Bacha, Edmar L., “Notes on the Brazilian experience with minidevaluations, 1968-1976”, Journal of Development Economics 6, (1979), 463–481.
Beveridge, S., and C. Nelson, “A new approach to decomposition of economic time series into permanent and transitory components with particular attention to the measurement of the business cycle,” Journal of Monetary Economics 7, (1981), 151–174.
Carrasquilla, A., “Estabilizacion macroeconomica y tasas de interes en Colombia: se agoto otro modelo?,” mimeo, Banco de la Republica (Colombia), (1992).
Corbo, V., and A. Solimano, “Chile’s experience with stabilization revisited,” in M. Bruno, S. Fischer, E. Helpman, and N. Liviatan, eds., Lessons of economic stabilization and its aftermath, (Cambridge, Massachusetts: MIT Press, 1991), 57–91.
Edwards, S., “Comments on: Chile’s experience with stabilization revisited”, in M. Bruno, S. Fischer, E. Helpman, and N. Liviatan, eds., Lessons of economic stabilization and its aftermath (Cambridge, Massachusetts: MIT Press, 1991), 92–98.
French-Davis, R., “Exchange rate policies in Chile: the experience with the crawling-peg,” in J. Williamson, ed., Exchange rate rules (New York: St. Martin’s Press, 1981).
Fontaine, J.A., “La administración de la política monetaria en Chile, 1985-1989,” Cuadernos de Economía 28 (Chile), (1991), 109–129.
Guidotti, P.A., and C.A. Végh, “Macroeconomic interdependence under capital controls: a two-country model of dual exchange rates,” Journal of International Economics 32, (1992), 342–367.
Herrera, S., “Qué tan grande es el desequilibrio cambiarlo en Colombia?,” Ensayos sobre política económica 20 (Banco de la Republica, Colombia), 145–174.
Johansen, S., “Estimation and hypothesis testing of cointegration vectors in Gaussian vector autoregressive models,” Econometrica 59, (1991), 1551–1580.
Lizondo, J.S., “Real exchange rate targets, nominal exchange rate policies, and inflation,” Revista de Análisis Económico 6, (1991), 5–22.
Miller, S., “The Beveridge-Nelson decomposition of economic time series: another economical computational method,” Journal of Monetary Economics 21, (1988), 141–142.
Montiel, P.J. and J.D. Ostry, “Macroeconomic implications of real exchange rate targeting in developing countries,” IMF Staff Papers 38, (1991), 872–900.
Montiel, P.J. and J.D. Ostry, “Real exchange rate targeting under capital controls: can money provide a nominal anchor?,” IMF Staff Papers 39, (1992), 58–78.
Osterwald-Lenum, M., “A note with quantiles of the asymptotic distribution of the maximum likelihood cointegration rank test statistic,” Oxford Bulletin of Economics and Statistics 54, (1992), 461–469.
Ostry, J. and C.M. Reinhart, “Private saving and terms of trade shocks: evidence from developing countries,” IMF Staff Papers 39, (1992), 495–517.
Reinhart, C.M. and C.A. Végh, “Nominal interest rates, consumption booms, and lack of credibility: a quantitative examination,” mimeo, IMF, (1992).
Reinhart, C.M. and C.A. Végh, “Intertemporal consumption substitution and inflation stabilization: an empirical investigation,” mimeo, IMF, (1993).
Stock, J., and M. Watson, “A simple MLE of cointegrating vectors in higher order integrated systems,” NBER Technical Working Paper 83, (1990).
Williamson, J., “The crawling peg in historical perspective,” in J. Williamson, ed., Exchange rate rules (New York: St. Martin’s Press, 1981).
This paper was prepared for the Sixth InterAmerican Seminar on Economics, held in Caracas, May 28-29, 1993, and will be published in a special issue of the Journal of Developments Economics. We are grateful to Miguel Kiguel, Saul Lizondo, Peter Montiel, Jonathan Ostry, Jorge Roldos, Andrés Velasco, Sweder van Wijnbergen, and seminar participants at Caracas, the University of Montreal, the 1993 Meetings of the Latin American Econometric Society (Tucuman, Argentina), and the International Monetary Fund for helpful comments and suggestions.
