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Mr. Le Fort, an economist in the Research Department when this paper was written, is now in the Asian Department of the Fund. He holds degrees from the Universidad de Chile and the University of California, Los Angeles.
This paper is based in part on the author’s Ph.D. dissertation (Le Fort (1985)). He is indebted to Edward Learner, Axel Leijonhufvud, and Sebastian Edwards for their guidance, and to his colleagues in the Fund for their valuable comments.
Similar structural reforms, ending in recessions and crises, were also attempted in Argentina and Uruguay. The experiences of these three countries, which have been analyzed at length in recent economic literature, provide important lessons on problems and policy mistakes during the process of economic liberalization.
Nontraded goods are goods that under the current technology, tastes, trade restrictions, transport costs, and international prices cannot be traded internationally, and thus their prices are determined domestically. Traded goods are goods that are actually exported or imported and their close substitutes produced domestically; their prices in international currency are determined abroad. The term “relative price of nontraded goods” refers to the price of nontraded goods relative to the price of exportable goods and can be interpreted as the real exchange rate.
The structural changes and the stabilization policy attempted in the Chilean economy during the late 1970s, as well as the results obtained, have attracted the interest of economists. Among the many studies produced are those by Corbo (1983, 1985a, 1985b); Corbo and de Melo (1986); Cortazar (1983); Diaz Alejandro (1981); Edwards (1983, 1986); Foxley (1983); Harberger (1982); Ramos (1984); and Zahler (1983).
Given the high inflation rate in the Chilean economy during most of the period analyzed, and that the direct effect of the exchange rate policy on relative prices takes place through real producer wages, the level of public sector wages relative to exportable goods prices was taken to represent the direct effect of the exchange rate policy on relative prices.
The posterior estimates were obtained using a program (SEARCH, Learner and Leonard (1983b)) that derives a contract curve between the prior beliefs and the least-squares estimates. The least-squares estimates were represented in this case by instrumental-variables estimates, and the prior beliefs were derived from each of the alternative theories.
The specification of prior beliefs, used in Section II below, is much easier when the parameters are elasticities.
This simplifying assumption is often used in the literature. According to the Balassa (1964) effect, technical change increases the price of nontraded goods relative to traded goods because technical progress is slower in the former sector. Limitations of the data prevent the use of a more general approach.
The implications of the sticky-wages model for the relative price of nontraded goods would not be different with intersectoral capital mobility.
A devaluation accompanied by an increase in public sector wages of the same proportion would not have any real effect according to this model. This restriction is consistent with rationality and helps to avoid estimation problems—that is. multicollinearity—that arise with the use of nominal explanatory variables in an inflationary environment.
It is also assumed that the rate of technical change is equal in the importables and exportables sectors and slower in the nontraded-goods sector, and that the rate of capital accumulation is the same in all the sectors. The derivation of the labor demand is presented in Appendix I.
In an open economy that can accumulate international assets, the domestic demand for goods in each particular period is restricted by real absorption, the sum of income, and the current account deficit. In this paper, the determination of absorption through intertemporal optimization is not modeled; explicit modeling is included in Le Fort (1985, Chapter 3).
E is nominal absorption; (E/P), real absorption; and P, the relevant price index. In the nontraded-goods market, fN is the supply price elasticity; gN, the demand price elasticity; gM, the demand cross-price elasticity; and g, the demand income elasticity.
A model with the same type of explanatory variables can be obtained with perfectly mobile factors if it is assumed that there are more factors than internationally traded goods. In the literature, several macroeconomic models of this type—that is, relating absorption and the price of nontraded goods—have been developed without explicit microeconomic foundations. See, for example, Dornbusch (1973).
The term (gN + fN − fNYN) represents the price elasticity of the excess supply of nontraded goods after the indirect effects of wages (fNYN) have been taken into account.
The assumptions of the PPP solution are, in general, not met in the short run; PPP implications have been rejected empirically for developed countries with flexible exchange rates (see Frenkel (1981)). An empirical question that is still open, however, is the length of the period needed for PPP conditions to hold.
See Balassa (1964). These results refer to partial effects of technical progress, assuming that the international prices of importables and exportables are constant. Technical change could reduce the relative price of nontraded goods even when the progress is faster in both traded-good s sectors than in the nontraded-goods sector, provided that the technical change in the capital-intensive sector (M) is faster than in sector T by an amount large enough to compensate for the Balassa effect.
̂Tec=(1−αN+αN/C)̂CT; ̂Tec is a proxy for technical change obtained from the economy-wide average labor productivity, αN represents the share of non-traded goods in GDP, ĈT is technical change in the ex portable-goods sector, and ĈT/c is technical change in the nontraded-goods sector. Equation (13) was estimated in the level of variables, with the values of the variables normalized to unity in the fourth quarter of 1977. The parameters represent elasticities around the value of the variables in that particular period.
This specification was selected because the rate-of-change linear model and the log-linear model indicated autocorrelation of the residuals that could not be corrected using a standard first-order autoregressive procedure. The results for the estimation that is linear in the variables did not show evidence of first-order autocorrelation.
To avoid excessive extension of the paper and a deviation from its main line of exposition, a structural equation for absorption is not specified, and only the set of instrumental variables used is presented.
The expected rate of change in the price of traded goods is equal to the sum of expected international inflation and expected currency devaluation. Expected international inflation was assumed to be formed adaptively. The expected rate of devaluation was constructed using a “peso problem” approach in which two events are possible for the exchange rate policy in each period: either the rate of devaluation is equal to the preannounced rate, or a major devaluation takes place. The probability of a policy break was obtained from a signal given by the change in international reserves and subjective information based on previous actions of the authorities.
The expected rate of change in the relative price of nontraded goods was obtained by assuming rational expectations and an information set limited to the value of the variables up to the last period. A linear projection of all the variables lagged one period was used to estimate the expected relative price for the next period; the expected rate of change of relative prices was calculated from that value.
The international creditworthiness index was constructed using variables that can affect the benefits and costs of debt default, including the debt-to-GDP ratio, the financial services-to-exports ratio, and the invcstment-to-GDP ratio, among others. The variables and coefficients were obtained from Edwards (1984).
The posterior estimates were computed using the program SEARCH (Learner and Leonard (1983b)). Table 4 presents statistics for the residuals in the semireduced-form estimations, including the coefficient of determination corrected by degrees of freedom (¯R2), the sum of squared errors (ESS), the Durbin-Watson statistic (DW), and the Durbin h test.
The sign of the long-run multiplier is given by the sum of the current and lagged (WP/PT) coefficients.
A devaluation increases the denominator of (WP/PT). The interim multipliers were developed for a once-and-for-all change in the explanatory variable and consequently, in this case, for a constant WP.
The indexation coefficient was assumed to equal 0.75 in the quarter immediately following the devaluation, 0.9 in the second, and unity in the third and following quarters. Nontraded goods represent 50 percent of the price index weights. The simulation was performed assuming that public sector wages are indexed to inflation in the preceding quarter.