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Money demand in China: The effect of economic reform
Money Demand in China: The Effect of Economic Reform DUO QIN
This paper reports an econometric study of China’s money demand relation, particularly for the period of the economic reforms. With respect to economic theory, two more factors representing transaction demand, additional to the commonly-used GDP or national income, are introduced: the ratio of the growth rates of total savings and loans, to capture the special feature of a centrally planned economy (CPE), and a monetization index to approximate the transitional feature of China’s economic reforms. With respect to econometric methods, the general+simple dynamic specification modelling approach is adopted to help search for relatively constant relations, using both quarterly data for the period 1978Ql-199144 and annual data for the period 1952-1991. The main findings are: (1) A relatively constant relation of money demand can be found, in spite of the considerable economic changes during the reforms; (2) The constancy is sustained by both the usual market variables, such as GDP and interest rate, and variables characterizing CPEs as well as the reform process; (3) The money demand relation will move towards a standard one in the long run, once the market system dominates; (4) Post-model forecast for 1992 renders satisfactory results; (5) The modelling results support further theoretical and policy studies of transitional economies from central planning towards market. (JEL E41,053, P21)
China’s banking system played a totally subordinate role in the material balancing mechanism of the centrally planned economy (CPE) prior to the economic reforms. The reforms have brought about gradual separation of fiscal and monetary policies, and brought to the limelight issues concerning the use of monetary instruments in macroeconomic management among Chinese economists and government policy makers. They have debated a great deal, during the last few years, upon the role of monetary policies in the rapid expansion of the aggregate demand. More specifically, their discussions are centered around the causality between money and inflation, and their views are roughly divided between those who regard the increase of money supply in excess of the economic growth as the main cause (e.g., Zhang, 1988; Chen, 1991) and those who disapprove of the existence of independent monetary policy and ascribe
Duo Qin Journal
l
Queen Mary and Westfield College, Department of Economics, Mile End Road, London El, 4NS, UK
of Asian Economics, Vol. 5. No. 2, 1994, pp. 253-271.
Copyright 0 1994 by JAI Press, Inc.
ISSN: 1049-0078
All rirhts of reoroduction in anv form reserved.
253
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the cause to the government credit policy, mainly the wage plan and the investment plan (e.g., Song, 1988; Jia, 1988; Dai 1992). To help resolve their debate and provide more specific answers to monetary policy issues, it is necessary to undertake careful empirical analyses of China’s aggregate money demand relation during the economic reforms. It seems a prevailing belief among Chinese economists that the reforms appear to have significantly altered and complicated the old money demand mechanism under a purely CPE regime (e.g., Ma, 1990). A list of possible causes are listed in Ma (1990), the key ones being: (1) the money demand function has been destabilized by both the marketization process and the rapid changes of government income policies and investment policies; and (2) the banking system is still subject to frequent interventions of various administrative mandates (pp. 798-799). But little empirical evidence is provided. On the other hand, relevant econometric studies appeared so far mainly estimated China’s money demand equations directly on the basis of the stylized quantity theory of money (e.g., Zhang, 1986; Chow, 1987; Li, 1990; Yi, 1993). Their findings are not focused on the questions of whether China’s money demand relation has undergone significant structural shifts since the reforms, and what the actual implications of their estimated relations are, particularly with respect to policy issues. Hence it is difficult to relate these empirical results to those problems in which the majority of Chinese economists are interested, as mentioned in the previous paragraph. The present study is an attempt to fill in the gap. It aims at finding an economically meaningful money demand equation with relatively constant coefficients, on the basis of the belief that any empirical results worthy of economic interpretation must exhibit certain degrees of parameter constancy, and that significant non-constancy found in econometric models serves as a strong indicator of missing variables. In other words, what those Chinese economists regard as changes of money demand mechanism is treated here as the case where certain factors inactive in the old mechanism start inserting effects. The model search is carried out under the generaljsimple dynamic specification approach. Two basic issues need to be clarified prior to the modelling: What do we use as the money variable? What do we mean by money demand? As for the first, narrow money (MO) is chosen mainly for the reason that MO has remained the principal control target of the People’s Bank so far. As for the second issue, the concept of ‘narrow’ demand for money used by Chinese economists (e.g., Ma, 1990) is adopted, which defines the demand solely from the viewpoint of the banking sector, separately from the fiscal budgetary sector of the government. The concept facilitates the analysis of different roles of government budgetary and monetary policies in relation to inflation, and the treatment of the aggregate investment behaviour of the sector of state-owned enterprises as a mainly non-market factor affecting the transaction demands for money.
Money Demand in China
1.
SPECIFYING
255
MONEY DEMAND
OF A MIXED ECONOMY
The core of various theoretical models of money demand comprises the following long-run equilibrium relation (e.g., Lucas, 1988; Goldfeld and Sichel, 1990): M’=f(k”,
R’)
(1)
where M: money Y: a flow variable representing transaction demands R: a rate variable representing portfolio demands e: denotes equilibrium. The relation is nested into certain stochastic and dynamic adjustment functions in econometric studies, to suffice measurement requirements. Over the last decade, the error-correction model (ECM) type has been opted for in many applied studies among all sorts of linear, dynamic adjustment functions: s
k
A.44,= CCXjW~-j + C PiAYf-i+ xy#?_i j=l
i=O
+ h(M+,M”) + E, E, -ZZV(O,O’)
(2)
i=O
where: ,r,s, k: are different end lags for the main reason that it accommodates static economic theory handily with non-stationary time series and expresses the long-run (theoretical) and short-run (dynamic) effects separately. Estimation of the ECM type of money demand relation with Chinese data has been pioneered by Burton and Ha (1990).’ They use national income as the flow variable and inflation as the rate variable. Two problems are immediately noticeable from their results: (1) all the long-run elasticities are too big to sustain equilibrium in the long run (similar results are found in other studies, e.g., Chow, 1987); (2) They have to use dummies for the sudden credit expansion and control in the mid 1980s. The problems indicate inadequate model representation of China’s money demand mechanism during the transitional period of China’s economic reforms. Possible ways to solve the problems lie in understanding the main special features of China’s money demand system prior to and since the reforms.’ A striking feature of the reforms is the marketization process, accompanied by rapid income increase of both individuals and enterprises. Of individuals, growth in farmer’s cash income is the most significant, especially during the early period of the reforms. As for enterprises, the increase occurs primarily with the rise of non-state owned enterprises, as well as with various micro policies allowing state-owned enterprises to retain certain profits. All these have boosted money demand. On the other hand, state economic planning still retains its vital role. A well-known feature of the planning is the subordination of the monetary system to the central government’s investment credit plan. Although the subordination has been weakened a great deal during the reforms, banking policies are still under the authority of the government’s major investment decisions. Finally, household savings have been quite sensitive to
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changes in price levels and interest rates, since the average income level is still low with respect to the average living cost in China. The last feature implies that inflation alone is not sufficient to represent portfolio demands. Thus the real interest rate is used in the present study, defined as the oneyear bank deposit rate net of inflation, which is measured by the rate of retail price indices following the common practice. The first two features are essentially of transaction demands. Here, it is important to note that either national income or GDP, the commonly-used measure for transaction demands, excludes sales and transfers of intermediate goods, as well as the effect of changes in the proportion of income distribution between households and the state (cf. Goldfeld and Sichel, 1990). It is precisely in these two areas that changes in transaction demands have been significant in China embodying the first two features. Hence, we ought to introduce more transaction variables than a single national income or GDP. Yi (1993) introduces the ratio of urban population to total population as a proxy of monetization process. A closer proxy of a unit-free index is composed here with respect to data availability, using the average of two ratios: the price ratio of agricultural to industrial output deflators, and the output ratio of the non-state-owned industry to the whole industry.3 The first ratio reflects the trend of rising rural money incomes through agricultural price reforms. The second ratio embodies the degree of growth of the non-state-owned enterprises. The index is expected to flatten out gradually as the market economy dominates, while isolating the normal transaction demands characterized by GDP in line with standard money demand theory both during and after the reforms. As for the effect of the government investment plan, the obvious candidates are state fixed investment, government budget deficit, the total credits issued.4 But direct inclusion of any of these variables would cause collinearity with the flow variable Y in the model. In other words, transactions of intermediate goods proportional to transactions of final goods should largely be captured by GDP already, and the likely missing effect is changes in the part of transactions of intermediate goods which are significantly out of proportion to GDP movement. In a CPE, such changes are mainly induced by the planning mechanism. Therefore, what is needed is a rate variable which approximates the oscillatory impact of government investment plans. Accordingly, the annual rates of the ratio between the total bank savings and bank loans is chosen (i.e., the ratio between the annual growth rate of savings and that of 10ans).~ To summarize, the following variables are used to model real money demand in China in the present study: real interest rates, R, and three elements for transaction demands: GDP, annual rate of savings/loan ratio, RSL, and a monetization index, IM.6 The money relation thus specified is expected to have the following properties: (1) The effect of RSL should be oscillating and opposite to the directions of changes in M; (2) The effect of R be negative; and (3) The transaction elasticity of GDP be proportional to M. The model search aims at finding out: (1) whether such a relation can be estimated from data; (2) whether the relation keeps relatively constant over time once estimated; and (3) whether it has any equilibrium power in the fluctuations of money movement.
