One of the key factors that determine the effect of a discrete change in the exchange rate of a country on its trade balance is the effect on that country’s costs and prices. The larger the rise in money costs, domestic prices, and export prices (in domestic currency) that is induced by a devaluation, the smaller will be the competitive price advantage achieved by the country from devaluation—and hence, ceteris paribus, the smaller will be the improvement in the trade balance.
This paper provides an empirical analysis of the probable effect of exchange rate changes on aggregate wage and price behavior in the United Kingdom. A simple and widely used two-equation model of wage and price behavior in the United Kingdom is employed in the study. One disadvantage of this model, however, is its partial equilibrium nature, which limits the analysis to providing information on only a few of the important effects of an exchange rate change. Specifically, the model is used to estimate only three effects of an exchange rate change: (1) the effect of a change in import prices on domestic prices (retail prices), (2) the effect of this initial change in retail prices on wage rates, and (3) the effect of this change in money wage rates on retail prices and any subsequent round effects on wage rates and retail prices. In other words, the analysis ignores or considers as exogenous these other effects of an exchange rate change: (1) the effect of an exchange rate change on import prices,1 (2) the effect of changes in import prices and domestic prices on the initiating country’s export prices,2 and (3) the effect of exchange-rate-induced changes in wages and prices on the supply and demand for factors of production, for commodities, and for money.3
The paper has four sections. In Section I, the basic model is introduced and briefly discussed, and the relevant parameters for assessing the impact of exchange rate changes on wages and prices are identified. In Section II, the price equation of the model is estimated (separately at first) to obtain information on the effect of changes in import prices on retail prices. Tests are also conducted to determine whether the devaluation of sterling in 1967 was followed by any change in the relationship between changes in import prices and retail prices. In Section III, the aggregate wage and price equations are estimated for the 1954—71 period (and for certain subperiods) in order to identify the effect of retail price changes on wage rate changes and the effect of wage rate changes on retail price changes. Tests are also conducted to determine whether the devaluation of sterling in 1967 was followed by any significant change in wage or price behavior. The main conclusions of the study are presented at the ends of Sections II and III and in Section IV.
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)| false Barker, Terence S., and J. R. C. Lecomber, “The Import Content of Final Expenditures for the United Kingdom 1954–1972,” Bulletin of the Oxford University Institute of Economics and Statistics, Vol. 32( February 1970), pp. 1– 17. 10.1111/j.1468-0084.1970.mp32001001.x
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Mr. Goldstein, an economist in the Special Studies Division of the Research Department, is a graduate of Rutgers University and of New York University. He was formerly a Research Fellow in Economics at The Brookings Institution.
In addition to his colleagues in the Research Department, the author wishes to express his appreciation to Orley Ashenfelter, George Johnson, Hugh Pitcher, and Robert S. Smith for helpful comments on an earlier draft of this paper. A previous draft of the paper was presented at the winter meetings of the Econometric Society, held in New York, December 1973.
For an empirical analysis of the effect of the 1967 devaluation on U.K. import prices, see Barker (1968).
A theoretical treatment of the effect of exchange rate changes on the supply and demand for factors of production and commodities is presented in Salop (1972). Also see Johnson (1972) for the effect of exchange rate changes on the demand for money.
In some wage equations, the change in the proportion of the labor force that is unionized is entered as an additional explanatory variable as a proxy for trade union aggressiveness; see Goldstein (1972) for a review of wage inflation models that use this variable. The unionization rate could not be used in this study because quarterly data on the proportion of the labor force unionized are not available.
Unexpected price changes are assumed to affect wage changes through the excess demand for labor (by means of the unemployment rate)—that is, an unexpected price increase lowers the real wage which, in turn, affects both the supply and demand for labor; see Friedman (1968).
Branson (1972, pp. 15–58) has shown that
In 1967, the United Kingdom devalued the pound sterling by 16.7 per cent in terms of pounds sterling per U. S. dollar (the relevant exchange rate measure for equation (6)) or by 14.3 per cent in terms of U. S. dollars per pound sterling (the more familiar exchange rate measure). If K = 1 and
It should be clear from the preceding analysis that one can estimate the long-run effect of
It is also possible that a devaluation could worsen a country’s competitive price advantage (and equation 11 would then be negative). This would occur if
The reduction in the real wage by 1 per cent in the above example reflects an increase in retail prices by 2 per cent and an increase in money wages by 1 per cent.
For a list of the policy measures adopted in the year following the 1967 devaluation, see National Institute of Economic and Social Research (1969, p. 5).
It should be noted, however, that an analysis of wage and price behavior after the 1949 devaluation of sterling would encounter many of the same problems facing this paper because the 1948:III–1950:III period was also a period of incomes policy. In order to obtain enough observations to test the above hypothesis adequately, it would probably be necessary to study exchange rate changes for many countries.
In view of (1) the difficulties associated with obtaining a reliable quarterly index of effective indirect tax rates, and (2) the finding by Lipsey and Parkin (1970) and Solow (1972) that indirect taxes did not have a significant effect on U. K. price changes, indirect taxes were not included in the price change regressions of this study. After this study was virtually completed, however, the author discovered a recent paper by Burrows and Hitiris (1972) that does find a significant effect for indirect tax changes on U. K. retail price changes, at least for the 1955–67 period. Clearly, additional empirical work on the effect of indirect taxes on U.K. prices is called for to determine the appropriate specification of the price equation.
