Schor, J. B. (1985). Changes in the Cyclical Pattern of Real Wages Evidence From Nine Countries,...

8/12/2019 Schor, J. B. (1985). Changes in the Cyclical Pattern of Real Wages Evidence From Nine Countries, 1955-80. the Ec

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Changes in the Cyclical Pattern of Real Wages: Evidence From Nine Countries, 1955-80Author(s): Juliet B. SchorSource: The Economic Journal, Vol. 95, No. 378 (Jun., 1985), pp. 452-468Published by: Wiley on behalf of the Royal Economic SocietyStable URL: http://www.jstor.org/stable/2233220 .Accessed: 11/12/2013 05:22

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The Economic ournal, 95 (June 1985), 452-468

Printed n Great Britain

CHANGES IN THE CYCLICAL PATTERN

OF REAL WAGES: EVIDENCE FROMNINE COUNTRIES, I955-80*

Juliet B. Schor

In the early 98os, the United States xperienced prolonged eriod ofrestrictivemonetary policy. In the United Kingdom, he monetarist ice has gripped heeconomy even longer, and similar deflationary olicies have been in effect nother Western conomies. nterestingly nough, hese policies were nstituted nthe context of a prolonged eriod of economic decline. Annual average ates ofgrowth of Gross Domestic Product or nine OECD countries' ell from 6-8 to5 o to 2 6 over the periods I955-66, i966-73, and I973-79.2 Similarly,unemployment as risen dramatically n the last io years.

The causes f slow growth and high unemployment re undoubtedly arious;however, it is clear that restrictive macroeconomic policy has played animportant ole. That is, the recent situation s at least in part attributable o adeliberate effort to induce an economic slump. The task of systematicallyexamining he origins and motivations f this policy still awaits us. However,this paper presents n account of a general development n the pattern of realwages over the business ycle which s a plausible determinant f current policy.The evidence ndicates hat short-run yclical movements n output no longerexert a determinate nfluence on real wage growth, a development which mayhave been the impetus or long-term deflation.

This account s based on an econometric tudy of nine countries over theperiod I955-80. The study has three noteworthy econometric eatures. Thefirst s that data from the nine countries re combined n a pooled model with

homogeneous oefficients. he use of a cross-national ample nsures gainst heattribution f general heoretical tatus o particular mpirical indings rom asingle country.3 Second, the units of observation re business cycle turningpoints, rather than chronological units. Third, changes n acceleration anddeceleration f wages are investigated, ather han rates of change hemselves.

* This paper is taken from the author's Ph.D. dissertation. She would especially like to thankSam Bowles. She would also like to thank John Sheahan, Lawrence Kahn, Morton 0. Schapiro,Gerald Epstein, Michele Naples, an anonymous referee, participants in the Williams Conference onthe Political Economy of Inflation and Unemployment and the Williams Faculty Seminar. HowardShapiro provided research assistance. Generous financial support was received from the BrookingsInstitution.

1 These countries are: Canada, France, Germany, Italy, Japan, Netherlands, Sweden, UnitedKingdom and United States.

2 These dates correspond roughly to business cycle turning points which are common to a majorityof the nine countries. The year 1979 is the latest for which data are available.

8 For instance, macroeconomists in the United States have generally failed to recognise the degreeto which real wages move procyclically. I suspect this is due to the fact that wages in the United Stateshave traditionally exhibited only weak procyclical movement relative to Western Europe.

[ 452 ]

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[JUNE I985] CHANGES IN THE CYCLICAL PATTERN OF REAL WAGES 453

I. REAL WAGES IN KEYNESIAN AND NEOCLASSICAL MODELS

Among Keynesian and neoclassical economists there is general agreement thatmoney wages exhibit a procyclical pattern. For real wages, there is theoretical

dissensus, and the econometric evidence is sparse.The debate on the issue dates from the I930s. Immediately following thepublication of the General Theory he cyclical pattern of real wages was discussedin the pages of this JOURNAL. Keynes had made the standard orthodoxargument that diminishing marginal returns to labour resulted in an inverserelation between real wages and employment.

. . . The change in real wages associated with a change in money wages,so far from being usually in the same direction, is almost always in the

opposite direction... This is because, in the short period, falling moneywages and rising real wages are each, for independent reasons, likely toaccompany decreasing employment; labour being readier to accept wage-cuts when employment is falling off, yet real wages inevitably rising in thesame circumstances on account of the increasing marginal return to a givencapital equipment when output is diminished. (p. IO)Dunlop (I938) and Tarshis (I939) challenged Keynes' position, concluding

that the evidence showed a tendency for real wages and output to move together.Keynes responded agnostically, agreeing that they had 'seriously shaken theassumptions on which the short-period theory of distribution s based', but that'we should not be too hasty in our revisions' (1 939).