Money Demand in China
257
TABLE 1. Unit Root Tests (critical value: -2.9) LnM DF ,ADF(1)
-0.859 -0.841
DF
ALnM -7.592
,4DF( 1)
-11.48
3.
LnGDP
R
IM
LnRSL
-1.57 -1.368 ALnGDP
-1.758 -3.272 AR
-1.133 -1.132 AIM
-1.959 -2.807 ARSL
-8.40
4.067
-8.004
-5.111
-12.74
-4.051
-7.432
A.191
MODELLING
MONEY DEMAND RELATION
The model search starts from a quarterly data set covering 197841-9144. Definitions and sources of variables are described in the appendix.7 Figure 1 plots real MO and real GDP, Figure 2 plots real interest rate, R, nominal rate IR and inflation PR, Figure 3 plots RLS and Figure 4 plots IM. Since these variables seem to exhibit non-stationarity, unit-root tests have been conducted, and the resulting Dickey-Fuller (DF) and augmented Dickey-Fuller (ADF) statistics are reported in Table 1. However, because the sample size of 56 is quite small, these unit root tests may not have much power. The uncertainty over the exact degrees of non-stationarity of individual variables makes it difficult to apply Johansen’s system estimation procedure 40QY
360-
320-
280-
240-
16B-
I C..,, ,. *. ,' ,' '__.,I .__ # , ,
12l3-
88-
,. ._,I
. ’ , \,' >'../' ' ,'
;*._. , 4a________*-._--
I
,-_*_-
*-____.----
e-'...".'.'.'..'.."".."".".'."'. 1980 1982
Figure I.
_- --___
1984
1986
1988
Real MO and GDP in 1980 Prices
1990
1992
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~...... ___ ,. F-----h .‘-..-.---.-~-‘-‘-.--“-‘ . :--.-.f.-...,,: IR
,.
.
..:
‘.
\
._.__
‘.._
___.’
__I
R
‘\
I
,*
.
;.*
<
-161
.'.
* 1982
'.
.' 1983
'.
.
1.. 1985
Figure 2.
..I
.
1987
.
.
I 1989
.
.
~.
I,
.
1991
R, IR and PR
9
1987
Figure 3.
LnRSL
. 1993
2
1982
.
1992
.
Money Demand in China
259
1982
1987 Figure 4.
1992
IM
(1988), since the procedure is quite sensitive to changes in the numbers of lags, variables and their degrees of non-stationarity. Moreover, the small sample size is likely to give biased estimates of the static regression (long-run) coefficients by the Engle-Granger (1987) two-step estimation method. In light with the situation, the following practical strategy is adopted: First, Hendry’s genera&simple reduction approach (1991) is used (i.e., the modelling starts from a log-linear, autoregressive distributed-lag (ADL) relation with LnGDP,, R,, IM,, LnRSL, in six lags and seasonal dummies, and then gradually reduces and reparameterizes into a parsimonious errorcorrection model. Johansen’s procedure is used afterwards to help check weak exogeneity of the obtained single-equation results, since the above procedure conditions crucially upon weak exogeneity. Table 2 presents the final results of the modelling search by OLS method. Figure 5 show,s the fit.8 Notice that a unit-elasticity restriction on GDP is imposed in the long-run relation imbedded in the error-correction term in accordance primarily with economics. The restriction is not rejected by data during the model reduction. It actually helps to isolate the monetization effect on transaction demands because there is a certain degree of collinearity between IM and GDP. Another point worth noting is that LnRSL, does not enter the EC term and only ALnRSL, is found significant as a short-run factor. To check whether the long-run equilibrium in the EC term is unique in the sense that GDP, R and IM are weakly exogenous with respect to M, we apply Johansen’s
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TABLE 2. c
G-I
-.258 (.102) Notes:
-.123 (.048)
OF ASIAN ECONOMICS,
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OLS Estimates of an EC Model of ALnM, (1978-1991) &LnGDP,
AL~(GDP/IV),~
AblRSL,
wt
Q2
,504 (.028)
,271 (.038)
-.414 (.143)
-.004 (.0018)
,100 (.014)
The figures in brackets are standard errors; A denotes first difference; A, denotes two-lag difference; Q2 is a seasonal dummy of the second quarter; R* = ,925; F(6,44) = 91.05; s = ,029s; DW = 2.36; Normality x’(2) = 0.7M); Residual autocorrelation F(4,40) = 2.475, ARCH x2(4) = 7.875 (lags IA); Heteroscedastic errors x2 (11) = 15.837; RESET F(l,43) = 0.987; Variance instability: 0.098; Joint instability: I.170 EC, = LnM, - LnGDP, + 0.01 R, -
I .4 IM,
(1988) procedure to the relation for both the cases of unrestricted and restricted GDP coefficient. The results are summarized in Tables 3a and 3b. It is seen from the tables that likelihood ratio tests for the both cases suggest the existence of at least one cointegrating vector, and that the coefficient estimates are quite sensitive to both the restriction and the number of lags. A more powerful check for exogeneity is parameter constancy (cf. Ericsson 1991). Figures 6-9 are some of the recursive results based on Table 2. The coefficients appear remarkably constant within their 95 percent confidence intervals through the sample
Figure 5.
Fitted and Actual ALnM, of the Quarterly Model
Money Demand in China
TABLE 3a.
261
COINTEGRATION
RESULTS BY JOHANSEN
PROCEDURE
(LnW, LnGDP,, R,, IM) H,: r = no. of cointegrating vectors (CV): r13 rS2
I LAGS
4LAGS
8LAGS
Critical Values 5%
0.126
1.037
0.421
9.094
6.518
20.168
3.491
rS1
11.49
25.82
r=O
28.67
127.3
EC coefficients
28.24
73.82
-0.162
35.068
107.0
.034
53.347
-.075
Estimated CVs for Ln M: LnM
1.0
Ln GDP
1.0
-2.01
1.0
3.453
R
0.002
IM
0.378
-1.32
0.080
0.032
-10.5
-0.64
period, especially in view of the rapidly changing economy during the reforms, and particularly the shock occurred in the summer of 1989. To investigate further the issue of constancy, the same modelling procedure is repeated on an annual data set for the period 1952-199 1. As GDP was not recorded before the reforms in China, national income (1980 price), NI, is used here instead. IM is set to constant prior to 1978 and to the annual average of the quarterly IM thereafter.’ The results of the annual model are summarized in Tables 4-5 and Figures 10-14. Comparing the annual model with the quarterly model results, we notice two particularly remarkable features: (1) The coefficient estimate of the EC term in the annual model (-0.557) is about four times of that of the quarterly model (-0.123) with respect to their confidence intervals; (2) the two long-run relations imbedded in the
TABLE 3b. -
Cointegration
Results by Johansen Procedure
(LnWLnGDP,,
H,., r = no. of cointegrating vectors (CV): rS2 0.60 rll
4.032
R,, IW I LAGS 1.084 5.861
4LAGS 0.001 8.939
rS0
33.08
31.72
27.37
EC coefficients Estimated CVs for Ln M:
-0.594
-0.05
-0.011
Ln M/GDP
1.0
1.0
R
0.004
0.094
IM
-1.63
-1.12
1.0 -0.21 1.141
8LAGS
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-.88
-.2c__---..________----
,.---__ ,*': : I' I
-.24-
;._.--
,1---.., ,-
1 ' < ' I ' q,_,'
-.32-
-.36
.-___-r
>*, ,' I< _ , \.-I
_c-r-._ ', ..I
-.28-
..
