Economic policies and demand for money: evidence from Canada

Abstract

This study identifies Canadian fiscal and monetary policy regime changes that could influence the services of money. It is argued that if these policy regime changes were not incorporated in the estimation of demand for real balances, the resulting estimate would be biased and unstable. Using Canadian monthly data for the January 1975 to June 2001 period, the paper estimates a standard demand-for-money (M1) function with and without these policy regime changes. It was found the demand for money in Canada is stable over the short- and long-run periods when these policy regime changes are incorporated and the estimated coefficients have correct signs.

Notes

¹ The assumption that the variable ownr is constant or zero is not only based on what has been practised in this literature, but is also based on the following evidence: (i) In our sample period, on average 42.44% of M1 is notes and coins in circulation. The remaining 57.56% of M1 is demand deposits. Only a small portion (no precise data available) of such demand deposits includes interest-bearing checking accounts. The interest rate on this small portion of M1 was constant at 3% from 1975 to the end of 1991 (the first 16 years of the sample period). From January 1992 until December 1998, the rate of interest on this small portion of M1 was constant for some months and then fell (as interest rates in general fell) and remained constant, and so on, with a value of 2.54% in January 1992 and 0.05% in December 1998 with an average rate of 0.47%. This rate remained constant at 0.03% from January 1999 to February 2001 and then fell (as interest rates, in general, fell) to 0.02% up to the end of the sample period. All of these rates are daily interest rates at the annual rate and the source of the data is CANSIM # B14035. As one can see, the interest rate on the small portion of M1 has been either constant or close to zero in the sample period. This evidence supports the assumption in the literature, including in this paper, that the own rate of M1 is either constant or zero.

² According to Friedman (1988), stock prices can influence the quantity of money demanded through four effects: wealth, risk-spreading, transaction and substitution: (i) A rise in stock prices results in a higher wealth which can be expected to increase demand for money. (ii) For a given risk aversion/preference, a rise in stock prices reflects an increase in the expected return from risky assets relative to safe assets, implying a higher relative risk. The higher risk can be offset by lowering the weight of long-term bonds in the portfolio and/or by increasing the weight of highly liquid fixed-income assets as well as money in the portfolio. (iii) An increase in stock prices may be taken to imply a rise in the dollar volume of financial transactions, resulting in an increase in the demand for money to facilitate transactions. (iv) An offsetting impact of these factors is a substitution effect of a change in stock prices. The higher the real value of stocks is, the more attractive stocks are as a component of the portfolio. Consequently, the sign of F_rtse is an empirical issue.

³ These dummy variables have a value of one for the appropriate month(s) in the period, a value of negative one in the immediate month after the period, and zero otherwise. This is because the effect of these postal strikes is believed to be an accumulation of money during the strike period which fell to the normal level after the period. This would mean a positive spike in the money growth in the first month of the period followed by a negative offsetting spike in the money growth in the immediate month after the strike. To clarify further assume the money supply before the strike is growing at 2% and as a result of the strike there is a growth of 10% in the supply of money, i.e. a positive spike of 8% in the growth rate in the first month of the strike. When the strike is over the growth of the money supply will fall by 8% for the immediate month after the strike (a negative spike of 8%) and after that it will be back to its normal level of 2% growth. So in a growth rate model, it makes more sense to have a dummy [… , 0, 1, … , 1, −1, 0, …]. Note that one effect of a postal strike is an increase in the money supply. Customers’ payments were delayed while firms’ obligations such as payrolls could not be postponed. Consequently, firms borrowed from the banking system. The Bank of Canada's policy has been to accommodate this additional and temporary demand for cash balances. At the same time, agents who had bills to pay had higher cash balances than desired. Since they mentally debited their accounts they thought they were holding their desired levels. Consequently, this phenomenon causes the standard demand for money to underestimate the demand for real cash balances, Gregory and MacKinnon (1980).

⁴ In January 1976 banks first began offering cash management services to large firms. Furthermore, in the early 1980s when interest rates were high, money demand shifted due to both the spread of corporate cash management services to smaller firms and the introduction of daily interest checkable savings accounts. Since this shift was not abrupt but rather continued into at least 1983 the dummy variable INOV80, which has a value of zero until January 1980 and then commences linearly upwards to a value of one for December 1982 and remains one after, is included.

