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)| false Hossain, Akhtar, 2002, “ An Examination of Some Key Issues in the Conduct of Monetary Policy in Bangladesh,” in Studies on Socio-Economic and Political Development of Bangladesh, ed.by ( Ashraf Ali, Ruhul Kuddus, and Syed Andaleeb Dhaka: University Press Limited, forthcoming).
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I would like to thank Donal Donovan, Andrew Feltenstein, Leonardo Hernandez, Eduardo Ley, Olin Liu, Reza Vaez-Zadeh, and Luis Valdivieso for their comments on earlier drafts. Rachel Lee processed some data for the paper, for which I am grateful. Any errors and omissions are my responsibility alone.
Devaluation may help to bring about macroeconomic stability when a country experiences external imbalances caused by, for example, expansionary monetary and fiscal policies. It is accepted that macroeconomic stability is necessary, if not sufficient, for sustained economic growth. However, because devaluation is believed to have effects on inflation and income distribution, these effects complicate the design of stabilization policies. See Aghevli, Khan, and Montiel (1991), Agénor and Montiel (1999), and Edwards (1989a,b) for detailed discussion of the economic effects of devaluation in developing countries.
This follows from the purchasing power parity proposition. At the microeconomic level, the literature on the exchange rate pass-through deals with the effect of devaluation on the price of tradables. The exchange rate pass-through, which is usually less than unity for manufactured products, differs a great deal across industries, depending on industrial organization factors including market concentration, product substitutability, and the market shares of domestic and foreign firms (Dornbusch, 1987; Feenstra, 1995; Kenen and Pack, 1980; Krugman, 1987).
Ceteris paribus, the expectation of a devaluation can increase an excess of money supply and thereby inflation by lowering the demand for domestic money. In a market economy operating under a flexible exchange rate system, this effect is captured by the nominal interest rate via the uncovered interest rate parity proposition. However, if the domestic nominal interest rate is determined administratively or remains unresponsive to any expected change in the exchange rate, the exchange rate may affect money demand directly (provided that asset holders can substitute alternative assets, including foreign currency) and thereby affect inflation. Under a fixed or a pegged exchange rate system, changes in exchange rates may also affect the money supply through developments in the balance of payments. This paper, given its limited scope, does not explore such possibilities of devaluation affecting inflation through various monetary channels.
This is reflected in the coefficient of variation of the change in the nominal effective exchange rate, which, in the case of monthly data, increased from 2.7 during the 1980s to 17.7 during the 1990s. For details see Appendix I.
I received conflicting suggestions from reviewers of this paper on the use of both annual and monthly data. Although the use of only monthly data would have been economical, one reviewer suggested that I use both annual and monthly data to find out whether the results are robust to both. This has become somewhat necessary as, in the case of monthly data, I use industrial output as a proxy for real GDP. Also, because there is no clear-cut choice between the narrow and the broad definition of money, I have used both definitions in my investigation.
In general, the causal linkage between inflation and devaluation can be examined by applying Granger causality tests within an error correction modeling framework, provided that both these series are stochastic in nature or generated through market mechanisms within a general-equilibrium framework. Apparently this cannot be said for Bangladesh, whose exchange rate has been pegged to a single currency or a basket of foreign currencies since the early 1970s but where the authorities have adjusted the official exchange rate at different intervals under different exchange rate regimes. Nevertheless, the Granger causality framework remains valid in the present context. Note that although the exchange rate is probably set through an implicit policy rule, it is a function of random variables such as the inflation rate and hence can be considered a random variable. In any case, even if the exchange rate only moves in discrete steps, what is needed for the approach to remain valid is that the coefficient of the time trend in the equations for changes in the exchange rate not be zero. I owe these classifications to Professor Adrian Pagan.
This is a restrictive assumption. In general, openness and economic growth have different effects on the share of tradables in total expenditure. For example, the share of tradables in total expenditure rises as the economy opens, provided that opening lowers trade restrictions and transactions costs, both implicit and explicit. However, a rise in income per capita, with or without an increase in openness, increases the relative demand for nontradables and thereby causes a structural transformation of the economy in favor of nontradables. Therefore, analytically, ϕ can be expressed as a function of the real exchange rate, which, in turn, depends on a set of real and nominal factors, including economic growth, capital flows, the terms of trade, and the stance of monetary and fiscal policies (Edwards, 1989a; Hossain, 2000a).