".
""I'. 1984
Figure 6.
1985
'I". 1986
RLS Estimated Coefficient
I'. 1987
""'I 1988
'
”
1989
of EC,_, (with its 95% confidence
199e
"'I'. 1991
.' 1992
interval as the dotted lines)
Figure 7. RLS Estimated Coefficient of ALnRSL, (with its 95% confidence interval as the dotted lines)
Money Demand in China
263
-.BQd-
.. .
-.ee8-
,.--.
: -.a1
-
-.lBL2L.
Figure 8.
_._4----,
I._._-.
‘.
.
’
1984
,
,’
.
,
,
_,----__._-___~_-----
,‘\ ,, <*
‘---.____--a
a. _-._,’
I
‘________-
.I f
. ,,’
.a
’
1985
.
.
.
’
1986
RLS Estimated Coefficient
.
.
‘.
.I
1987
.
1988
.
1
1989
of AR, (with its 95% confidence
.
I
1990
.
I
1991
.
.
,
1992
interval as the dotted lines)
1. Ia l-
.95 l-
____---_________---____~~~__________~__________________~~-~~~~~-----~_~~_~____~_~~___. t 52 critical
.sl-
value
.81
.72
.63 I-
.54 l-
.36
.27
.18
I
1984
.
1985
Figure 9.
I
1986
.
.
I
1987
.
.
I
1988
.
.
I
1989
,
1998
Forecasting Chow Tests (dotted line is 5% critical value)
,
1991
.
,
1992
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-.18
OF ASIAN ECONOMICS,
:
;
‘< -.24
(5)2,1994
".
'I 1968
"'
'I 1965
"""".""". 1970
“’
”
1975
1980
I’
".
1985
199e
'
“.
1995
Figure IO. Fitted and Actual ALnM, of the Annual Model
-.4-
-.45-
_--.-__ ,' __-.______,.~~-~..-_.._ I. _,,' _ -._____..~" *.*A...-..,____\__ ,_/_..I~'~~~--..._.__,..~~-................_.~
-.5-
-.55-
-.6\
-.95[
’
1
Figure II.
’
’
1975
RLS Estimated Coefficient
1980
of EC,,
’
’
1985
(with its 95% confidence
.
.
1990
’
.
’
.
’
1995
interval as the dotted lines)
Money Demand in China
Figure 12.
265
RIS Estimated Coefficient of AzLnRSL, (with its 95% confidence interval as the dotted lines)
-.erz-
. ,_.__-_
,_,______..._~.~~....._~_~~........~ -.l315-
__,* ..
. . . . . _.
,’
..
,,*’
..*._..
,I’
*. ,,’ ..\,_.,_
,.*. ,’ -.818_,_,..’ :
,,’
:
.,_-.-.-
-.821-,.-
-.a24
,/’
Il....
I....#....,..,.,*. 1975
Figure 13.
,., 1980
RLS Estimated Coefficient
1985
of AR, (with its 95% confidence
1990
1995
interval as the dotted lines)
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1.04r
1.... . . .___. ...... ..... .........._.... ........_._.._.._
... .. ..___________._.......................
‘-“t””
.96 c
5X critical
level
.ea .8 .72
t
.64
-56t .4a
.4
.32i
:::?._,..-----._-.eal..
.
.
1.
.
1975
LtlM
DF ADF(1)
-1.710 (.341) Nok-
. 1985
1
.
1 1990
J 1995
Unit Root Tests (critical value: -2.9) LnNI
R
IM
LnRSL -3.685
1.142
0.372
-2.926
4.941
1.038 ALnM
0.064 ALnNI
-2.997
1.75
-2.527
AR
AIM
A2LnRSL
-5.283 -3.622
4.25 4.866
-6.154 -5.654
-2.93 -1.655
4.092 11.698
TABLE 5. C
.
Forecasting Chow Tests (dotted line is 5% critical value)
TABLE 4.
DF
1.. 1980
Figure 14.
ADF(1)
.
OLS Estimates of an EC Model of ALnM, (19521991)
ECt-I
-.557 (.076)
MLnNI,
AQf,
A&nRSL,
A&
,216 (.065)
2.250 (.324)
-.675 (.260)
-.0076 (.0029)
M denotes taking tirst difference twice; RZ = ,817; s = ,072; DW = 2. II; Nomality x2(2) = 0.372; Residual autocorrelation F(2, 29) = 1.089, ARCH x’(l) = 0.414 (lag I); Heteroscedastic errors x*(10) = 4.284; RESET F(1.30) = 0.537; Variance mstability: 0.165; Joint instability: 0.987 EC, = LnM,-LnY,
+ O.OlR, -1.65 IM,
Money Demand in China
TABLE 6.
267
Post-Model Prediction of 1992 by the Quarterly Model
Actual MO (billion) 311.7 315.6 355.9 433.6
Predicted MO
Prediction Errors %
316.7 310.6
1.6
347.9
1.6 2.2
413.0
4.7
EC terms are virtually identical.” Moreover, the annual model appears similarly constant (see Figures 11-14). However, it is difficult to compare the short-run dynamic effects of the two models, because the units of the lag differences are incompatible. Finally, post-model prediction is made for the year 1992 using the quarterly model, and prediction errors are found below five percent (see Table 6), in spite of the fact that the annual growth rate of MO shot up from around 20 percent in 1991 to above 36 percent in 1992. The consistency between the quarterly and annual models and the relatively satisfactory econometric results lay the ground for the following tentative arguments.
4.
ECONOMIC
IMPLICATIONS
OF MODEL RESULTS
The first argument is that China’s economic reforms do not seem to have induced changes in the standard propensities of the transaction and portfolio money demands represented by GDP, or NI, and real interest rate. The fact that simple, standard money demand models have suffered from ‘structural breaks’ since the reforms does not necessarily indicate changes in these propensities. More often than not, it indicates insufficient model formulation in the sense that certain important variables representing the specific features of the economy or the regime are missing. The present modelling search has identified two of such variables, IM and RSL, and established a relatively constant model. These two variables deserve further explanation since they are non-standard money demand variables and they play a crucial role in the present model. First, let us look at IM. Although a proxy, the variable has segregated the trend of marketization from the normal transaction demands represented by GDP or NI. The segregation helps to clarify the confusion that some of Chinese economists have between excess money supply and faster MO growth than GDP growth due to the marketization effect on money demand. It also helps to maintain the economic sense of the long-run relation imbedded in the ECM. Take for example the long-run relation under equilibrium from the quarterly model: LnM = Lnc + 1.65ZM + LnGDP - O.OlR or M = cZM’.~~~.“GDP
(1)
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where c is a constant. Since the values of IM are smaller than one, its exponential by 1.65 will not cause the growth of MO accelerate faster than the growth of GDP, unlike those models which have produced estimates of LnGDP coefficient greater than one. Furthermore, IM will converge to a constant once the monetization process ends, reducing the long-run relation under equilibrium to a simple and standard one involving only GDP and R: M = kR4.”