This debate was abandoned for 30 years, until Bodkin (I969) presentedevidence supporting the view that real wages and employment are positivelyrelated. However, Bodkin's findings are consistent with various Keynesianassumptions about wage and price responsiveness. For instance, if prices respondto wages with a considerable ag, real wages will exhibit procyclical movement,at least for some portion of each business cycle phase. Similarly, if nominal

variables are subject to substantial inertia, as Keynesians have claimed (Perry,I980; Okun, I98I) then cases in which price inertia dominates wage inertia willresult in procyclical movement. In the United States, the structure of labourmarket institutions (long-term employment, multi-year contracting, anddecentralised bargaining) and the postwar data have led observers to the viewthat wages are more prone to inertia than prices. However, this may not neces-sarily be the case in some Western European countries which have differentlabour market institutions and significantly more open economies. For instance,in an Aukrust-type model (I977), prices are likely to respond less than propor-

tionately to money wage increases. Thus, while the thrust of most Keynesianmodels is to posit countercyclical movement of real wages, this prediction is notstrictly necessary.

Bodkin's evidence was challenged by Canzoneri (I978) and Neftci (I978).Neftci used a neoclassical model, but was agnostic about its predictions. Forinstance, ' In a market such as the one for labor, the supply and demand curveswill shift over the business cycle and generate a sequence of observations on real



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5/18

I985] CHANGES IN THE CYCLICAL PATTERN OF REAL WAGES 455

described by Marx. They can be summarised in the following two-equationsystem:

zbt = f(ut) (T < ?)) (I)

Ut = 9(fbi-n) (V > ?) (2)

where w is the real wage, and u is the unemployment rate. Much of the currentwork in this area has been directed toward the first equation, namely the pro-position that real wage change depends on the level of unemployment.' Thereasoning s that economic expansion eventually results n increased strength forlabour relative to capital, which yields a rise in real wages. This may occur forvarious reasons. First, the growth of labour supply may not keep pace with thegrowth of labour demand because population adjusts slowly. Second, workers'willingness to resist firms' wage offers and production strategies may increase asstrike funds grow and competition for jobs diminishes. Similarly, firms may bemore willing and able to grant concessions if demand and profits are high.Finally, shortages of particular types of skilled labour may develop. To theextent that customary wage differentials are maintained, wages of all workerswill rise. An analogous argument s applicable to periods of slow capital accumu-lation during which excess labour supplies can be expected to exert downwardpressure on wage rates, thereby creating the conditions for renewed profitability.

These arguments are generally cast in macroeconomic terms, however, for,amodel in which microfoundations are developed, see Bowles (I985). In that

model, the reserve army mechanism operates through the conflict betweenworkers and firms over the intensity of work.2The foregoing analysis has been set in real wages with the implicit pre-

sumption that money wages are driven by real wages. Marx did not develop atheory of price movements over the cycle under conditions of imperfect competi-tion. However, in models which permit price-making behaviour by firms, it isnot obvious that money and real wages will move together, as Keynes pointedout in I 939. For instance, Ka]ecki (I 97 I) argued the converse ' . . . and even therise in wage rates resulting rom the strong bargaining power of the workers s less

likely to reduce profits than to increase prices, and thus affects adversely only therentier interests' (p. I4V). Rowthorn (I977) takes the view that competitors'capacity constraints ead firms to raise prices freely during expansions.

On the other hand, various observers have argued that price increases cannotfully offset increases n money wages because of the discipline of a world market,where prices are set competitively (Glyn and Sutcliffe, I 972; Crotty and Rapping,I976). However, Epstein (I982) suggests hat the switch from a fixed to a floatingexchange rate regime may have undermined this discipline.

Marxian theory, like neoclassical and Keynesian, is thus theoretically in-conclusive on the issue of the cyclical pattern of real wages. Marx's own analysis

1 The description of reserve army theory is taken from the following: Boddy and Crotty, I975;Kalecki, 197I; Glyn and Sutcliffe, 1972; Desai, 1975; Rowthorn, I977; and Kahn, I980.

2 In Marxian theory, this issue is dealt with through the distinction between labour and labourpower. Labour is defined as the production input and labour power as the ability to work, the object ofmarket trading. Under conditions of alienated labour, or even weaker assumptions, the extraction oflabour from labour power is a problematic process for firms. For a fuller exposition of this problem, seeMarglin (1974), Gintis (1976) or Bowles (I985).