⁵ Note that there is no reason to believe that any of the dummy variables in the set DUM affects the long-run parameters. Examples of structural breaks, which affect the long-run (cointegrating) parameters, include the unification of Germany, the creation of the European Monetary System (EMS) and the collapse of the Soviet Union. For instance, in their estimation of demand for M3 for Germany for the sample period 1975 to 1996, Saikkonen and Lütkepohl (2000) allow the level shift in long-run parameters due to the unification of Germany. Johansen et al . (2000) analyse the Uncovered Interest Parity hypothesis between Germany and Italy for the period 1973 to 1995. They introduce two structural breaks in the study. The first break coincides with the creation of the EMS and the second break corresponds to the exit of Italy from the EMS as well as the unification of Germany.

⁶ M1 (CANSIM # B2033) is in millions of dollars and consists of currency outside banks and chartered bank deposits, the real total industrial production is in millions of dollars (CANSIM # I57001) and TSE 300 Composite Index is at closing quotations at month-end (CANSIM # B4237). All of these variables deflated by the Consumer Price Index (CANSIM # P100000). The 30-day corporate paper rate is at the annual percentage rate (CANSIM # B14039).

⁷ Choudhry (1996), footnote 4, provides a good explanation, with the relevant literature, as to why the use of seasonally unadjusted data is preferable to the seasonally adjusted data.

⁸ The Augmented Dickey–Fuller and non-parametric Phillips–Perron tests were used to investigate the stationarity property of the variables. According to the test results, all variables are integrated of degree one (non-stationary). They are, however, first-difference stationary. For the sake of brevity, these results are not reported, but are available upon request.

⁹ It should be noted that in ECM we allow agents to be backward looking (reacting to previous deviations from equilibrium) while they may also be forward looking if at least one of the variables in the system has a statistically significant and instantaneous impact on the demand for real balances and fails to be superexogenous.

¹⁰ Note that the complete elimination of reserve requirements and the introduction of NAFTA took place within six months apart in 1994.

¹¹ Note that an ECM of the extended version of the model, i.e. including also the real exchange rate and one-month US covered corporate paper rate, was also estimated. Similar to the above result the estimated coefficients of the extended model were found to be jointly unstable when the impact of policy dummy variables was ignored and stable, otherwise. This result is not reported but is available upon request. Other estimated coefficients were not materially different than what was reported for our standard model. Furthermore, the estimated coefficient of the US covered interest rate was not statistically significant and was dropped. This result can be due to a high colinearity between the Canadian rate and the US covered rate, as it was evidenced by Kia (1996). In fact, when the model was re-estimated without the domestic rate, but with the US covered rate, the estimated coefficient of the US covered rate had a correct sign and was statistically significant.The estimated coefficient of the growth of the real exchange rate was negative implying the existence of dollarization over the short term. However, this result should be interpreted cautiously. We can write:

where ex is the nominal exchange rate, pf and p are US and Canada price levels, respectively. Consequently, instead of the change of the log of real exchange rate, the change of the log of nominal exchange rate and the US–Canada inflation rates differential were included and the extended model was re-estimated. The estimated coefficient of both the change of the log of exchange rate and the US–Canada inflation rates differential was positive and statistically significant. All other coefficients remained the same or were not materially different than before.The estimated positive coefficient of the growth of the exchange rate, for a given US–Canada inflation rates differential, indicated that a depreciation of the Canadian dollar increases the consumption of domestically produced goods and, therefore, the demand for real cash balances in Canada would go up. Namely, there was no evidence of dollarization in Canada during the sample period. Similar to our standard model, the error term had a linear effect on the demand for real money when the impact of economic policy changes was not incorporated. However, with the correct specification, the impact of the error correction term was nonlinear. In sum, we conclude again that when the impact of fiscal and monetary policy regime changes which enhance/weaken the services of money is ignored the estimated demand for real balances will be biased and unstable.

¹² Using a base of a three-month average of the seasonally adjusted level of M1, beginning April 1975, the Bank specified a target range of 10–15%. The following changes were announced: 8–12% in September 1976, 7–11% in October 1977, 6–10% in October 1978, 5–9% in January 1980, and 4–8% in March 1981 (Deaves, 1991).