GDP
(2)
Next, let us look at RSL. As seen from Tables 2 and 5, only ALnRSL is found significant and its negative coefficient estimates comply with the prior expectation. which represents the acceleration/deceleration rates of the proportion/disproportion of the growth rates between savings and loans. Since ‘investment hunger’ almost always exists and the Chinese government has maintained low and stable income policies and fixed interest rate policies, significant fluctuations in ALnRSL have been induced much more often by drastic credit plans than by saving fluctuations. However, expansionary investment decisions alone is insufficient to stimulate money demand unless the expansion exceeds significantly the economy’s saving potential. Now the question arises that this variable seems to be a supply-side factor from the viewpoint of a standard market economy. But as mentioned earlier, an important feature of a CPE is that its government plays not only the role of the economic regulator, but also the role of the chief representative of the state-owned sector, which forms the major part of the economy. If we analyze such an economy in the form of a real section versus a financial section, government credit decisions then reflect the investment behavior of the state-owned sector, and therefore should be considered as affecting transaction demands for money through fluctuations in the sales of intermediate goods. From this perspective, the significance of ALnRSL appears to correspond to Komai’s conjecture that ‘soft budget constraint syndrome may show up in the formation of excess demand’ (1986), since excessive investment expansions exhibit the syndrome of ‘soft budget constraint’ at the macro level. Notice that LnRSL does not enter the long-run relation, suggesting that ‘soft budget’ in the form of excessive investments at the macro level is transient and bound by the long-run market equilibrium condition. The model result with RSL also suggests that modelling government investment behavior is more fundamental than modelling money demand, in the sense that money is ‘passive’ not only with respect to the usual demand factors of a market economy, but also with respect to special factors of a CPE. Finally, let us look at R briefly. The interest rate variable is absent in most previous econometric studies of China’s money demand, except perhaps only (Portes and Santorum, 1987). Here, the significance of both the long-run and short-run coefficients of R indicates that the cost of holding money is not a negligible factor in studying China’s aggregate money demand behavior. Since the government has implemented fixed interest rates policies, much of the dynamic effect of AZ? is due to inflation fluctuations, as shown in Figure 2. ‘I The relatively constant coefficient estimates found for this variable as well as the long-run cointegration tests suggest that inflation is
Money Demand in China
269
weakly exogenous for money, since a regression with constant coefficients and non-constant marginal distribution of exogenous variables cannot produce a constantcoefficient inverse regression (cf. Ericsson, 1991). Although weak exogeneity does not carry the full capacity of determining ‘causality’ in the common sense, the present finding with R is supportive to those Chinese economists who argued that inflation was not fundamentally caused by active money expansion during the reforms (e.g., Dai, 1992). More conclusive arguments entail further econometric studies of the price determination mechanism in China.
APPENDIX (A) Quarterly data 1978(l)-1992(4) MO GDP R
IM
RSL Source:
money in circulation deflated by GDP deflator (billion yuan) gross domestic products in 1980 prices (billion yuan) IR - PR (IR: interest rate of one-year deposit savings; PR: inflation rate = annual rate of overall retail price indices) 0.5 RP + 0.5 RO (RP: ratio of agricultural output deflator to industrial output deflator; RO: ratio of non-state-owned industrial output to total industrial output) annual rates of the ratio of total bank savings to total loans = &Ln(SIL)
the State Statistical Bureau of China
(B) Annual data (19521991) MO Y R
money in circulation deflated by national income deflator (billion yuan) national income in 1980 prices (billion yuan) interest rate of one-year deposit savings - inflation rate = 0.5 for 1952-77, the annual average of the above IM for 1978-91 annual rates of the ratio of total savings to total loans = ALn(SIL)
IM RSL Sources:Statistical
Yearbook of China (various years); interest rates for 1952-77 from Byrd (1983)
Acknowledgment: An earlier version of this paper has come out in the Discussion Paper series of the Department of Economics, Queen Mary and Westfield College (No. 281) under the joint authorship with Tongsan WANG, Mingwu ZHOU and Guoqiang LIU. I sincerely acknowledge the Ford Foundation research grant 910-1107, which enabled me to work on modelling the Chinese economy with the help of WANG, ZHOU and LIIJ. The present study in money demand is an extension of the modelling exercise and has been carried out solely by the current author. Nevertheless, I am very grateful to them three. Thanks should also go to Lizuo JIN, Sheng HUA, Weiying ZHANG, Guy LIU, as well as many other Chinese economists, for their insightful views on China’s economy, and to Chenggang XU, Christopher Gilbert, members of the Economics Department of QWM and three anonymous referees for their helpful comments on the earlier versions of this paper.
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NOTES 1. Burton and Ha (1990) estimated various money demand, such as household demand, enterprise demand, savings deposits, as well as MO, Ml, and M2, with quarterly data for the period 1983-1988. 2. For more detailed description, see Portes and Santorum, 1987, Yi, 1991; 1992 and Ma, 1990. 3. Modelling with either ratio as the index has been experimented. The results are slightly worse than the combined index. Equal weights for the two ratios in the combined index are imposed solely for simplicity. 4. In fact, the aggregate level of credit funds has been used by the central bank in China as one of its main intermediary targets of the monetary policy, cf. Ma (1990). 5. The ratio is also an important index that the Chinese government monitors cf. Ma. 6. Real money is calculated from nominal money using the same deflator of real GDP to keep the measurement of these two flow variables as consistent as possible. 7. The issue of data accuracy is ignored here on the assumption that there are no significant changes in the data collection process as far as accuracy is concerned, since what all applied economic studies do is to explain and predict given data. 8. 2SLS and IV methods for the variable RSL, have been experimented. The results do not differ significantly from Table 2, and are not reported here. 9. The constant is set at 0.5 on the simple assumptions that the price ratio was one and that the industrial output ratio was zero. 10. The only difference lies with the coefficient estimates of IM,, which trends with GDP or NI since the reform. The following simple regression for the period 1978Ql-199144 shows: GbP=
1.24NI=,E>&
Hence the annual estimate of the coefficient of IM, ought to be slightly larger than the quarterly estimate. 11. The alternative of estimating separately the inflation effect and nominal interest rate effect has also been experimented during the modelling. The resulting estimates do not contradict the composition of a single variable of real interest rate.
REFERENCES Burton, David and Ha, Jibing. 1990. “Economic Reform and the Demand for Money in China.” International Monetary Fund Working Paper, 42. Byrd, William. 1983. China’s Financial System. Boulder, CO: Westview Press. Chen, Shiqiang. 1991. “Assessment of current macroeconomy and some policy suggestions.” China Finance (in Chinese), I: 21-23. Chow, Gregory C. 1987. “Money and price level determination in China.” Journal of Comparative Economics, 11: 319-333. Dai, Yuan-then. 1992. Nota Free Choice: InflationandStrategy During the Reform (in Chinese). Beijing: Engle, R.F. and Granger, C.W.J. 1987,“Cointegration and error correction: representation, estimation and testing.” Econometrica, 55: 251-276. Ericsson, Neil R. 1991. “Cointegration, exogeneity, and policy analysis: an overview.” International Finance Discussion Papers, No.415, Board of Governors of the Federal Reserve System of USA. Goldfeld, Stephen M. and Sichel, Daniel E. 1990. “The demand for money.” Pp. 299-356 in B.M. Friedman and F. H. Hahn (eds), HandbookofMonetary Economics, Vol. 1. Amsterdam: North-Holland. Hendry, David. 1991. “Lectures on Econometric Methodology,” a draft. Oxford University. Jia, Kang. 1988. “Tight money, or reduce investment scale.” Economic Research(in Chinese), 5: 10-14. Johansen, S. 1988. “Statistical analysis of cointegration vectors.” JoournaZof Economic Dynamics and Control, 12: 231-254. Komai, Janos. 1986. “The soft budget constraint.” Kyklos, 39: 3-30.