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456 THE ECONOMIC JOURNAL [JUNEand early Marxian scholarship stressed the importance of a real wage-inducedprofit squeeze. However, these treatments failed to account adequately for priceformation under conditions of less than perfect competition. Given the lack ofdefinitive empirical evidence on the question, the theoretical discussion is

unlikely to be clarified and advanced until strong econometric evidence can befound and accepted. For we are not yet clear how the theory stands vis-a-visthe stylised facts. I will therefore proceed to the empirical investigation.

III. EVIDENCE ON THE RESERVE ARMY EFFECT

In order to investigate the cyclical pattern of wages I have paid explicit attentionto business cycle turning points, by constructing a series of theoretical, ratherthan chronological, observations.' These observations are cyclical peaks and

troughs, rather than conventional annual or quarterly observations. It is a prioripreferable to use theoretical observations, as it is a relationship between theo-retically-defined units which we are attempting to understand n this study. Thispoint may be clarified by considering the fact that because the length of cyclicalupturns and downturns is generally variable, the use of chronological obser-vations has the effect of weighting the data, whereas the use of theoreticalobservations accords equal weight to each upturn or downturn. This method alsoresults in a smaller measurement noise to true signal ratio. A similar procedurehas been used by Schultze (I98I) and of course by Phillips in his original article.

There are two major disadvantages to this time-sequence method. The first isa reduction in the number of available observations. I have addressed thisproblem by pooling time-sequence data for nine countries. The second is thepossibility that the analysis will be sensitive to the particular peak and troughpoints chosen. The selection of turning points may be a matter of controversy,insofar as the behaviour of a number of distinct indicators is considered. Schor(I982) contains a detailed discussion of the choice of peak and trough years, aswell as a re-estimation with alternative peak and trough datings of the basic

equations shown here. The results presented here are robust with respect to thealternative datings.A second point of method is the use of changes in the rates of change, rather

than the rates of change themselves. This method addresses recent criticisms ofthe use of inflation and unemployment levels. (See Lucas, I98I and Taylor,1980.)

As a first step I have calculated a measure of acceleration and deceleration ofthe rate of change of wages around business cycle turning points, corrected or theseverity of the business cycle, and labelled it the relative cyclical effect. Positive

values of the relative cyclical effect reflect procyclical movement of wages,negative values reflect countercyclical movement.

The relative cyclical effect is presented n Table I.2 Peak and trough years are

1 For a discussion of the difference between theoretical and chronological, or historical time, seeAlthusser (1970).

2 Because the number of cycles varies slightly by country, cycle numbers were allocated so that theywould correspond to roughly the same time periods.



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Tablei

Relative

Cyclical

Effect*

Real

wages

A-

5

All-country

Canada

France

Germany

Italy

Japan

Neth.

Sweden

U.K.

U.S.

cycle

mean

CycleI

0217

2-66

0463

1154

?-154

-

o678

0O331

0I361

o0752

CycleII

o04I8

-o0372

I789

I936

3-836

I538

-2-70I

0?445

OI75

o785

Cycle

III

-o208

-2*299

0-787

I760

o654

0-209

-0-42I

I562

0255

0-256

Cycle

IV

-o

263

I

I55I

0380

-01I39

I*3I2

-I

.004

-0o7I4

OI84

o0i63

CycleV

-o0780

-II83

-o-260

-I426

-o-o6i

0O442

-4

633

-0'27I

-1020

Average,

o0I42

-O-004

.OI

I*62

I.55

0874

-o8I5

O0779

o264

o6oo

cycles

I-III

Average,

-o*I23

-o0299

o*866

o076i

o889

o875

-o862

-o-602

o0I4I

O1185

all

cycles

*

The

relative

cyclical

effectis

the

sumof

the

peakto

trough

and

the

troughto

peak

changein

the

annual

rateof

changeof

real

wages,

dividedby

the

sumof

the

peakto

trough

and

troughto

peak

changein

the

output

gap.

(zbt-

dtj)+

(Wtj~-

d8+1)

-

+

(q~~-

where

j=cycle

number

(q,j-

qtj)+

(qtj-

qj+,,)

p=

peak

year

t=

trough

year.

wi,

Annual

rateof

changeof

hourly

compensationin

manufacturing

deflated

by

the

consumer

price

index.

q,

Actual

minus

potential

outputin

manufacturing.

All

annual

ratesof

change

are

calculated

as

ioo

(log

xt-

log

xt-i).



