V Estimating Money Demand in Lebanon
Author:
Mr. Jose Martelino
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Mr. S. Nuri Erbas
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Mr. Adnan Mazarei
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Ms. Sena Eken
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Mr. Paul Cashin
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Language:
English
Keywords:
OP; exchange rate; Lebanese pound; economy; exchange rate movement; currency substitution; net foreign currency position; US$ exchange rate series; Lebanese economy; Exchange rates; Monetary aggregates; Dollarization; Personal income; Middle East; South America; Africa; Europe; currency holding; dollar-denominated deposit; nominal income; unit elasticity; nominal monetary aggregate; Income; Monetary base; Bank deposits; Demand for money
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Abstract

Estimating money demand for Lebanon has been made more difficult by the war-induced increase in currency substitution that occurred during the sample period. Empirical tests suggest several promising monetary aggregates that result in a stable money demand function, giving the Lebanese monetary authorities scope to influence future developments in both the price level and income. These results provide econometric evidence that, despite the economic turmoil of the civil war, stable equilibrium relationships do exist in the long-run series for money, prices, and output in Lebanon.36 Specifically, in the context of monetary programming, the results also suggest that the domestic component of total liquidity is preferred as the monetary target for money denominated in Lebanese pounds, and the foreign component of total liquidity is preferred as the monetary target for dollar-denominated money.

Estimating money demand for Lebanon has been made more difficult by the war-induced increase in currency substitution that occurred during the sample period. Empirical tests suggest several promising monetary aggregates that result in a stable money demand function, giving the Lebanese monetary authorities scope to influence future developments in both the price level and income. These results provide econometric evidence that, despite the economic turmoil of the civil war, stable equilibrium relationships do exist in the long-run series for money, prices, and output in Lebanon.36 Specifically, in the context of monetary programming, the results also suggest that the domestic component of total liquidity is preferred as the monetary target for money denominated in Lebanese pounds, and the foreign component of total liquidity is preferred as the monetary target for dollar-denominated money.

Dollarization and the Definition of Money

U.S. dollars in circulation in Lebanon serve as both a store of value and a medium of exchange. However, the substitutability between Lebanese pounds and U.S. dollars can be gauged only for dollar-denominated bank deposits and pound-denominated money (currency plus deposits). As previously mentioned, dollar currency in Lebanon is unobservable, and no official data exist.

Notwithstanding these difficulties, for purposes of monetary targeting by the Lebanese authorities to achieve desired outcomes for key economic variables, a necessary requirement is that a stable demand relationship exist between the monetary aggregate of choice and these variables. In this sense, a suitable monetary aggregate is defined as one with a stable money demand function, and several possible candidates are examined here: base money (BM), narrow money (M1), foreign currency deposits (FCD), the Lebanese pound component of M2 (M2LL), and total liquidity, including FCD (M2).

Behavior of Velocity

The wide variation in the velocity of the five monetary aggregates over the sample period is depicted in Chart 9. Velocity for all five aggregates declined during 1964–82, thereafter rising. The fall in the velocity of BM and M1 after 1974 reflects the increase in Lebanese pound currency holdings in the early years of the civil war. The sharp spike in the velocity of BM, M1, and M2 LL around 1986 reflects the abandonment by Lebanese residents of pound-denominated money at this time, and is matched by an increase in dollar-denominated deposits both inside and outside Lebanon. For most of the sample period, foreign currency deposits were increasingly used to avoid the opportunity cost of holding pound-denominated assets in a high-inflation environment.37

The time path of the magnitudes of the various monetary aggregates is shown in Chart 10 and mirrors the results of Chart 9. Of greatest interest is the rapid relative growth of foreign currency deposits, particularly during 1977–80 and 1982–86. A clear picture of the composition of total liquidity (money plus quasi money) is given in Chart 11, which demonstrates the sharp collapse (rapid rise) in the relative weight of pound-denominated (dollar-denominated) deposits in total deposits after 1985 and a similar fall in the share of Lebanese pound currency holdings at that time. After 1987, the share of pound-denominated assets in total liquidity began to rise, but there is evidence of hysteresis in this process; between 1991 and 1993, the share of such assets in total liquidity remained constant at about 40 percent, far below the 1985 level of about 70 percent.