For an application of the input-output method to estimating the effect of
The correlation is spurious only in the statistical sense that part of the dependent variable and part of the
The distributed lags tried for
Whether or not a decrease in the lag between
Of course, there could be other factors responsible for the decrease in the lag between
Recall from equation (4) that the coefficient on
The average import content of consumption expenditure was 19.2 per cent in 1954, and the estimated figure for 1972 is 15.2 per cent (see Barker and Lecomber, 1970, p. 7). The average import content of consumption is used in the above analysis as a proxy for the share of imports in the retail price index because estimates of the latter are, to the author’s knowledge, unavailable.
t= 1.40 < t0.05 = 2.000.
Smith (1968) also used half-yearly data to estimate some price equations.
In these equations, the coefficient on
The results without the incomes policy dummies were nearly identical, except that the Durbin-Watson statistic was lower.
When the slope dummy on wages was included in the price equation without the slope dummy on import prices present, the coefficient on DV -
Lipsey and Parkin (1970) estimated separate price equations for periods of incomes policy, periods of no incomes policy, and for the 1948–68 period as a whole. The estimated coefficient on
From footnote 29, note that equation (a) implies that
Suppose a slope dummy DIC6
A similar point was, of course, made long ago by Orcutt (1950) with respect to the price elasticity of demand (for imports and exports) for large and small price changes. Kindleberger (1963, p. 157) has also argued that the price elasticity of demand is likely to be greater for large than for small price changes because consumers have more inducement to overcome their inertia and the cost of shifting to substitutes. A similar argument for producers, based on the costs of changing prices, can be made for large versus small changes in costs.
That is, performing a t test on the coefficient c3 in equation (b) is equivalent to performing a t test on the equality of the coefficients a2 and a3 in equation (a).
See Rowley and Wilton (1973) for an analysis of what happens to the significance of estimated coefficients in quarterly wage equations when an efficient generalized least-squares estimator is employed to remove the autocorrelation in these equations. In brief, Rowley and Wilton find that when the generalized least-squares estimator is employed, the t statistics on the coefficients in the equation are reduced by 50 per cent or more in many cases. The examples used in the Rowley-Wilton study, however, are wage equations for the United States and Canada; no tests are done on U.K. wage equations.
In the future, it would probably be useful to re-estimate the wage equations reported in this paper with a generalized least-squares estimator (as outlined by Rowley and Wilton) to see whether the results are significantly affected.
See, however, the recent paper by Ashenfelter and Pencavel (1974) for a model of earnings changes in the United Kingdom.
Recall from equation (4) that the coefficient on
Lipsey and Parkin (1970) also found that a linear form of the unemployment rate provided the best results.
As a further example of this instability, compare the coefficients in equations (23) and (25) with those in the following wage equation estimated (also by ordinary least- squares) for the 1950:I–1967:II period.
Godfrey and Taylor (1973) also found a positive coefficient on the registered unemployment rate in their earnings function for the United Kingdom over the 1955–70 period.
A thorough investigation into why the wage equation shifted upward in the period 1967–71 (aside from any effects of the 1967 devaluation) was considered to be beyond the scope of this paper. It might be mentioned, however, that no convincing explanation for this upward shift in the wage equation has yet been found, to judge from the current U.K. wage literature (see the papers in Johnson and Nobay (1971), especially the study by Artis (1971)).*
The calculated F statistic for the two slope dummies in equation (27a) is 5.9, which is significant at the 1 per cent level. Thus, the null hypothesis that the explanatory power of the wage equation is not significantly improved by addition of the slope dummies on UW5t and Pt is rejected. Strictly speaking, the F test cannot be applied to an equation with a significant autocorrelation, but this problem is ignored here.
It might also be the case that expected price changes are a function of past price changes and exchange rate changes—that is, a devaluation might well lead to an increase in people’s price expectations. Unfortunately, a test of this hypothesis really requires direct survey data on price expectations, and such data are generally not available.
The author wishes to thank Rudolf Rhomberg for helpful discussion on this point.
Once again, it should be kept in mind that the autocorrelation causes the significance of the coefficients to be overstated.
The estimated size of this announcement effect should be treated with some caution because the announcement dummy and the last incomes policy dummy (DIC6) are quite collinear.
It should also be mentioned that a very similar test was conducted for the price equation but with quite different results. More specifically, when a shift dummy for the 1968:I–1968:IV period was included in the price equation, its coefficient turned out to be small (0.29) and quite insignificant (t = 0.42).
Lipsey and Parkin (1970) estimated separate wage equations for periods of incomes policy, periods of no incomes policy, and for the 1948—68 period as a whole. The estimated coefficient on
As previously demonstrated, the fact that the values for β3, α2, and β1 were higher in the 1967–71 period does not necessarily imply that they were significantly different from the values for 1954–71 in a statistical sense.
The values for β3, α2, and β1 cited above are obtained by adding the estimated coefficients on the slope dummies for the 1967–71 period to the estimated coefficients on
Cooper (1968) has also estimated the effect of a U.K. devaluation on the United Kingdom’s competitive price advantage. In terms of the notation in this paper, Cooper assumes that K = 0.85, (β3 = 0.19, α2 = 0.7 and β1 = 0.7. This implies that about 62 per cent of the initial price advantage achieved by devaluation will be retained (see Cooper, pp. 191–92).