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László Halpern is Senior Research Fellow and Deputy Director of the Institute of Economics of the Hungarian Academy of Sciences, and a Research Fellow of the Center for Economic Policy Research (CEPR) in London. He graduated from Karl Marx University of Economics (Budapest). Charles Wyplosz is Professor of Economics at the Graduate Institute of International Studies in Geneva, a Research Fellow of CEPR in London, and Managing Editor of Economic Policy. He received his Ph.D. from Harvard University. This paper was initially written while Charles Wyplosz was visiting the IMF’s European I Department, which provided invaluable support, advice, and help with data. László Halpern acknowledges support from the Hungarian Science Fund (OTKA 3231). Many individuals were involved in the data collection and advice regarding the data; the authors are deeply grateful to them all. They particularly thank Massimo Russo, Michael Deppler, and Gérard Bélanger for making this project at all possible and for useful advice, many IMF staff members for comments, Phi-Anh Plesch and Harmen Lehment for additional help with data, and David Maxwell and Nadine Orosa for excellent research assistance.
Let N = W/EW* be the conventional wage-deflated real exchange rate, where W and W* are, respectively, the domestic and foreign nominal wages. W/E = N if we assume that W* is constant and normalized to unity. Note that we do not average dollar and deutsche mark wages, as we do with the other exchange rates. The reason is that the dollar wage has become a highly recognized measure in the transition countries.
Note that the figure exhibits the dollar wage itself, not an index. As a result, the scale (shown on the right vertical axis) is often different from that used for the two relative price indices (shown on the left axis).
For a formal analysis of demand effects on the real exchange rate, see De Gregorio, Giovannini, and Wolf (1994).
For an extensive analysis of the cyclical and structural reasons behind the rapid fall in tax income, see Bélanger and others (1994).
Lipschitz and McDonald (1992) correctly suggest that international competitiveness is best captured by measures of producer profitability (the ratio of relative producer prices or value-added deflators to relative labor costs). Unfortunately, such data are not available for the transition countries.
We assume away the distinction between traded and nontraded good prices abroad. A symmetric treatment would lead to the following version of equation (8b) below: λ = κ + γθ + θ(ρT − ρN) + γ(aT − aN) − γ*(a*T − a*N) as in De Gregorio, Giovannini, and Wolf, 1994. We adopt the small country assumption because we do not have data on the “foreign country” to match the productivity terms a*T and a*N.
We define aggregate variables as x = γxN + (1 − γ) xT, for x = a, ρ. There is no compelling reason to use the same weight γ as for the CPI. The more general case is straightforward and qualitatively identical.
A “normalization” is under way but this contributes to pollution of the data. Furthermore, it proceeds at different speeds from one country to another.
The ILO Statistical Yearbook does not always provide monthly wages: in some countries it publishes instead hourly, weekly, or annual wages. We therefore need information to convert them into monthly wages. In many cases, we have requested information from the IMF country desk economist.
Given the size of our sample, the rejection of normality is unlikely to indicate serious misspecification.
Once we allow for slope effects, there is no significant fixed effect.
The negative sign for Southeast Asia supports the result by Young (1994) that these countries’ output has grown fast not because of gains in total factor productivity but because of fast, and not particularly efficient, increases in the use of inputs. This interpretation applies even more strongly to the planned economies. The positive value for the African dummy is more difficult to interpret. It might correspond to the opposite effect: while inputs grow very slowly, scarcity leads to a more efficient use. Clearly, more research is called for to elucidate these results.
The measured short-run effect may correspond to such inflation effects as capital flight and exchange rate overshooting.
Sensitivity analysis is straightforward: Table 2 shows that setting the benchmark inflation rate at 0 percent (20 percent) raises (lowers) the estimated equilibrium dollar wage by 1.4 percent, a trivial amount.
In regression 3, this dummy appears significantly when interacted with the GDP variable and with the relative shares of agriculture and industry in GDP (see the notes to Table 2 for details).
Most data have been provided by the IMF. For 1995 and 1996, as well as for PPP-adjusted GDP, we have used PlanEcon estimates.
As reported in Rogoff (1996), tests of PPP indicate that the convergence of the exchange rate toward its equilibrium (assumed to he PPP in most studies) is very slow, with a half-life of three to five years.
When the size of the labor force undergoes a rapid decline, employment and unemployment do not move systematically in opposite directions and can therefore be treated as two distinct regressors.
Of course, we could replace de with the rate of inflation, but this would be exactly the same regression. This observation may also explain why the velocity of money is not significant. Being strongly influenced by the rate of inflation, velocity may just be a proxy of inflation, which is highly collinear with the rate of depreciation.
Here we use λ, Looking at equations (8) and (9), it can he checked that the expressions for μ and ω are similar. Indeed, μ only differs from λ. by the “quality” parameter κ. while to includes aggregate productivity and the wage distortion factor ρ.
No reliable methodology is available to test stationarity in the panel estimation of an error correction model. For this reason it is not possible to properly test for cointegration. However, the real exchange rate, the real producer wage, labor productivity, employment, and unemployment are cointegrated, which gives some support for the estimation method used here. Columns (1) and (7) show that the intuitive panel estimation of the error correction mechanism should not be rejected. Misspecification systematically emerges when we interact the regressors with country-specific dummy variables, Heteroscedasticity must be assumed for our heterogeneous sample. The usual treatment—weighted ordinary least squares (OLS), using standard errors from single country estimations as weights—was tried but did not help.
The half-life is the lime it takes for a discrepancy between the actual and equilibrium exchange rate to be reduced by half, lt is computed as ln 2/α.
While we focus on the transforming economies, the results obtained here can be readily applied to any country.