Chart 9
Chart 9

Velocity of Monetary Aggregates

Sources: Bank of Lebanon: International Monetary Fund. international Financial Statistics, and IMF staff estimates.

Specifications of Money Demand

Two simple models of money demand are posited. The first model follows a standard specification;

Mt=β0+β1Pt+β2yt+β3πt+β4D+ϵt,(1)

where Mt is the log of the nominal monetary aggregate (alternatively defined as above); Pt is the log of the price level; yt is the log of real income; πt is expected inflation (proxied by the difference of the log of the actual price level); D is a deterministic dummy variable to account for the civil war years of 1975–90 (taking a value of unity in those years and zero elsewhere); and εt is the error term. In this specification, money demand is expected to increase with rises in the general level of prices and real incomes and to fall with increases in the rate of change of the price level (which raises the opportunity cost of holding money, particularly currency holdings, which have a zero nominal rate of return).38 The deterministic shift parameter is included to account for the shock to money demand caused by the civil war. (See Giovannini and Turtel-boom (1992) and Orden and Fisher (1990) for uses of this approach.)39 The second model is the constant-velocity model of money demand, where the nominal monetary aggregate is deemed to have a stable relationship with nominal GDP. It is important to note that any such stability in the income velocity of money requires that the data not reject the restriction of a unit elasticity of money demand with respect to output and prices (see Orden and Fisher (1990)).

Chart 10
Chart 10

Monetary Aggregates

(In millions of Lebanese pounds)

Sources: Bank of Lebanon: and International Monetary Fund, International financial Statistics.
Chart 11
Chart 11

Composition of Total Liquidity

(In percent)

Sources: Bank of Lebanon; International Monetary Fund, International Financial Statistics; and IMF staff estimates.

Unit Root Tests for Stationarity40

Recent developments in time-series econometrics have made clear that models that ignore the nonstationarity of individual series will generate spurious regression results and that if the (similarly integrated)41 series in any given model are cointegrated,42 then linear combinations of these variables will converge to stationary long-run equilibrium relationships. Stationarity can be achieved through (1) differencing the data; or (2) formulating a regression model with nonstationary variables, a linear combination of which produces stationary error terms, using the concept of cointegration.43

To ensure that all variables in an equation such as (1) form a cointegrating relationship with a stationary error term, it is necessary to establish that all individual time series are of the same data-generating process (or order of integration) I(d), d >0, and that the residual term in (1) is stationary, thus minimizing the probability of obtaining spurious results (Granger (1986)).44 In testing whether or not there is any long-run relationship between the variables of interest, the cointegration technique makes use of the fact that nonstationary series generally evolve over time. A stationary error term means that the variables have tended to move together over the long run and that if an exogenous shock drives the variables out of equilibrium, then there is a tendency for them to move together again. In this sense, cointegration of money, income, and prices implies that these variables move together over time and revert to an equilibrium relationship in response to shocks.

Phillips-Perron (1988) integration tests were applied to the following series, which were also subsequently used in the cointegration analysis: real pound-denominated GDP (yLL); real dollar-denominated GDP (y$); the Lebanese consumer price index (P*), the U.S. consumer price index (P*); the Lebanese inflation rate (π); the U.S. inflation rate (π*); the rate of interest on dollar deposits, proxied by the six-month U.S. dollar deposit LIBOR rate (LIBOR6A); the Lebanese discount rate (DISRLBN); the various Lebanese nominal monetary aggregates: base money (BM), narrow money (M1), dollar-denominated deposits (FCD$), the Lebanese pound component of broad money (M2LL), broad money including FCD, valued in Lebanese pounds (M2); and their real counterparts: BM-P, M1-P, FCD$-P*, M2LL-P, and M2-P. All monetary data were taken either from the International Financial Statistics data base or from the BDL, and the real sector data were supplied by IMF staff and BDL estimates, with interpolations carried out where necessary (see Section II for details).

The results of the stationarity tests for 1964–93, presented in Table 10, clearly indicate that all of the variables tested are not stationary. Further, most of them do not contain a deterministic trend component, with the possible exception of M2, which could well be due to measurement error arising from its exclusion of U.S. dollar currency in circulation. All the statistics testing for one unit root versus no unit roots are insignificant, indicating that the null hypothesis of a unit root cannot be rejected. Alternatively, all the statistics testing for two unit roots versus at most one unit root are significant (not reported here). Hence, all series are integrated of order one (I(1)), and the logs of the levels of the individual time series are nonstationary. This finding has two important implications. First, shocks to the economic system (such as the civil war) will move the above variables permanently above or below any previous trend in their evolution. Second, the level form of these variables can be modeled in a cointegrating equation and tested for the existence of a long-run relationship.

