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  • 1, International Monetary Fund
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Appendix I. Data Description and Sources

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The non-administered subcomponent of CPI is obtained by subtracting the administered component from overall CPI The administered component was made up of the following items and weights from 1997/98–2001/02: bread, 1.03 percent; sugar, 0.55 percent; vegetable oil, 0.9 percent; medicines, 1.1 percent; water, 0.35 percent; fuel, 0.81 percent; electricity, 0.47 percent; inter-urban bus transport, 0.27 percent; and inter-city air transport 0.13 percent. Slightly different weights were applied for the period from 1990/91–1996/97.


Appendix II. Time-Series Properties: Tests of Stationarity

1. This study uses a cointegration approach to identify a long-run equilibrium relationship in the money market in Iran. A set of variables which are integrated of order one (I(1)) are said to be cointegrated with each other if there is at least one linear combination of these variables which is stationary (I(0)). The order of integration of the variables which enter the money market equilibrium relationship given in equation (3) (m1p, y, dcpi, dpar) was investigated using the Augmented Dickey-Fuller (1979) and Phillips-Perron (1988) tests.22 For all the variables except y, the hypothesis of nonstationarity cannot be rejected at 5 percent confidence by at least one of the tests. The stationarity of y is rejected only at 10 percent confidence, but for samples ending before 2001 :Q4, at higher levels of confidence. For the first differences of the same variables, the hypothesis of nonstationarity is rejected at least 10 percent confidence, suggesting that the variables are I(1). Table 1 presents a summary of the unit-root test results based on the Augmented Dickey-Fuller (1981) procedure.

Table 1.

Unit Root Augmented Dickey-Fuller Test Statistics: 1990:Q3–2001:Q4

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Notes: Variables are as defined in the text. The test statistic is the coefficient of the first lag of the variable in a regression of first difference of the variable on its lags and a constant term, divided by its standard error. The criteria for lag selection is a modified version of the Akaike information criterion, as described by Pantula et. al. (1994). The critical values of the tests are taken from MacKinnon (1994). The asterisks * and ** indicate that the test statistic is significant at the 5 percent and 10 percent levels, respectively.


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  • Sundararajan, V., Michel Lazare, and Sherwyn Williams, 1999, “Exchange Rate Unification, the Equilibrium Real Exchange Rate, and the Choice of the Exchange Rate Regime: The Case of the Islamic Republic of Iran,” IMF Working Paper 99/15 (Washington: International Monetary Fund).

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We are grateful to Fredric Malek and Payman Ghorbani of the Bank Markazi Joumhouri Islami Iran for discussions and great help with obtaining much of the data used in the analysis. We thank Abdelali Jbili, Vitali Kramarenko, and Jean Le Dem for very useful comments and Alina Milasiute for excellent research assistance. All numerical results were obtained using PcGive version 10.1.


Liu and Adedeji (2000) present an econometric analysis of money demand and inflation for the period 1989/90–1998/99, and conclude that inflation was mainly a monetary phenomenon.


The Iranian fiscal year begins in March.


While this methodology allows for economic theory to guide the specification of long-run economic relationships, the specification of short-run dynamics is taken to be an empirical matter. See Kuijs (1998) for a useful outline of this methodology with an application to Nigeria.


See Sundararajan, Lazarc, and Williams (1998) for an account of the 1993 exchange rate unification.


Tradable goods comprise food, beverages and tobacco; clothing and footwear; and household goods, whereas nontradables comprise all remaining items in the CPI.


All the variables are in logarithms. Data description and sources are presented in Appendix A.


The levels of nominal interest rates are not used as they are administered and fairly constant over much of the sample period.


The parallel market for foreign exchange is an amalgam of several closely linked and integrated markets, and throughout the sample period, the only foreign exchange market where Iranian agents could obtain foreign currency for most capital account activities, including the purchase of foreign currency as a financial instrument that provides some hedge against domestic inflation.


The results of the unit-root tests arc summarized in Appendix B.


Liu and Adedeji (2000) report an estimated output elasticity of real M2 of 0.63 for a sample from 1989/90–1998/99 and Pesaran (2000) reports an elasticity of 0.53 for 1979/80–1995/96, both of which are reasonably close to our estimate of 0,57.


The concept of weak exogeneity is described in Johansen (1992).


The cointegration test was also carried out for real M2 balances, real output, inflation rate and the parallel market depreciation rate, and the Johansen trace statistic indicated the statistical significance of a single cointegrating vector. However, all variables except the level of real M2 balances were found to be weakly exogenous to the cointegration relationship. Given that inflation was found not to adjust in response to disequilibria in broad money, the level of real Ml balances was used as the relevant measure of money supply.


The condition for a given level of inflation to reduce real money supply more than it reduces real money demand is -0.4dcpii+1<0.6dcpit, which is always satisfied for nonnegative rates of inflation.


The non-administered component of CPI excludes the prices of bread, sugar, vegetable oil, medicines, water, fuel, electricity, inter-urban bus transport and inter-city air transport, which comprise about 5 percent of the consumer price index basket.


Taking the change in overall CPI as the dependent variable does not change any of the subsequent results, including those relating to the stability of the model.


The exclusion restrictions were tested at each stage of the model reduction and only statistically insignificant restrictions were accepted.


Standard errors are in parentheses. Asterisks * and ** denote statistical significance of the coefficients at the 1 and 5 percent levels, respectively.


In the period ahead, the empirical relevance of the parallel market exchange rate for inflation is likely to decrease and to be replaced by that of the official rate. The ongoing process of opening to international trade is likely to increase the direct effect of the official exchange rate on CPI inflation. Also, as foreign currency transactions for capital account purposes are gradually liberalized and shifted to the official market, the relevance of the parallel market for asset substitution purposes might diminish.


From mid-1999 until the March 2002 exchange rate unification, all imported goods excluding imports of subsidized goods were imported at the TSE exchange rate. Following March 2002, most imports were applied the official exchange rate prevailing at the interbank market for foreign exchange, but essential imports were subsidized. Both the TSE exchange rate during 2000–2002 and the interbank rate after March 2002 were very stable.


This effect is likely to have been partially captured by our model, as the stability of the TSE and interbank exchange rates during the period from 2000–2002 was matched by the stability of the parallel rate, which is included as an explanatory variable in the model.


Two versions of the tests were carried out, by allowing a constant, and a constant and a trend to enter the equations. The test results were not significantly different for the two versions, and only the tests including a constant term are reported in the paper.

An Analysis of Money Demand and Inflation in the Islamic Republic of Iran
Author: Mr. Mangal Goswami and Oya Celasun