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458 THE ECONOMIC JOURNAL [JUXElisted in Appendix I, and data sources are listed in Appendix I.1 At this pointI will discuss only the data for the years 1955-70, because of the hypothesisedchange in the pattern of cyclical variability. I will first present evidence that realwages moved procyclically in the period 1955-70, and in the next section, show

the alteration of this pattern.The relative cyclical effect calculations reveal the following: First, all-countryaverages show strong procyclical movement for the first three cycles. In thethird cycle, the relative cyclical effect is considerably smaller, although nearlyhalf the decline is accounted for by the large countercyclical pattern in France,excluding which the cyclical average is 0575. Second, seven of the nine countriestypically exhibit procyclical movement in real wages. The two exceptions areFrance and Sweden.2

Table 2

Relative Cyclical Effect, Expansions nd Contractions

Contractions Expansions

Cycle I -o-267 0-486Cycle II -0 534 0-251Cycle III 0o020 0-276Cycle IV -o0020 0?143Cycle V 0o025 -I-OOI

Period averagesCycles I-III -o-26o o.338Cycles IV-V o-oo3 - 0429Cycles I-V -0_155 0?031

The relative cyclical effect is highly aggregated. An important form ofdisaggregation is between business cycle expansions and contractions. Thesecalculations are shown in Table 2, averaged for all countries. They conform to

the expected pattern, that is, the rates of change of real wages rise in expansionsand fall in contractions.

The foregoing calculations have been supplemented by econometric estimatesfor each of the nine countries individually and for a pooled model. Pooling thedata is desirable for two reasons. First, the macroeconomic theory which under-lies the model is a priori applicable to all advanced capitalist economies. Poolingprovides a test of whether the reserve army effect is sufficiently operative to

1 Manufacturing, rather than aggregate economy data have been used because they are consider-ably more reliable and consistent across countries than aggregate economy data. Also, output gaps areonly available for manufacturing. Output gaps have been used because unemployment rates arewidely recognised to be poor indicators of cyclical conditions in Western Europe.

2 Previous treatments have identified France and Sweden as atypical. Boyer (1979) argues thatlabour market developments have ceased to play a significant role in wage determination in France.This view coincides with the findings of Perry (I975), and Sachs (I979). Similarly, Swedish wagedevelopments have been analysed in idiosyncratic terms. For instance, if Swedish workers face a weakcapitalist class, a monopoly pricing model may be appropriate. This analysis should be augmented byan Aukrust model incorporating the degree of openness in the economy (I977).



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i985] CHANGES IN THE CYCLICAL PATTERN OF REAL WAGES 459

override national institutional or other differences. Second, pooling increasesthe available degrees of freedom.'

The model used is derived from a standard wage equation, such as equation (i)above, in which wage change depends on the level of unemployment. Becausethis study is investigating

acceleration and deceleration in rates of change,second differences were taken from the standard form. This yields the followingmodel:

w =o+b,4+el (3)

where w is the real wage, defined as the money wage deflated by the contem-poraneous vaJue of the consumer price index, q is the output gap and el is astochastic error term, assumed to have standard properties. A double dot overa variable indicates a second difference, a single dot a first difference. Thesevariables are defined as x =

(xi+,-

xi)where i goes from I to n and n is the number

of peak to trough and trough to peak observations. Verbally, w represents peakto trough (or trough to peak) changes in the rate of change of real wages at thepeak (trough). All estimates also include a constant term, for reasons discussedbelow. For a detailed discussion of the choice of data used to estimate theseequations consult Schor (I982).

In addition to estimating a real wage equation, it may be of interest to considerproduct wages, i.e., money wages deflated by a value-added index. Cross-nationally consistent value-added deflators are available beginning in I96I,

therefore the sample period for the product wage equation is slightly shorter.These estimates are included with the real wage equations.Equation (3) was estimated for the period I955-70 in order to discern the

pattern of wages before the hypothesised decline in cyclical variability took place.The results of the estimation of equation (3) are consistent with the evidenceprovided by the relative cyclical effect measures. In the real wage equation, theoutput gap is significant and positive, indicating procyclical movement, andconsistency with a reserve army hypothesis, as well as certain Keynesian and neo-classical models. In the product wage equation, the results are essentially thesame. These results are presented in Table 3.2

Results from individual country equations are not shown here, but can beseen in Schor (I982). In those estimates the output gap variable is significant atthe IO level or above in all countries but France and Sweden. The usefulness

1 In this model both the intercepts and the slope coefficients are assumed to be equal. Before poolingit is necessary to test whether this assumption is warranted, by means of an F test for which the nullhypothesis is the assumption that there are no significant differences in the coefficients and intercepts.Equation (3) was estimated. The resulting F-ratio was F(3 5') 0-745 which is not significant atconventional confidence levels, thus the null hypothesis is not rejected and the decision was made topool the data. For a discussion of this F-statistic, see Maddala (I977). When either slope or interceptdummies for individual countries were added, they were not statistically significant, nor did they alterthe basic results.