It is certainly not the only one. Chile, for instance, has also used an interest rate on an indexed bond (i.e., a real interest rate) as its main policy instrument since 1985 (see Fontaine (1991)).
The PPP exchange rate was computed as the ratio of the domestic CPI to that of the United States, and based by giving it the same value as the actual exchange in 78.01 for Brazil, 85.07 for Chile, and 86.01 for Colombia. (The United States inflation is used for simplicity; more comprehensive indices could be used.) In the case of Brazil, the base date coincides with the beginning of the sample; in the case of Chile and Colombia, the base date marks the beginning of a period during which a PPP rule was in effect (see below).
In work in progress, we have analyzed these episodes in detail, and provided econometric evidence on the existence of PPP rules. Specifically, using monthly data we test for Granger-causality between inflation and the nominal exchange rate. As expected, we find that during the periods in which a PPP rule was in effect in both Chile and Colombia, the rate of devaluation was Granger-caused by inflation. In Brazil, considering the whole sample with the exception of the Cruzado and Collor plans, inflation also Granger-causes the rate of devaluation.
Early work by Dornbusch (1982) focused on the effects of PPP rules on the trade-off between output and price-level variability. Many of the issues raised by Adams and Gros (1986) had already been discussed at length in the volume edited by Williamson (1981), mainly in the context of Latin America.
The essential mechanism provided in Lizondo is retained in the other three papers.
The elasticity of money demand is assumed to be below unity, so that higher inflation implies higher revenues from the inflation tax.
Lizondo (1991) implicitly assumes that government expenditure on traded goods acts as the residual variable in that it accommodates the higher tax revenues.
The welfare effects, however, differ. Targeting a more depreciated level is welfare reducing because there is no initial distortion. In contrast, keeping the real exchange rate constant is welfare improving because it offsets the distortion introduced by the temporary external shock.
Note that in all the empirical work reported in this paper, a rise in the real exchange rate index denotes a depreciation.
To simplify notation and without loss of generality, we assume that, aside from endowment of future output, the individual and the country have no wealth as of time 0.
The real exchange rate is defined as e=EP*/P, where E is the nominal exchange rate, in units of domestic currency per unit of foreign currency, P* is the (constant) foreign currency price of traded goods, and P is the nominal price of home goods.
To ensure the existence of a steady-state and eliminate inessential dynamics from the case of perfect capital mobility, it will be assumed that β=r.
For the sake of simplicity, no new notation is introduced to denote equilibrium values.
This property does not hold in Lizondo (1991, 1992) and in Montiel and Ostry (1991, 1992), because in those models changes in the revenues from the inflation tax are not returned to the household in a lump-sum way, and thus affect private wealth. Hence, in these other models it is (implicitly) assumed that fiscal policy changes with inflation.
In more realistic examples, the real exchange rate associated with the social optimum may not be constant over time. However, one could prove in a wide variety of cases that any attempt to modify the real exchange rate via monetary policy (specifically, by variable-rate-of-devaluation policy) would result in an inferior social outcome.
As discussed below, for the resource constraint (10) to hold, e¿ must be below its initial steady-state.
Figure 3, Panel C assumes that the nominal interest increases with respect to its initial level (see the discussion below).
Naturally, this interpretation assumes that the inflation tax is used to finance in part a given level of government spending. As long as government spending does not change (i.e., as long as the fiscal adjustment involves only substituting the inflation tax for conventional taxes), the above model can be readily reinterpreted in this way since Ricardian equivalence holds.
We are assuming, of course, that private sector expectations are not altered by the policy response itself.
If the system converges to a steady state, however, the domestic real interest rate converges to the subjective rate of discount β which, by assumption, equals the international real interest rate (in terms of tradables), r.
Recall that, by assumption, foreign inflation is zero.
Notice that we can choose the same values of c*1 and c*2 as in the case of perfect capital mobility, because the economy as a whole faces the same overall budget constraint (10), independently of the degree of international capital mobility.
Note that it cannot be the case that (1/α + β)λ0 = uc*(c*1, y)/α. If that were the case, λt = λ0 for 0 ≤ t < T, and terminal condition (15) would be violated.
