Agénor, Pierre-Richard, 1990, “Stabilization Policies in Developing Countries with a Parallel Market for Foreign Exchange: A Formal Framework.”, Staff Papers, International Monetary Fund, Vol. 37 (September) pp. 560–92.
Doornik, Jurgen, and David Hendry, 1997, Modelling Dynamic Systems Using PcFiml 9.0 for Windows (London: International Thomson Business Press).
Edwards, Sebastian, 1994, “Real and Monetary Determinants of Real Exchange Rate Behavior: Theory and Evidence from Developing Countries,” Estimating Equilibrium Exchange Rates, ed. By John Williamson (Washington: Institute for International Economics).
Prepared by Louis Kuijs.
The model allows for a long-run effect of import prices on domestic prices. However, this effect is shown empirically not to be significant.
In the absence of non-oil exports, the national accounts identity for the non-oil sector is
Y = C + I - MPU - MPR, where C is non-oil consumption, I represents investment, and MPU denotes public sector imports. By definition, total domestic demand, DT, can be written as DT = C + I = Y + MPU + MPR. Total domestic demand exceeds production because of the net proceeds of oil revenues. Non-oil goods are produced either by the public sector or the private sector. But value added of the public sector equals, by definition, the public sector wage bill. Hence, non-oil private sector income stems from both private and public sector production, plus the amount of net oil revenues allocated to the private sector. Net oil revenues allocated to the private sector are a net transfer to the private non-oil sector. Non-oil private sector income can then be defined as D = Y + MPR. Assuming that savings equal investment for the non-oil private sector, expenditure equals income.
It would have been preferable to base the analysis on total imports, including non factor services, of the private sector. Data limitations necessitated the use of imports of goods. There are no source data available for nonfactor services; IMF staff estimates are made using a constant share of imports of goods. According to those estimates, nonfactor services of the private sector are relatively small.
The relative exchange rate is defined as RER = P$ * E * (1+T/100) / P, with E the formal exchange rate (naira per U.S. dollar) and T the average import tariff.
Since the first quarter of 1995, the formal exchange rate has been allowed to adjust to market forces. Consequently, the premium has been reduced strongly and virtually disappeared during 1997.
The enrollment ratio is an imperfect proxy for the stock of human capital. However, no data on the stock of human capital (or education of the work force) was available, and the construction of a stock variable would require information on the starting level, demographic trends, and the ‘depreciation rate’.
The parallel market premium is, over a long-enough time period, stationary. Although its inclusion does influence the estimated level for output capacity in specific periods, it does not alter the estimated long-run rate of technological progress.
Fxs is stationary at the 5% significance level, INF at the 8% level.
Although P is strictly speaking I(2) rather than I(1) at the 5 percent significance level, the test statistic (-2.5) is close to the critical value (of -2.9). Moreover, P was found to be I(I) when the unit root test was repeated over a longer period.
A stationary variable forms a co-integrating relationship on its own. Ignoring the stationarity of variables included in a co-integration analysis leads to an overestimation of the number of ‘proper’ co-integrating vectors, and could distort the identification process.
The imposition of the sum of the coefficients of d and p being 2.
The analysis was initially carried out for the period 1983-1996. A repetition of the exercise for the period 1981-1996 rendered an almost identical relationship, which was, however, more robust to tests. Therefore, the results for 1981-1996 were used in the dynamic model.
As discussed above, the unit root tests suggests that fxs is not unambiguously I(1); hence, fas could be the one stationary relationship suggested by the rank testing. However, because none of the other variables could be removed from the identified co-integrating vector and the vector was shown to be stationary, the identified relationship seems to be the one cointegrating vector suggested by the rank testing.
In 1990 constant prices.
INF, which was found to be co-integrated by itself (stationary) in the co-integration analysis for the money-price block, was also included in (lagged) level form. However, it was not found to be significant in any of the dynamic equations.
R2 does not allow for the mean since there is no constant in the equation.
As is the case for the dynamic equation for the real exchange rate, the distribution of the residuals is, strictly speaking, not normal (see attached table). However, this problem disappeared when the equation was estimated over 1987-96.