2 The reader may notice the absence of the customary Durbin-Watson statistic for auto-correlationof residuals. This is because the D.W. statistic was developed for time-series, not time-sequence data.In Schor (I982) I discuss the question of auto-correlation in these estimates and present estimateswhich are corrected for auto-correlation. The results shown here are robust with respect to this cor-rection.



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460 THE ECONOMIC JOURNAL [JUNE

Table 3Wage Equations, Early Period

Number of

Dependent variable Constant observations R2 S.E.

(i) Real wages 0o284 01202 46 0o336 3-09I(0955-70) (4*872) (0o439)

(2) Product wages 0o434 -037I 3I 0o205 5-628

(i962-70) (2*953) (-0o367)

Figures in parentheses are t-statistics. Dependent variables are change in rate of change of real andproduct wages. All observations are calculated at business cycle peaks and troughs. Real wages arehourly compensation deflated by the consumer price index. Product wages are hourly compensationdeflated by the value-added deflator. All data are for the manufacturing sector. (See data appendix forsources.) All estimates are Ordinary Least Squares. S.E. is standard error of equation. All annual ratesof change are calculated as IOO (log xt- log Xt-).

of these equations is rather limited, however, as there are few degrees offreedom.'

IV. CHANGES IN THE CYCLICAL VARIABILITY OF WAGES

The results of the previous section provide support for the existence of procyclicalreal and product wages in the period I955-70. What can be said about theIO years following I970? In this section I present evidence that over the fullperiod I955-80 there was a substantial decline in the cyclical variability of realwages. By the end of the period there appears to be countercyclical movement.

In the I970S the previously stable relation between wage change and un-employment appeared to have been altered. In the United States, the I975Annual Report f the Council of Economic Advisers contained evidence for theview that there is 'more resistance to moderating cyclical forces after the down-turn in recent cycles than in earlier ones'. At the Tenth Anniversary of theBrookings Panel on Economic Activity, the two papers which contained

empirical estimates of the relation between nominal wage change and un-employment adopted a similar view.2While there has been a fair amount of theoretical attention to this issue, the

empirical evidence is limited to a few studies of nominal wages. Sachs (I980)provided the first correctly specified model and found evidence of a fall in thecoefficient of the cyclical variable for the United States.3 Schultze's (i 98 I) study

1 The wage equations presented here were also estimated for nominal wages, with virtually identicalresults for both the I955-70 period, and for the hypothesis of declining cyclical variability.

2 For instance, George Perry (i98o) observes that 'the economic experience of the past decade hasconfirmed the limitations of stabilization policy for slowing inflation' (p. 207). James Tobin (I980)contends that 'the price and wage-setting institutions of the economy have an inflationary bias'(p. 64).

3 The original paper on this issue was by Wachter (1976). It contained results showing increasingcyclical variability but the model was misspecified because Wachter constrained the coefficient of theprice term to remain constant over the period and omitted an intercept dummy or a time trend. If infact an outward shift due to a rising coefficient on prices, rather than a slope change were occurring,Wachter's specification would force a specious steepening. Wachter reports equations which includethe time trend and the results for his product terms are no longer robust. Two papers by Perry ( 978,I980) also yield unsatisfying results. For more, see Baily's comments on Perry's I978 paper.



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I985] CHANGES IN THE CYCLICAL PATTERN OF REAL WAGES 46I

revealed reduced flexibility for both real and nominal wages in the twentiethcentury United States. Cagan (I975) presents similar evidence for pricebehaviour. I know of only one study which uses European data. For France,Boyer (I 979) documents a marked decline in the role of unemployment in wage

determination from the period before the second world war to that after.As a first step in identifying trends in the cyclical variability of wages, I haveexamined the ielative cyclical effect measures for the last two cycles. (SeeTable I.) There is a decline between the three-cycle and five-cycle averages innearly all nine countries. The generality of this result is certainly striking. Theall-country cycle means also indicate a monotonic downward trend from thesecond to the fifth cycles. Examining the disaggregated data (Table 2) one cansee that there has been declining cyclical variability in both the contraction andexpansion periods, however, the change is considerably arger for the latter.

Econometrically, I have chosen two methods for investigating the hypothesisedchange in the output gap coefficient: the estimation of the equation with asuccessively onger sample period and the inclusion of an interactive term betweena time trend and the output gap. The model with an interactive term is asfollows:

w-=b0+bj4+b2(T() +b3T+e2 (4)

where all variables are as previously defined, T is a time trend, and e2 is astochastic error term. This specification ncludes a constant and a time trend in

order to avoid biasing the slope change term in the event that the intercept is notequal to zero and/or the position of the output gap-wage change curve ischanging over time. Again, both real and product wage equations wereestimated.

