MOHAMMAD S HASAN The third issue is the assumption of constant and stable income velocity. Chen and Hou(1986) pointed out that income velocity is generally quite stable in China compared with income velocity in market economies, because of the role of the state banking system in handling the receipts and expenditures of industrial and commercial enterprises. However, the Chinese banking authorities indicated that income velocity, V, declined during the postreform period from 1980 to 1982 Chen and Hou, 1986). Chow(1987) found that the elasticity of the price level with respect to the ratio of money stock to real output, Mo/Y, is smaller than unity, which appears to suggest that velocity is not constant. Although the declining trend in income velocity, PY/M, in the 1980s certainly reduces the significance of the traditional quantity theory prediction that monetary growth is the sole cause of inflation, this trend does not necessarily eliminate the validity of the quantity theory in the long run. As long as a decline in income velocity is will be less than proportional to a change in money stoo oa not completely offset by an increase in money stock, change in the price level Finally, we do not claim that Chinese inflation can be explained in a classical paradigm, but rather that the empirical techniques we describe here will help to identify and estimate how much of the movement in price can be explained by monetary forces. Our general model is more acceptable and subject to less criticism than a pure quantity theory model 3. EMPIRICAL RESULTS To estimate the general inflation model, we identify six variables. They broad money (M,), the true price level (P), an output gap(g), wages(h agricultural productivity(AP), and a measure of industrial productivity (IP) Descriptions and sources for the variables and data used for estimation are given in the data appendix. The sample spanned the period from 1952 to 1993. Figure I plots the logarithms of M and P. A causal glance at the graphs clearly indicate that both series are trending together in the long run Santorum, 1987). Blejer et al. (1991), however, noted that currency demand has an unusual patte nges in expected inflation. The uncertainty surrounding the precise timing of the phases of the response pattern and difficulties in observing changes in expected inflation restrict the usefulness of currency as an intermediate target variable. Moreover, using M3 has the added advantage that the overall credit plan of the monetary authority is incorporated This study uses 42 annual observations for the following reasons. First, the impact and adjustment lags of various macroeconomic relations, such as money income and the money-price relationship, are too long for monthly quarterly observations to reflect the actual correlation between these macroeconomic variables. Although annual observations yield smaller degrees of freedom, the noisy effects associated with monthly or even quarterly observations tend to average out with annual data which better approximate the money-income or money-price relationship(Masih and Masih, 1995; Spencer, 1989). Second, Hakkio and Rush(1991)and Berg and Jayaneti(1993) contend that cointegration is a long run concept and, hence, requires a long span of data to give it much explanatory power. Finally, Chinese national income data are available only annually
The third issue is the assumption of constant and stable income velocity. Chen and Hou (1986) pointed out that income velocity is generally quite stable in China compared with income velocity in market economies, because of the role of the state banking system in handling the receipts and expenditures of industrial and commercial enterprises. However, the Chinese banking authorities indicated that income velocity, V, declined during the postreform period from 1980 to 1982 (Chen and Hou, 1986). Chow (1987) found that the elasticity of the price level with respect to the ratio of money stock to real output, M0/Y, is smaller than unity, which appears to suggest that velocity is not constant. Although the declining trend in income velocity, PY/M, in the 1980’s certainly reduces the significance of the traditional quantity theory prediction that monetary growth is the sole cause of inflation, this trend does not necessarily eliminate the validity of the quantity theory in the long run. As long as a decline in income velocity is not completely offset by an increase in money stock, change in the price level will be less than proportional to a change in money stock. Finally, we do not claim that Chinese inflation can be explained in a classical paradigm, but rather that the empirical techniques we describe here will help to identify and estimate how much of the movement in price can be explained by monetary forces. Our general model is more acceptable and subject to less criticism than a pure quantity theory model. 