Money Demand in China
271
Li, Bo-xi (ed) 1990. Economic Development and Model of China (in Chinese). Beijing: Publisher? Lucas, Robert E. 1988. “Money demand in the United States: A Quantitative Review.” Pp. 137-168 in Carnegie-Rochester Conference Series on Public Policy, Vol. 29. Ma, Hong et al. (eds) 1990. Encyclopedia ofFinancial Knowledge (in Chinese), Vol. 1. Beijing: Publisher? Portes, Richard and Santorum, Anita. 1987. “Money and the Consumption Goods Market in China.” Journal of Comparative Economics, I I: 354-37 1. Song, Guoqing. 1988. “Monetary policy under the dual price system.” Economic Research (in Chinese), 5: 18-25. Yi, Gang. 1991. ‘The monetization process in China during the economic reform.” China Economic Review, 2: 75-95. -. 1992. “The money supply mechanism and monetary policy in China.” Journal of Asian Economics, 3: 217-238. -. 1993. “Toward estimating the demand for money in China.” Economics of Planning, 26: 243-270. Zhang, Jue. 1986. “Quantitative analysis of money in circulation.” In Wu, J-P. (ed), Applications of Econometric Methods in China (in Chinese). Zhang, Shuguang. 1988. “Preliminary analysis of China’s inflation problem.” China: Development and Reform (in Chinese), 3: 30-39.
Received July 1993; Revised March 1994
This paper reports an econometric study of China’s money demand relation, particularly for the period of the economic reforms. With respect to economic theory, two more factors representing transaction demand, additional to the commonly-used GDP or national income, are introduced: the ratio of the growth rates of total savings and loans, to capture the special feature of a centrally planned economy (CPE), and a monetization index to approximate the transitional feature of China’s economic reforms. With respect to econometric methods, the general+simple dynamic specification modelling approach is adopted to help search for relatively constant relations, using both quarterly data for the period 1978Ql-199144 and annual data for the period 1952-1991. The main findings are: (1) A relatively constant relation of money demand can be found, in spite of the considerable economic changes during the reforms; (2) The constancy is sustained by both the usual market variables, such as GDP and interest rate, and variables characterizing CPEs as well as the reform process; (3) The money demand relation will move towards a standard one in the long run, once the market system dominates; (4) Post-model forecast for 1992 renders satisfactory results; (5) The modelling results support further theoretical and policy studies of transitional economies from central planning towards market. (JEL E41,053, P21)
China’s banking system played a totally subordinate role in the material balancing mechanism of the centrally planned economy (CPE) prior to the economic reforms. The reforms have brought about gradual separation of fiscal and monetary policies, and brought to the limelight issues concerning the use of monetary instruments in macroeconomic management among Chinese economists and government policy makers. They have debated a great deal, during the last few years, upon the role of monetary policies in the rapid expansion of the aggregate demand. More specifically, their discussions are centered around the causality between money and inflation, and their views are roughly divided between those who regard the increase of money supply in excess of the economic growth as the main cause (e.g., Zhang, 1988; Chen, 1991) and those who disapprove of the existence of independent monetary policy and ascribe
Duo Qin Journal
l
Queen Mary and Westfield College, Department of Economics, Mile End Road, London El, 4NS, UK
of Asian Economics, Vol. 5. No. 2, 1994, pp. 253-271.
Copyright 0 1994 by JAI Press, Inc.
ISSN: 1049-0078
All rirhts of reoroduction in anv form reserved.
253
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the cause to the government credit policy, mainly the wage plan and the investment plan (e.g., Song, 1988; Jia, 1988; Dai 1992). To help resolve their debate and provide more specific answers to monetary policy issues, it is necessary to undertake careful empirical analyses of China’s aggregate money demand relation during the economic reforms. It seems a prevailing belief among Chinese economists that the reforms appear to have significantly altered and complicated the old money demand mechanism under a purely CPE regime (e.g., Ma, 1990). A list of possible causes are listed in Ma (1990), the key ones being: (1) the money demand function has been destabilized by both the marketization process and the rapid changes of government income policies and investment policies; and (2) the banking system is still subject to frequent interventions of various administrative mandates (pp. 798-799). But little empirical evidence is provided. On the other hand, relevant econometric studies appeared so far mainly estimated China’s money demand equations directly on the basis of the stylized quantity theory of money (e.g., Zhang, 1986; Chow, 1987; Li, 1990; Yi, 1993). Their findings are not focused on the questions of whether China’s money demand relation has undergone significant structural shifts since the reforms, and what the actual implications of their estimated relations are, particularly with respect to policy issues. Hence it is difficult to relate these empirical results to those problems in which the majority of Chinese economists are interested, as mentioned in the previous paragraph. The present study is an attempt to fill in the gap. It aims at finding an economically meaningful money demand equation with relatively constant coefficients, on the basis of the belief that any empirical results worthy of economic interpretation must exhibit certain degrees of parameter constancy, and that significant non-constancy found in econometric models serves as a strong indicator of missing variables. In other words, what those Chinese economists regard as changes of money demand mechanism is treated here as the case where certain factors inactive in the old mechanism start inserting effects. The model search is carried out under the generaljsimple dynamic specification approach. Two basic issues need to be clarified prior to the modelling: What do we use as the money variable? What do we mean by money demand? As for the first, narrow money (MO) is chosen mainly for the reason that MO has remained the principal control target of the People’s Bank so far. As for the second issue, the concept of ‘narrow’ demand for money used by Chinese economists (e.g., Ma, 1990) is adopted, which defines the demand solely from the viewpoint of the banking sector, separately from the fiscal budgetary sector of the government. The concept facilitates the analysis of different roles of government budgetary and monetary policies in relation to inflation, and the treatment of the aggregate investment behaviour of the sector of state-owned enterprises as a mainly non-market factor affecting the transaction demands for money.
Money Demand in China
1.