For simplicity, the rate of expected inflation or the nominal interest rate (a proxy for the opportunity cost of holding money) is assumed constant.
In the simplified equation (2.8), the growth of real money is expressed as the differential between the growth of the money stock and the rate of inflation.
This is a variant of the conventional monetary model of inflation, such as Harberger’s (1963) model of inflation for Chile. This restricted model can be derived from the flow equilibrium condition of the money market, such that Δln (M/P) = Δln md(y).
Although the Bangladeshi authorities did not explicitly follow the real exchange rate targeting approach to exchange rate policy, such an approach was somewhat revealed by or could be discerned from the movement of the real exchange rate, especially since the early 1980s. However, this proposition is essentially a working hypothesis as part of establishing the point that external balance, rather than price stability, was the main objective of the government’s monetary and exchange rate policy. For a detailed discussion of monetary policy in Bangladesh, see Hossain (2002).
See Appendix II for the time-series properties of all variables used in the empirical investigation.
The equilibrium condition is ln (M/P) = ln md(y, i), where M is the money stock, P is the price level, and md is demand for real balances, which depends positively on real income (y) and negatively on the nominal interest rate (i). The interest rate can be ignored in the specification if it is determined administratively or does not change much over time. This equilibrium condition gives a price-level equation that depends positively on the money supply and negatively on real income.
Further tests are conducted by the Johansen procedure (Johansen, 1988, 1991; Johansen and Juselius, 1990). The results, which are consistent with those obtained by the Engle-Granger procedure, are not reported here but will be available from the author upon request.
One reviewer questioned the appropriateness of using the first-difference terms of cointegrating variables while estimating a cointegral relationship, as it might lead to a misspecification error. The use of such first-difference terms is a common practice in the applied literature, however. As pointed out in the text, these first-difference terms are considered unimportant but are included to lower the finite sample bias in estimates of long-run coefficients of variables in level form (Phillips, 1988; Reserve Bank of New Zealand, 1991). Insofar as the present paper indicates, the first-difference terms are I(0) and therefore can be used in a regression in which long-run relationships are formed among variables that arc I(1).
The value of the CRDW statistic is relatively low, however, conflicting with the ADF test results.
Given that a stable money demand function is a key component of the inflation model, one reviewer suggested that I estimate a money demand function, examine its stability, and test for causality between money supply growth and inflation. In a recent paper (Hossain, 2002), I adopted this approach to investigate the inflationary process in Bangladesh. To avoid repetition, therefore, I have not estimated a money demand function for this paper. However, the main finding of the earlier paper on money demand remains consistent with empirical results of the present paper.
One reviewer suggested that I use the basic Granger causality framework where the error correction term docs not appear. However, I think that because an error correction modeling approach is adopted here, the augmented causality framework is more appropriate. Without an error term in the specification, the results could be biased (Roca, 2000).
One reviewer suggested that I use the AIC to select the lag length. As indicated in the text, the reasoning behind a sequential increase in the number of lagged terms is that the causal inference drawn from causality tests is sensitive to the choice of lag length. By following this strategy, I have avoided selecting a lag length using a statistical criterion that is sometimes considered unreliable.
The detailed results are not presented here but are available from the author upon request.
The overall results should be interpreted with caution. As indicated earlier, I have not taken into account some indirect effects of devaluation on inflation, which would require specifying and estimating a structural model. Moreover, I have adopted a pragmatic, judgmental approach in conducting tests and interpreting results.
This view is reflected in a widely cited early study by Aghevli, Khan, and Montiel (1991, p. 13), who report that “the inflation performance of the countries that have operated under a fixed exchange rate regime has been, on the whole, superior to that of the group operating under more flexible arrangement.” However, a subsequent IMF study (1997, p. 87) raises doubt about such a relationship: “… it is not the case mat flexible exchange rates arc necessarily associated with higher inflation, as there are a number of countries with flexible exchange rate arrangements that have had relatively low inflation (and robust growth).”
The detailed results are not presented here but are available from the author upon request.
When both the AIC and the SBC were applied, results with optimal lag lengths were also generated. However, these results are not reported, because they are similar to those obtained with common lag lengths.
Conflicting results were found for ln M2. The results obtained following the defined testing procedure show that ln M2 is I(2), However, when the ADF(12) command facility of Pesaran and Persaran (1997) is applied, the series, reported with an asterisk mark, is found to be I(1). The latter result is accepted for the present purposes as it is consistent with the properties of other series.