3. EMPIRICAL RESULTS To estimate the general inflation model, we identify six variables. They are broad money (M3), the true price level (P), an output gap ( g), wages (W), agricultural productivity (AP), and a measure of industrial productivity (IP). Descriptions and sources for the variables and data used for estimation are given in the data appendix. The sample spanned the period from 1952 to 1993.7 Figure 1 plots the logarithms of M and P. A causal glance at the graphs clearly indicates that both series are trending together in the long run. Santorum, 1987). Blejer et al. (1991), however, noted that currency demand has an unusual pattern of responses to changes in expected inflation. The uncertainty surrounding the precise timing of the phases of the response pattern and difficulties in observing changes in expected inflation restrict the usefulness of currency as an intermediate target variable. Moreover, using M3 has the added advantage that the overall credit plan of the monetary authority is incorporated. 7 This study uses 42 annual observations for the following reasons. First, the impact and adjustment lags of various macroeconomic relations, such as money income and the money–price relationship, are too long for monthly or even quarterly observations to reflect the actual correlation between these macroeconomic variables. Although annual observations yield smaller degrees of freedom, the noisy effects associated with monthly or even quarterly observations tend to average out with annual data which better approximate the money–income or money–price relationship (Masih and Masih, 1995; Spencer, 1989). Second, Hakkio and Rush (1991) and Berg and Jayaneti (1993) contend that cointegration is a long run concept and, hence, requires a long span of data to give it much explanatory power. Finally, Chinese national income data are available only annually. 674 MOHAMMAD S. HASAN
MONEY AND INFLATION N CHINA 7.0 12 65 6.0 5.0 6 4.5 556065707580859 FIG. 1 Log level of money stock(-)and prices(---). As a first step, the data were checked for stationarity using the Augmented Dickey-Fuller(ADF) test in each of the six variables. The number of augmentation terms in the ADF regression was determined by examining the significance of the final lag, up to three, and the serial correlation of residuals. The results of the ADF tests, as reported in Table 1, indicate that each of the six variables is nonstationary in level but not in first difference form. In the next step, the data series are further checked for the presence of cointegration using Hansen's(1990) procedure to see whether stochastic trends of these variables move together in the long run. Given the relatively small sample size, we apply the single equation estimation method. Under the single equation method, although statistical inference on the parameter of the integrating vector is possible, readers are urged to exercise some caution Hansen(1990)has noted that the power of the Engle-Granger(1987)tests falls substantially as the size of the system increases. Hansen(1990)has suggested a simple test of cointegration by applying a Cochrane-Orcutt procedure to correct for first-order serial correlation in the residuals of t cointegration equation. The procedure is modified to take into account possible structural break in the postreform period; e.g., Huang(1995)iden- tifies 1979 as the turning point of economic liberalization The results of the cointegration regression as well as the cointegration tests are reported in Table 2. The Dickey-Fuller t-statistics for both price and money
As a first step, the data were checked for stationarity using the Augmented Dickey–Fuller (ADF) test in each of the six variables. The number of augmentation terms in the ADF regression was determined by examining the significance of the final lag, up to three, and the serial correlation of residuals. The results of the ADF tests, as reported in Table 1, indicate that each of the six variables is nonstationary in level but not in first difference form. In the next step, the data series are further checked for the presence of cointegration using Hansen’s (1990) procedure to see whether stochastic trends of these variables move together in the long run. Given the relatively small sample size, we apply the single equation estimation method. Under the single equation method, although statistical inference on the parameter of the cointegrating vector is possible, readers are urged to exercise some caution. Hansen (1990) has noted that the power of the Engle–Granger (1987) tests falls substantially as the size of the system increases. Hansen (1990) has suggested a simple test of cointegration by applying a Cochrane–Orcutt procedure to correct for first-order serial correlation in the residuals of the cointegration equation. The procedure is modified to take into account a possible structural break in the postreform period; e.g., Huang (1995) identifies 1979 as the turning point of economic liberalization. The results of the cointegration regression as well as the cointegration tests are reported in Table 2. The Dickey–Fuller t-statistics for both price and money FIG. 1. Log level of money stock (—) and prices (- - -). MONEY AND INFLATION IN CHINA 675