SPECIFYING
255
MONEY DEMAND
OF A MIXED ECONOMY
The core of various theoretical models of money demand comprises the following long-run equilibrium relation (e.g., Lucas, 1988; Goldfeld and Sichel, 1990): M’=f(k”,
R’)
(1)
where M: money Y: a flow variable representing transaction demands R: a rate variable representing portfolio demands e: denotes equilibrium. The relation is nested into certain stochastic and dynamic adjustment functions in econometric studies, to suffice measurement requirements. Over the last decade, the error-correction model (ECM) type has been opted for in many applied studies among all sorts of linear, dynamic adjustment functions: s
k
A.44,= CCXjW~-j + C PiAYf-i+ xy#?_i j=l
i=O
+ h(M+,M”) + E, E, -ZZV(O,O’)
(2)
i=O
where: ,r,s, k: are different end lags for the main reason that it accommodates static economic theory handily with non-stationary time series and expresses the long-run (theoretical) and short-run (dynamic) effects separately. Estimation of the ECM type of money demand relation with Chinese data has been pioneered by Burton and Ha (1990).’ They use national income as the flow variable and inflation as the rate variable. Two problems are immediately noticeable from their results: (1) all the long-run elasticities are too big to sustain equilibrium in the long run (similar results are found in other studies, e.g., Chow, 1987); (2) They have to use dummies for the sudden credit expansion and control in the mid 1980s. The problems indicate inadequate model representation of China’s money demand mechanism during the transitional period of China’s economic reforms. Possible ways to solve the problems lie in understanding the main special features of China’s money demand system prior to and since the reforms.’ A striking feature of the reforms is the marketization process, accompanied by rapid income increase of both individuals and enterprises. Of individuals, growth in farmer’s cash income is the most significant, especially during the early period of the reforms. As for enterprises, the increase occurs primarily with the rise of non-state owned enterprises, as well as with various micro policies allowing state-owned enterprises to retain certain profits. All these have boosted money demand. On the other hand, state economic planning still retains its vital role. A well-known feature of the planning is the subordination of the monetary system to the central government’s investment credit plan. Although the subordination has been weakened a great deal during the reforms, banking policies are still under the authority of the government’s major investment decisions. Finally, household savings have been quite sensitive to
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changes in price levels and interest rates, since the average income level is still low with respect to the average living cost in China. The last feature implies that inflation alone is not sufficient to represent portfolio demands. Thus the real interest rate is used in the present study, defined as the oneyear bank deposit rate net of inflation, which is measured by the rate of retail price indices following the common practice. The first two features are essentially of transaction demands. Here, it is important to note that either national income or GDP, the commonly-used measure for transaction demands, excludes sales and transfers of intermediate goods, as well as the effect of changes in the proportion of income distribution between households and the state (cf. Goldfeld and Sichel, 1990). It is precisely in these two areas that changes in transaction demands have been significant in China embodying the first two features. Hence, we ought to introduce more transaction variables than a single national income or GDP. Yi (1993) introduces the ratio of urban population to total population as a proxy of monetization process. A closer proxy of a unit-free index is composed here with respect to data availability, using the average of two ratios: the price ratio of agricultural to industrial output deflators, and the output ratio of the non-state-owned industry to the whole industry.3 The first ratio reflects the trend of rising rural money incomes through agricultural price reforms. The second ratio embodies the degree of growth of the non-state-owned enterprises. The index is expected to flatten out gradually as the market economy dominates, while isolating the normal transaction demands characterized by GDP in line with standard money demand theory both during and after the reforms. As for the effect of the government investment plan, the obvious candidates are state fixed investment, government budget deficit, the total credits issued.4 But direct inclusion of any of these variables would cause collinearity with the flow variable Y in the model. In other words, transactions of intermediate goods proportional to transactions of final goods should largely be captured by GDP already, and the likely missing effect is changes in the part of transactions of intermediate goods which are significantly out of proportion to GDP movement. In a CPE, such changes are mainly induced by the planning mechanism. Therefore, what is needed is a rate variable which approximates the oscillatory impact of government investment plans. Accordingly, the annual rates of the ratio between the total bank savings and bank loans is chosen (i.e., the ratio between the annual growth rate of savings and that of 10ans).~ To summarize, the following variables are used to model real money demand in China in the present study: real interest rates, R, and three elements for transaction demands: GDP, annual rate of savings/loan ratio, RSL, and a monetization index, IM.6 The money relation thus specified is expected to have the following properties: (1) The effect of RSL should be oscillating and opposite to the directions of changes in M; (2) The effect of R be negative; and (3) The transaction elasticity of GDP be proportional to M. The model search aims at finding out: (1) whether such a relation can be estimated from data; (2) whether the relation keeps relatively constant over time once estimated; and (3) whether it has any equilibrium power in the fluctuations of money movement.
Money Demand in China
257
TABLE 1. Unit Root Tests (critical value: -2.9) LnM DF ,ADF(1)
-0.859 -0.841
DF
ALnM -7.592
,4DF( 1)
-11.48
3.
LnGDP
R
IM
LnRSL
-1.57 -1.368 ALnGDP
-1.758 -3.272 AR
-1.133 -1.132 AIM
-1.959 -2.807 ARSL
-8.40
4.067
-8.004
-5.111
-12.74
-4.051
-7.432
A.191
MODELLING
MONEY DEMAND RELATION
The model search starts from a quarterly data set covering 197841-9144. Definitions and sources of variables are described in the appendix.7 Figure 1 plots real MO and real GDP, Figure 2 plots real interest rate, R, nominal rate IR and inflation PR, Figure 3 plots RLS and Figure 4 plots IM. Since these variables seem to exhibit non-stationarity, unit-root tests have been conducted, and the resulting Dickey-Fuller (DF) and augmented Dickey-Fuller (ADF) statistics are reported in Table 1. However, because the sample size of 56 is quite small, these unit root tests may not have much power. The uncertainty over the exact degrees of non-stationarity of individual variables makes it difficult to apply Johansen’s system estimation procedure 40QY
360-
320-
280-
240-
16B-
I C..,, ,. *. ,' ,' '__.,I .__ # , ,
12l3-
88-
,. ._,I
. ’ , \,' >'../' ' ,'
;*._. , 4a________*-._--
I
,-_*_-
*-____.----
e-'...".'.'.'..'.."".."".".'."'. 1980 1982
Figure I.
_- --___
1984
1986
1988
Real MO and GDP in 1980 Prices
1990
1992
JOURNAL OF ASIAN ECONOMICS,
(5)2,1994
~...... ___ ,. F-----h .‘-..-.---.-~-‘-‘-.--“-‘ . :--.-.f.-...,,: IR
,.
.
..:
‘.
\
._.__
‘.._
___.’
__I
R
‘\
I
,*
.
;.*
<
-161
.'.
* 1982
'.
.' 1983
'.
.
1.. 1985
Figure 2.
..I
.
1987
.
.
I 1989
.
.
~.
I,
.
1991
R, IR and PR
9
1987
Figure 3.
LnRSL
. 1993
2
1982
.
1992
.
Money Demand in China
259
1982
1987 Figure 4.
1992
IM
(1988), since the procedure is quite sensitive to changes in the numbers of lags, variables and their degrees of non-stationarity. Moreover, the small sample size is likely to give biased estimates of the static regression (long-run) coefficients by the Engle-Granger (1987) two-step estimation method. In light with the situation, the following practical strategy is adopted: First, Hendry’s genera&simple reduction approach (1991) is used (i.e., the modelling starts from a log-linear, autoregressive distributed-lag (ADL) relation with LnGDP,, R,, IM,, LnRSL, in six lags and seasonal dummies, and then gradually reduces and reparameterizes into a parsimonious errorcorrection model. Johansen’s procedure is used afterwards to help check weak exogeneity of the obtained single-equation results, since the above procedure conditions crucially upon weak exogeneity. Table 2 presents the final results of the modelling search by OLS method. Figure 5 show,s the fit.8 Notice that a unit-elasticity restriction on GDP is imposed in the long-run relation imbedded in the error-correction term in accordance primarily with economics. The restriction is not rejected by data during the model reduction. It actually helps to isolate the monetization effect on transaction demands because there is a certain degree of collinearity between IM and GDP. Another point worth noting is that LnRSL, does not enter the EC term and only ALnRSL, is found significant as a short-run factor. To check whether the long-run equilibrium in the EC term is unique in the sense that GDP, R and IM are weakly exogenous with respect to M, we apply Johansen’s
260
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TABLE 2. c
G-I
-.258 (.102) Notes:
-.123 (.048)
OF ASIAN ECONOMICS,
(5)2,1994
OLS Estimates of an EC Model of ALnM, (1978-1991) &LnGDP,
AL~(GDP/IV),~
AblRSL,
wt
Q2
,504 (.028)
,271 (.038)
-.414 (.143)
-.004 (.0018)
,100 (.014)
The figures in brackets are standard errors; A denotes first difference; A, denotes two-lag difference; Q2 is a seasonal dummy of the second quarter; R* = ,925; F(6,44) = 91.05; s = ,029s; DW = 2.36; Normality x’(2) = 0.7M); Residual autocorrelation F(4,40) = 2.475, ARCH x2(4) = 7.875 (lags IA); Heteroscedastic errors x2 (11) = 15.837; RESET F(l,43) = 0.987; Variance instability: 0.098; Joint instability: I.170 EC, = LnM, - LnGDP, + 0.01 R, -
I .4 IM,
(1988) procedure to the relation for both the cases of unrestricted and restricted GDP coefficient. The results are summarized in Tables 3a and 3b. It is seen from the tables that likelihood ratio tests for the both cases suggest the existence of at least one cointegrating vector, and that the coefficient estimates are quite sensitive to both the restriction and the number of lags. A more powerful check for exogeneity is parameter constancy (cf. Ericsson 1991). Figures 6-9 are some of the recursive results based on Table 2. The coefficients appear remarkably constant within their 95 percent confidence intervals through the sample
Figure 5.