Table 4 presents estimates of the original model (equation (3)) for the fullsample period, I955-80. A comparison of these estimates to those shown earlierreveals that the coefficient of the output gap in the real wage model declinedfromr284 to 020I. In the product wage model, the coefficient fell from o0434to o-o87.

The foregoing result is supported by the estimates of equation (4), also shownin Table 4. In the full-period real wage model the interactive term between timeand the output gap has the expected sign and is statistically significant.' Evalua-tion of the coefficient at the beginning and end of the period reveals a changefrom o0428 to 0-025 as shown in Table 4A, which records the value of the outputgap coefficient at the beginning, midpoint and end of the period. For the productwage model, the interactive term is also significant, and the coefficient values gofrom o-683 to - 0-472. I suspect that the difference between the two results islargely attributable to the fact that the consumer price index is generally morevolatile than the value-added deflator due to the inclusion of components fromflex-price sectors, such as food and energy.

These estimates also show a statistically significant negative time trend in the1 These results are robust with respect to various specifications, such as the separate estimation of

the periods I955-69 and I970-80, or the use of a binary slope variable between the two periods. Inaddition, the basic equation was also estimated without the United Kingdom in view of the extremevalue for the relative cyclical effect in the last cycle. The results are robust with respect to this change.



12/18


Table 4Wage Equations, Full Period

Number ofDependent obser-variable q T Constant vations R2 S.E.

(i) Real wages 020I _ - 0-360 86 o-I62 3715(I955-80) (4-I7I) (-0.894)

(2) Real wages 3I962 -I6.I30 - I 74-029 342.256 86 0-240 3-526(I955-80) (2-32 ) (-2-305) (2 -75 ) (2 749)

(3) Product wages 0-087 -0563 7I 0-005 5.009(I962-80) (I * 76) (-0-945)

(4) Product wages I26-6I6 -64-I86 - 62-820 123-I26 7I 0-247 4-358(i962-80) (4.883) (-4.879) (-0

595) (0-592)

Figures in parentheses are t-statistics. All variables are as previously defined. Time is a linear timetrend which takes the value of I -955 in the first observation and increases by o-ooi in each calendaryear. The interactive term is time times the output gap variable. To evaluate the coefficient of theoutput gap, add the estimated coefficient of the interactive term times the value of time to the estimatedcoefficient of the output gap. A negative sign on the product term indicates a declining output gapcoefficient over time.

Table 4 AFull Coefficient n Output Gap

1955 I962 I968 1980

Real wages 0-428 0-218 0-025Product wages - o-683 0298 - 0472

real wage model. The presence of a downward trend may appear to be sur-prising given that there is a general presumption that the nominal wage-un-employment curve has shifted upward. However, because the present model is insecond differences, he negative time trend indicates that over time the amountof acceleration or deceleration around business cycle'turning points has fallen,not that the position of the curve has shifted. It is thus a second piece of evidencein support of declining wage responsiveness. Estimates of models without theinteractive term (not shown here) reveal that both the time trend and theinteractive term contribute independent explanatory power to the equation.'

1 The calculations of the relative cyclical effect and the foregoing econometric estimates assume thatthe expansion and the contraction phases of the cycle are symmetric with respect to the effect ofunemployment on wage change. There is no a priori reason to presume an asymmetric response. How-ever, the question of whether symmetry does in fact exist is an important one. In the disaggregatedrelative cyclical effect measures, the decline in wage responsiveness during expansions was greaterthan the decline in wage responsiveness during contractions. However, preliminary econometricresults (not shown here) do not indicate this difference. Therefore, the empirical evidence thus farmust be viewed as ambiguous. Clearly, this issue is relevant to whether expansionary and contrac-tionary macroeconomic policy has symmetric effects on inflation rates, real wages and profits. Inorder to investigate a possible link between the changes in cyclical variability discussed here and thelong-term decline in profit rates found by Hill (I979), this issue should be pursued. At present, itremains unresolved.



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Individual country equations for equation (4) were also estimated. In allcountries but Sweden the slope change interaction term had the expected sign.The interaction term was significant at the IO level in five countries (Canada,France, Italy, the United Kingdom and the United States).

