production, hog prices had been expected to
average around $50 per cwt. in the first half of 1981,
nearly $16 above the depressed levels of 1980. But
with pork production likely to fall only about 6 per­
cent in the first half, prices are not likely to reach
profitable levels. This would indicate more cutbacks
in pork production in the second half of 1981.
Poultry and Eggs
After suffering losses in the first half of 1980, broiler
producers planned to reduce production in the sec­
ond half of 1980. This, coupled with an unusually
hot summer that caused a substantial unplanned re­
duction, resulted in higher prices.
Reduced breeding flocks, the result of last summer’s
hot weather, and higher production costs are expected
to limit production increases to around 3 percent
above 1980. As meat supplies decline, broiler prices
are expected to rise in 1981. Wholesale prices may
average around 52 cents per pound in the first quar­
ter, increasing to around 55 cents in the second quar­
ter and 56 cents in the second half.
Since turkey production has generally been profit­
able since 1977, producers have sharply increased
output. Increasing year-round consumption of tur­
keys has meant a substantial increase in demand. De­
spite higher feed costs, producers have increased the
number of poults hatched for slaughter purposes in
recent months, and output may increase around 7
percent to 8 percent in the first half of 1981. Prices
may average 67 cents to 70 cents per pound in the
first half of 1981, compared with 57 cents in the first
half of last year.
31

Egg production was not profitable in 1980. During
the first half of the year, prices were low due to a
weak economy and large supplies of competing pro­
tein foods. In the second half, rising costs largely off­
set price increases. As a result, producers have cut
back egg production and output is expected to be
down about 1 percent from 1980. Most of the reduc­
tion will occur in the first quarter, which should
cause egg prices to rise in the first half of the year.
Milk
Milk production has expanded since 1979 as favor­
able prices to producers have prevailed, largely due
to government price supports. Production last year
was about 3 percent larger than in 1979. Milk prices
are expected to rise this year, but higher feed prices
will reduce producers’ profits to levels below those of
the past couple of years. While larger dairy herds
should result in higher milk production, higher feed
costs will slow total output per cow, so that produc­
tion will likely rise by about 2 percent.
The government support price of manufacturing
milk for the marketing year beginning October 1 was
set at the minimum required level of 80 percent of



parity — $12.80 per cwt. This will be adjusted again
on April 1 to reflect changes in the index of prices
paid by all farmers. In price support operations,
government purchases of milk in the first nine months
of 1980 totaled 7.35 billion pounds, almost 8 per­
cent of all milk marketed, compared with 1.31
billion pounds in 1979. Commercial use of milk and
dairy products was down 1.6 percent in 1980, but an
increase may occur in 1981. Despite this increase,
USDA purchases of dairy products in price support
operations are expected to continue if the gains in
milk production occur.

CONCLUSION
Retail food prices in 1981 are expected to increase
12.5 percent (range from 10 to 15 percent). General
inflation underlies much of the increase, though food
prices may rise somewhat faster than overall prices.
This reflects such adverse supply factors as reduced
feed grain supplies resulting from last summer’s
drought, reactions of hog producers to unfavorable
profit opportunities and reduced sugar supplies. As
a result, substantially higher livestock and sugar
prices will contribute to higher retail food prices and
substantially higher net profits of farm operators.