Fitted and Actual ALnM, of the Quarterly Model
Money Demand in China
TABLE 3a.
261
COINTEGRATION
RESULTS BY JOHANSEN
PROCEDURE
(LnW, LnGDP,, R,, IM) H,: r = no. of cointegrating vectors (CV): r13 rS2
I LAGS
4LAGS
8LAGS
Critical Values 5%
0.126
1.037
0.421
9.094
6.518
20.168
3.491
rS1
11.49
25.82
r=O
28.67
127.3
EC coefficients
28.24
73.82
-0.162
35.068
107.0
.034
53.347
-.075
Estimated CVs for Ln M: LnM
1.0
Ln GDP
1.0
-2.01
1.0
3.453
R
0.002
IM
0.378
-1.32
0.080
0.032
-10.5
-0.64
period, especially in view of the rapidly changing economy during the reforms, and particularly the shock occurred in the summer of 1989. To investigate further the issue of constancy, the same modelling procedure is repeated on an annual data set for the period 1952-199 1. As GDP was not recorded before the reforms in China, national income (1980 price), NI, is used here instead. IM is set to constant prior to 1978 and to the annual average of the quarterly IM thereafter.’ The results of the annual model are summarized in Tables 4-5 and Figures 10-14. Comparing the annual model with the quarterly model results, we notice two particularly remarkable features: (1) The coefficient estimate of the EC term in the annual model (-0.557) is about four times of that of the quarterly model (-0.123) with respect to their confidence intervals; (2) the two long-run relations imbedded in the
TABLE 3b. -
Cointegration
Results by Johansen Procedure
(LnWLnGDP,,
H,., r = no. of cointegrating vectors (CV): rS2 0.60 rll
4.032
R,, IW I LAGS 1.084 5.861
4LAGS 0.001 8.939
rS0
33.08
31.72
27.37
EC coefficients Estimated CVs for Ln M:
-0.594
-0.05
-0.011
Ln M/GDP
1.0
1.0
R
0.004
0.094
IM
-1.63
-1.12
1.0 -0.21 1.141
8LAGS
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-.88
-.2c__---..________----
,.---__ ,*': : I' I
-.24-
;._.--
,1---.., ,-
1 ' < ' I ' q,_,'
-.32-
-.36
.-___-r
>*, ,' I< _ , \.-I
_c-r-._ ', ..I
-.28-
..
".
""I'. 1984
Figure 6.
1985
'I". 1986
RLS Estimated Coefficient
I'. 1987
""'I 1988
'
”
1989
of EC,_, (with its 95% confidence
199e
"'I'. 1991
.' 1992
interval as the dotted lines)
Figure 7. RLS Estimated Coefficient of ALnRSL, (with its 95% confidence interval as the dotted lines)
Money Demand in China
263
-.BQd-
.. .
-.ee8-
,.--.
: -.a1
-
-.lBL2L.
Figure 8.
_._4----,
I._._-.
‘.
.
’
1984
,
,’
.
,
,
_,----__._-___~_-----
,‘\ ,, <*
‘---.____--a
a. _-._,’
I
‘________-
.I f
. ,,’
.a
’
1985
.
.
.
’
1986
RLS Estimated Coefficient
.
.
‘.
.I
1987
.
1988
.
1
1989
of AR, (with its 95% confidence
.
I
1990
.
I
1991
.
.
,
1992
interval as the dotted lines)
1. Ia l-
.95 l-
____---_________---____~~~__________~__________________~~-~~~~~-----~_~~_~____~_~~___. t 52 critical
.sl-
value
.81
.72
.63 I-
.54 l-
.36
.27
.18
I
1984
.
1985
Figure 9.
I
1986
.
.
I
1987
.
.
I
1988
.
.
I
1989
,
1998
Forecasting Chow Tests (dotted line is 5% critical value)
,
1991
.
,
1992
264
JOURNAL
-.18
OF ASIAN ECONOMICS,
:
;
‘< -.24
(5)2,1994
".
'I 1968
"'
'I 1965
"""".""". 1970
“’
”
1975
1980
I’
".
1985
199e
'
“.
1995
Figure IO. Fitted and Actual ALnM, of the Annual Model
-.4-
-.45-
_--.-__ ,' __-.______,.~~-~..-_.._ I. _,,' _ -._____..~" *.*A...-..,____\__ ,_/_..I~'~~~--..._.__,..~~-................_.~
-.5-
-.55-
-.6\
-.95[
’
1
Figure II.
’
’
1975
RLS Estimated Coefficient
1980
of EC,,
’
’
1985
(with its 95% confidence
.
.
1990
’
.
’
.
’
1995
interval as the dotted lines)
Money Demand in China
Figure 12.
265
RIS Estimated Coefficient of AzLnRSL, (with its 95% confidence interval as the dotted lines)
-.erz-
. ,_.__-_
,_,______..._~.~~....._~_~~........~ -.l315-
__,* ..
. . . . . _.
,’
..
,,*’
..*._..
,I’
*. ,,’ ..\,_.,_
,.*. ,’ -.818_,_,..’ :
,,’
:
.,_-.-.-
-.821-,.-
-.a24
,/’
Il....
I....#....,..,.,*. 1975
Figure 13.
,., 1980
RLS Estimated Coefficient
1985
of AR, (with its 95% confidence
1990
1995
interval as the dotted lines)
266
JOURNAL
OF ASIAN ECONOMICS,
(5)2,1994
1.04r
1.... . . .___. ...... ..... .........._.... ........_._.._.._
... .. ..___________._.......................
‘-“t””
.96 c
5X critical
level
.ea .8 .72
t
.64
-56t .4a
.4
.32i
:::?._,..-----._-.eal..
.
.
1.
.
1975
LtlM
DF ADF(1)
-1.710 (.341) Nok-
. 1985
1
.
1 1990
J 1995
Unit Root Tests (critical value: -2.9) LnNI
R
IM
LnRSL -3.685
1.142
0.372
-2.926
4.941
1.038 ALnM
0.064 ALnNI
-2.997
1.75
-2.527
AR
AIM
A2LnRSL
-5.283 -3.622
4.25 4.866
-6.154 -5.654
-2.93 -1.655
4.092 11.698
TABLE 5. C
.
Forecasting Chow Tests (dotted line is 5% critical value)
TABLE 4.
DF
1.. 1980
Figure 14.
ADF(1)
.
OLS Estimates of an EC Model of ALnM, (19521991)
ECt-I
-.557 (.076)
MLnNI,
AQf,
A&nRSL,
A&
,216 (.065)
2.250 (.324)
-.675 (.260)
-.0076 (.0029)
M denotes taking tirst difference twice; RZ = ,817; s = ,072; DW = 2. II; Nomality x2(2) = 0.372; Residual autocorrelation F(2, 29) = 1.089, ARCH x’(l) = 0.414 (lag I); Heteroscedastic errors x*(10) = 4.284; RESET F(1.30) = 0.537; Variance mstability: 0.165; Joint instability: 0.987 EC, = LnM,-LnY,
+ O.OlR, -1.65 IM,
Money Demand in China
TABLE 6.
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Post-Model Prediction of 1992 by the Quarterly Model
Actual MO (billion) 311.7 315.6 355.9 433.6
Predicted MO
Prediction Errors %
316.7 310.6
1.6
347.9
1.6 2.2
413.0
4.7
EC terms are virtually identical.” Moreover, the annual model appears similarly constant (see Figures 11-14). However, it is difficult to compare the short-run dynamic effects of the two models, because the units of the lag differences are incompatible. Finally, post-model prediction is made for the year 1992 using the quarterly model, and prediction errors are found below five percent (see Table 6), in spite of the fact that the annual growth rate of MO shot up from around 20 percent in 1991 to above 36 percent in 1992. The consistency between the quarterly and annual models and the relatively satisfactory econometric results lay the ground for the following tentative arguments.