A PPP-type of rule would lead to similar results, as discussed above for the perfect capital mobility case. Hence, keeping the real exchange rate constant in the face of a temporary external shock that would tend to appreciate would entail high and rising real interest rates relative to the situation in which there is no policy response.
This point should be obvious as the path of c is the same in both cases.
See Reinhart and Végh (1992) for a similar numerical exercise under perfect capital mobility aimed at assessing the quantitative importance of imperfect credibility in exchange rate-based stabilization programs.
For the sake of concreteness, we will assume that the nominal interest rate does not change across steady-states and is equal to r (i.e., steady-state inflation is zero). Hence, in the first period the nominal interest rate is higher than in the initial steady-state.
It should be noted that, when q = 1, equation (26) reduces to the closed-form solution that would hold in a one-good model (see Reinhart and Végh (1992)). Thus, this two-good setting makes clear that the intertemporal elasticity of substitution relevant for traded-goods consumption, η*, may differ from the elasticity of substitution of aggregate consumption, η. Note that if q = 1, then η* = η. However, if η < 1 (the relevant case in practice, see below), η* > η.
Note that i is an interest rate which is capitalized instantaneously. Hence, if i is 3 percent, the corresponding yearly effective rate is 3.05 percent and the monthly rate is 0.25 percent. In what follows, and unless otherwise noticed, all interest rates are in expressed in effective terms.
The following parameter values were used: α = 0.15, q = 0.4, r = 0.03. Without loss of generality (since we are only concerned with percentage changes), y* and e were normalized to one. The values of α and q were chosen based on actual data. Specifically, the ratio of M1 to private consumption was 20.0 for Brazil in 1987, 10.7 for Chile in 1991, and 16.7 for Colombia in 1988 (data from IFS). The value of q, based on the average share of traded-goods consumption during the period 1978-1986, was 0.47 for Brazil and 0.32 for Colombia (see Reinhart and Ostry (1992)).
Reinhart and Végh (1993) use for estimation purposes a monetary model very similar to the present cash-in-advance model, but which allows for variable velocity. Depending on the parameters of the money demand, estimates for η lie in the range 0.18-0.19 for Argentina, 0.07-0.13 for Brazil, 0.24-0.80 for Chile, 0.16-0.22 for Israel, 0.11-0.19 for Mexico, and 0.04-0.17 for Uruguay. With the exception of Uruguay for some parameters of the money demand, all estimates are statistically significantly different from zero. In a real model, using panel data which includes Brazil and Colombia, Ostry and Reinhart (1992) estimate the intertemporal elasticity of substitution to be around 0.4.
The steady-state value of the real interest rate is 0.25 percent per month. In all the entries of Table 2, the value of the real interest rate at time 0 is higher than the steady-state value, although in some cases this is not evident from the table as figures have been rounded to two decimals.
In other models, there exists a steady-state relationship between inflation and the real exchange rate (see below).
Of course, other policy actions not discussed in this paper, such as changes in government spending on nontraded goods will have permanent effects on the real exchange rate (see for example, Edwards (1989)).
When the test results give conflicting results (for instance, inflation in Colombia), a higher weight is attached to the P-P tests, which allow for more general forms of heteroskedasticity.
This is not surprising, since the two variables have different orders of integration any relationship is bound to be spurious.
See Herrera (1991) and Carrasquilla (1992), who note for the case of Colombia that periods of real exchange rate disequilibria (in the sense of an overdepreciated real exchange rate) were characterized by inflationary pressures.
As discussed above, for the countries in our sample, inflation is stationary while the real exchange rate is not. Hence, no direct steady-state relationship can exist between these two variables.
It is assumed that real money demand is inelastic, so that an increase in inflation raises revenues from the inflation tax.
These results are not reported but are available upon request.
Indeed, the variables appear with the anticipated signs, but since these parameter estimates are neither unbiased nor consistent no conclusions are possible.
See Roldos (1990) for some empirical evidence on the effects of government spending and other exogenous variables on the real exchange rate.
It is important to note, that the figures for Colombia understate importantly the recent surge in capital inflows, as some of the capital inflows were being recorded as transfers in the current account.