The foregoing discussion has focussed exclusively on the cyclical variability ofwages, without reference to supply side phenomena. In I973, however, theworld economy experienced productivity and energy shocks. The effect on wagesof these shocks has presumably been a decline in the trend rate of 'feasible' realwage growth, that is, a downward shift in the wage growth-unemploymentcurve. This issue has been investigated by Grubb et al. (I982) and Sachs (I979),among others. What is the relation between the fall in the trend growth of wagesand the decline in cyclical variability? In particular, are the results presentedhere merely artifacts caused by the failure to include these factors?

Table 5Wage Equations, Final Expansion Excluded

Numberof

Dependent obser-variable Tq T Constant vations R2 S.E.

(i) Real wages 42 977 -21745 -86 56o 170-456 76 0 229 3349(2 600) (-2 588) (_ I* I 62) (I.I63)

(2) Product wages 144 375 -73'210 I 104837 205 768 6i 0'239 4'519(4.044) (-4.039) (-0?737) (0?734)

Figures in parentheses are t-statistics. All variables are as previously defined.

For a number of reasons, I think it is implausible that supply shocks areartificially generating the results found here. First, the decline in cyclicalvariability began long before the shocks, as Table I reveals. Second, to myknowledge there is no theoretical reason to believe that changes in the trend rateof growth of wages have an effect on cyclical variability. Thus, the argument thatthese results are an artifact is most plausible in the case of an econometric bias,which presumably would exist if the number of expansion observations includedin the I973-80 sample period exceeded the number of contraction observations.If this were true, then the downward shift in wage growth occurring in theexpansion years would bias the results in favour of the hypothesis by appearingto indicate countercyclical movement. There are ten contraction observationsin that period and nine expansion observations.

In order to examine this issue, two tests have been carried out. First, equation(4) was estimated without the last expansion. The estimates indicate that theearlier results are robust with respect to this test, suggesting that decliningcyclical variability is not an artifact caused by supply side shocks.

Second, productivity and energy price variables were added to the model. Theproductivity terms were included both in the standard form for all othervariables (changes in rates of change between business cycle turning points) and



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as first differences (rates of growth obtaining at the cycle turning points). Energyprices were proxied by the world market price of Saudi Arabian crude oil. Inneither case were these variables significant, nor did they alter the basic results.

V. EXPLANATIONS OF THE DECLINE IN CYCLICAL VARIABILITY

No tests of a causal model have been carried out in this study. However, thereare at least four plausible explanations for the decline in the cyclical responsive-ness of wages. They are: increasingly accurate expectations of present and futurebusiness cycle conditions by workers and firms; a rise in the degree of labourmarket segmentation; an increase in the duration of collective bargainingagreements; and a growth in income-replacing social welfare expenditures.

The first explanation is that over time, workers and firms develop increasinglyaccurate expectations of business cycle conditions. A learning process occurs inwhich economic agents come to know the regular cyclical pattern of output,employment, wages, and prices. Because any current state of affairs s known tobe temporary, its ability to influence behaviour is reduced. This process s likelyto be especially pronounced in those countries where macroeconomic policy ispredictably practised by the state, whether it is procyclical or countercyclical.Baily (I978) has presented circumstantial evidence which indicates that in theUnited States learning about countercyclical policy may have reduced firms'variations of employment and output and that workers' expectations may havebeen similarly affected.

A second explanation is a rise in the degree of labour market segmentation.Segmentation results in relatively increased isolation from market pressures nthe restricted sector. To the extent that wage developments in the restricted, orprimary sector dominate aggregate wage movements, the increase in segmenta-tion may lead to an attenuation of wage variability. Wage differentials and thedegree of segmentation tend to narrow in economic expansions and widen inrecessions. Thus we would expect more segmentation on average n the I 970s due

to lower than average growth. The evidence for the United States is that seg-mentation has increased over the relevant period.'Third, there is a large body of literature on the effects of long-term contracting.2

These models show that increased contract duration results n less wage adjust-ment to current labour market conditions. Unless workers and firms accuratelyforecast abour market conditions 2 or 3 years ahead, multi-year contracting willlead to less contemporaneous wage response. In the United States there has beenan increase in the duration of contracts from a mean duration of I year in thelate I940s, to 3 years by the I970S.3 For countries other than the United States,only Canada and Italy have multi-year contracting, although there may havebeen a lengthening of duration in the United Kingdom over the period as well.4

1 For the movement of wage differentials over the cycle, see Okun (1973). For evidence on thedegree of segmentation see Gordon et al. (I982).

2 See Baily (1975), Okun (I98I), or Taylor (1979).3 Data from BLS Bulletin No. 2095 and earlier issues in the series.4 Sachs (I979) cites this development.