4.
ECONOMIC
IMPLICATIONS
OF MODEL RESULTS
The first argument is that China’s economic reforms do not seem to have induced changes in the standard propensities of the transaction and portfolio money demands represented by GDP, or NI, and real interest rate. The fact that simple, standard money demand models have suffered from ‘structural breaks’ since the reforms does not necessarily indicate changes in these propensities. More often than not, it indicates insufficient model formulation in the sense that certain important variables representing the specific features of the economy or the regime are missing. The present modelling search has identified two of such variables, IM and RSL, and established a relatively constant model. These two variables deserve further explanation since they are non-standard money demand variables and they play a crucial role in the present model. First, let us look at IM. Although a proxy, the variable has segregated the trend of marketization from the normal transaction demands represented by GDP or NI. The segregation helps to clarify the confusion that some of Chinese economists have between excess money supply and faster MO growth than GDP growth due to the marketization effect on money demand. It also helps to maintain the economic sense of the long-run relation imbedded in the ECM. Take for example the long-run relation under equilibrium from the quarterly model: LnM = Lnc + 1.65ZM + LnGDP - O.OlR or M = cZM’.~~~.“GDP
(1)
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where c is a constant. Since the values of IM are smaller than one, its exponential by 1.65 will not cause the growth of MO accelerate faster than the growth of GDP, unlike those models which have produced estimates of LnGDP coefficient greater than one. Furthermore, IM will converge to a constant once the monetization process ends, reducing the long-run relation under equilibrium to a simple and standard one involving only GDP and R: M = kR4.”
GDP
(2)
Next, let us look at RSL. As seen from Tables 2 and 5, only ALnRSL is found significant and its negative coefficient estimates comply with the prior expectation. which represents the acceleration/deceleration rates of the proportion/disproportion of the growth rates between savings and loans. Since ‘investment hunger’ almost always exists and the Chinese government has maintained low and stable income policies and fixed interest rate policies, significant fluctuations in ALnRSL have been induced much more often by drastic credit plans than by saving fluctuations. However, expansionary investment decisions alone is insufficient to stimulate money demand unless the expansion exceeds significantly the economy’s saving potential. Now the question arises that this variable seems to be a supply-side factor from the viewpoint of a standard market economy. But as mentioned earlier, an important feature of a CPE is that its government plays not only the role of the economic regulator, but also the role of the chief representative of the state-owned sector, which forms the major part of the economy. If we analyze such an economy in the form of a real section versus a financial section, government credit decisions then reflect the investment behavior of the state-owned sector, and therefore should be considered as affecting transaction demands for money through fluctuations in the sales of intermediate goods. From this perspective, the significance of ALnRSL appears to correspond to Komai’s conjecture that ‘soft budget constraint syndrome may show up in the formation of excess demand’ (1986), since excessive investment expansions exhibit the syndrome of ‘soft budget constraint’ at the macro level. Notice that LnRSL does not enter the long-run relation, suggesting that ‘soft budget’ in the form of excessive investments at the macro level is transient and bound by the long-run market equilibrium condition. The model result with RSL also suggests that modelling government investment behavior is more fundamental than modelling money demand, in the sense that money is ‘passive’ not only with respect to the usual demand factors of a market economy, but also with respect to special factors of a CPE. Finally, let us look at R briefly. The interest rate variable is absent in most previous econometric studies of China’s money demand, except perhaps only (Portes and Santorum, 1987). Here, the significance of both the long-run and short-run coefficients of R indicates that the cost of holding money is not a negligible factor in studying China’s aggregate money demand behavior. Since the government has implemented fixed interest rates policies, much of the dynamic effect of AZ? is due to inflation fluctuations, as shown in Figure 2. ‘I The relatively constant coefficient estimates found for this variable as well as the long-run cointegration tests suggest that inflation is
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269
weakly exogenous for money, since a regression with constant coefficients and non-constant marginal distribution of exogenous variables cannot produce a constantcoefficient inverse regression (cf. Ericsson, 1991). Although weak exogeneity does not carry the full capacity of determining ‘causality’ in the common sense, the present finding with R is supportive to those Chinese economists who argued that inflation was not fundamentally caused by active money expansion during the reforms (e.g., Dai, 1992). More conclusive arguments entail further econometric studies of the price determination mechanism in China.
APPENDIX (A) Quarterly data 1978(l)-1992(4) MO GDP R
IM
RSL Source:
money in circulation deflated by GDP deflator (billion yuan) gross domestic products in 1980 prices (billion yuan) IR - PR (IR: interest rate of one-year deposit savings; PR: inflation rate = annual rate of overall retail price indices) 0.5 RP + 0.5 RO (RP: ratio of agricultural output deflator to industrial output deflator; RO: ratio of non-state-owned industrial output to total industrial output) annual rates of the ratio of total bank savings to total loans = &Ln(SIL)
the State Statistical Bureau of China
(B) Annual data (19521991) MO Y R
money in circulation deflated by national income deflator (billion yuan) national income in 1980 prices (billion yuan) interest rate of one-year deposit savings - inflation rate = 0.5 for 1952-77, the annual average of the above IM for 1978-91 annual rates of the ratio of total savings to total loans = ALn(SIL)
IM RSL Sources:Statistical
Yearbook of China (various years); interest rates for 1952-77 from Byrd (1983)
Acknowledgment: An earlier version of this paper has come out in the Discussion Paper series of the Department of Economics, Queen Mary and Westfield College (No. 281) under the joint authorship with Tongsan WANG, Mingwu ZHOU and Guoqiang LIU. I sincerely acknowledge the Ford Foundation research grant 910-1107, which enabled me to work on modelling the Chinese economy with the help of WANG, ZHOU and LIIJ. The present study in money demand is an extension of the modelling exercise and has been carried out solely by the current author. Nevertheless, I am very grateful to them three. Thanks should also go to Lizuo JIN, Sheng HUA, Weiying ZHANG, Guy LIU, as well as many other Chinese economists, for their insightful views on China’s economy, and to Chenggang XU, Christopher Gilbert, members of the Economics Department of QWM and three anonymous referees for their helpful comments on the earlier versions of this paper.
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NOTES 1. Burton and Ha (1990) estimated various money demand, such as household demand, enterprise demand, savings deposits, as well as MO, Ml, and M2, with quarterly data for the period 1983-1988. 2. For more detailed description, see Portes and Santorum, 1987, Yi, 1991; 1992 and Ma, 1990. 3. Modelling with either ratio as the index has been experimented. The results are slightly worse than the combined index. Equal weights for the two ratios in the combined index are imposed solely for simplicity. 4. In fact, the aggregate level of credit funds has been used by the central bank in China as one of its main intermediary targets of the monetary policy, cf. Ma (1990). 5. The ratio is also an important index that the Chinese government monitors cf. Ma. 6. Real money is calculated from nominal money using the same deflator of real GDP to keep the measurement of these two flow variables as consistent as possible. 7. The issue of data accuracy is ignored here on the assumption that there are no significant changes in the data collection process as far as accuracy is concerned, since what all applied economic studies do is to explain and predict given data. 8. 2SLS and IV methods for the variable RSL, have been experimented. The results do not differ significantly from Table 2, and are not reported here. 9. The constant is set at 0.5 on the simple assumptions that the price ratio was one and that the industrial output ratio was zero. 10. The only difference lies with the coefficient estimates of IM,, which trends with GDP or NI since the reform. The following simple regression for the period 1978Ql-199144 shows: GbP=
1.24NI=,E>&
Hence the annual estimate of the coefficient of IM, ought to be slightly larger than the quarterly estimate. 11. The alternative of estimating separately the inflation effect and nominal interest rate effect has also been experimented during the modelling. The resulting estimates do not contradict the composition of a single variable of real interest rate.
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Received July 1993; Revised March 1994