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I985] - CHANGES IN THE CYCLICAL PATTERN OF REAL WAGES 465

The final explanation concerns the growth of welfare state expenditures. Inthe model developed by Bowles (I 985), a growth in social welfare expenditures ofthe income-replacing type is likely to result in a fall in the responsiveness of wagesto business cycle conditions. This is because firms' wage offers are a function of the

interaction between unemploymentand social welfare spending which together

determine the cost ofjob loss to the worker. Evidence on this issue for the UnitedStates is presented in Schor (I985). Comprehensive and consistent data onincome-replacing social welfare expenditures are difficult to find; however, moreaggregated measures of social welfare spending do exist for the period I955-77.The two measures are likely to be highly correlated. For the United States, whereit has been possible to differentiate between income-replacing and non-income-

Table 6Ratio of Public to Private Consumption

All-countryYear Canada France Germany Italy Japan Neth. Sweden U.K. U.S. average

1955 OII2 0237 o0288 o-i62 0-070 OI48 O0I70 0-145 0'077 O'1571970 0-21I o0288 0o370 0-281 0-09I o0487 0-423 0o2i8 o*i62 0o28I1977 0274 053I 0'544 0-364 0oI35 0o80I 0927 o-286 0-244 0456

The ratio of public to private consumption is calculated as total social security expenditures divided

by the quantity, private final consumption minus transfers to households. The socialsecurity measure

includes all social welfare spending except housing and education. 1977 is the latest year for which dataare available.

replacing social welfare expenditures, the two measures are very closely related.'I have constructed a measure of social welfare spending which is the ratio ofsocial welfare spending, or publicly-provided consumption, to privately-provided consumption. These data are presented in Table 6. Large increases inthe ratio of public to private consumption occurred n all nine countries over theperiod. This pattern is consistent with the scope and pace of decline of the outputgap coefficient reported earlier.

VI. CONCLUSION

This study has yielded evidence on the cyclical pattern of real wages. In theperiod I 955-70 real wages displayed a strong procyclical pattern, a finding whichis consistent with Marxian reserve army models, as well as neoclassical and

Keynesian models, under certain assumptions. Furthermore, this patternchanged over the entire period I 955-80, as measured by a variety of econometrictechniques. This evidence was the basis of speculation at the beginning of thepaper that the current long-term deflationary policies may have been motivatedby the failure of chianges n output to moderate inflation and real wage growth.

1 See the Appendix to Schor and Bowles (1985).



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It remains to be seen whether a long period of restraint will in fact significantlylower the long-term growth of real wages.'Harvard University

Date ofreceipt ffinal typescript: October 984

APPENDIX I

Dating of Cyclical Peaks and Troughs f Output Gaps

Peak Trough Peak Trough

Canada 1956 1958 Netherlands 1961 i 9631959 1961 I965 I967I966 I968 1970 1972

I969 1970 1974 19751974 1975 1976'979

Sweden 1955 1958France 1957 1959 I962 I963

I964 I965 I965 I967I966 I968 1970 19721974 1975 1974 1978'977

United Kingdom 1955 1958Germany 1955 1958 1960 I963

I96I I963 I964 I967

I965 I967 I969 19721970 1972 '973 '975'973 '975 '977'979

United States 1955 1958Italy 1955 1958 1959 196I

I962 I965 I966 I967I967 I968 I968 19701970 1972 '973 '9751974 1975 19781976 1978

Japan 1957 1958I96I I963I964 I9651970 1972

'973 '975I1980

APPENDIX II

Data Sources

Wages. Wages are hourly compensation in manufacturing. All wage andproductivity data are from the BLS, Office of Productivity and Technology,'Underlying Data for Indices of Output Per Hour, Hourly Compensation, andUnit Labor Costs in Manufacturing, Eleven Countries, I 950-I 980', May I 98 I.

1 In the United States at least, the data on peak to trough changes in real wages show a strengtheningof the countercyclical movement in the last cycle due to the substantial fall in price inflation. Thissuggests that the results presented in this paper are not historical artifacts.



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Prices. Prices are consumer price indices from the International MonetaryFund, International inancial tatistics, arious issues, Washington: IMF.

Value-added eftators. alue-added deflators are for manufacturing. They wereobtained as unpublished data from the International Monetary Fund.

Output gaps. Output gaps are from Jacques Artus and Anthony Turner,'Measures of Potential Output in Manufacturing in Ten Industrial Countries,I955-80', Research Department, International Monetary Fund.

Social Security pending. ublic and private consumption were calculated fromInternational Labour Office, The Cost of Social Security, arious issues, Geneva:ILO, and Organization for Economic Co-operation and Development, I98I,National Accounts tatistics, Volume : Aggregates, 950-79, Paris: OECD.

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