Table 10

Phillips-Perron Tests for Stationarity, 1964–931

article image
Source: IMF staff calculations.

The Phillips-Perron (1988) tests are based on the following model for any variable x: xt = βt + αxt - 1 + εt. The test relies on rejecting a unit root (α = 1) in favor of stationarity. The H0: α = 1 is tested by Z(α), Z(tα): and α = 1, β = 0 is tested by Z(Ф3). The statistics in columns 3–5 are computed by restricting the coefficient β to be zero, and if the null hypothesis cannot be rejected (is not significant), then the series has a unit root. The (*) indicates significance at the 5 percent level, A finding of statistical significance means that the relevant null hypothesis can be rejected. All test statistics are calculated at a lag truncation order of three for the residual autocorrelations used in calculating the estimated residual variance. The statistics were similar for other lag lengths.

All series (except for the interest rates) are in logs.

In later sections, the Engle-Granger (1987) procedure is applied to the cointegrating vector that contains the real monetary aggregates (thus imposing homogeneity, a unit price elasticity). The cointegrating vector is then further restricted by imposing a value of unity on the coefficient on real income, thus regressing the nominal monetary aggregates on nominal income. The validity of these restrictions is then tested using the values of the Phillips-Ouliaris cointegration tests. Note that in selecting the appropriate variables to enter the vector of cointegrating variables, t-statistics from ordinary least-squares (OLS) cointegrating regressions should not be relied upon, as the estimated standard errors are biased and inconsistent (see Engle and Granger (1987)). Instead, it is better to use the cointegration test results to guide both the selection of explanatory variables and an assessment of the validity of restrictions. (See also Muscatelli and Papi (1990) on this point.)

Money Demand: Cointegration Results and Policy Implications

For each of the five monetary aggregates, the preferred OLS results from the cointegrating regressions are listed below for 1964–93.45 All analyses were carried out with data in logarithms and with a linear functional form assumed. A deterministic dummy variable is used to control for the effects of the civil war years of 1975–90. The best results for each monetary aggregate are given below, with (*) denoting that the Z(α) test statistic of Phillips and Ouliaris (1990) is significant:46

BM-P=1.67+0.79y-1.20π+0.77D(2)R2=0.782,Z(α)=-25.82*M1⁢-P=-2.77+1.12y-1.47π+0.77D(3)R2=0.775,Z(α)=-29.69*FCD$=-7.78+2.22P*+0.64y$+0.17π+0.32D(4)R2=0.978,Z(α)=-18.99M2LL-P=0.38+0.96y-1.31π+0.89D(5)R2=0.801,Z(α)=-23.73*
M2=-1.74+1.01P+0.80y-0.06π+0.77D(6)R2=0.993,Z(α)=-15.84

The critical values for the Z(α) tests are given in Phillips and Ouliaris (1990), where the null hypothesis of no cointegration is rejected if the error term from the cointegrating regression is stationary.47 The results for each of the monetary aggregates are examined below.

The results for the regressions on real BM, real M1, and real M2LL of equations (2), (3), and (5) indicate that a cointegrating (long-run) relationship among the money, output, price, and inflation series exists, provided that a single deterministic shift variable is included in the model to account for the effects of the civil war on money demand. In equations (2), (3), and (5), the Z(α) statistic is significant at the 0.075, 0.05, and 0.010 percent levels, respectively. As expected, the coefficient on real output in each of the three equations is positive, and that on inflation is negative, indicating that holding fiat money and bank deposits has high opportunity costs in times of inflation. For example, long-run M2LL demand has a coefficient on real income of 0.96, indicating that a 1 percent increase in real income raises real M2LL by 0.96 percent.

As noted above, all regressions were run after imposing homogeneity. This imposes a unit elasticity on the responsiveness of money demand to changes in the price level and was not rejected by the data in equations (2), (3), and (5). That is, there is a proportional relationship between increases in money stocks and changes in the price level for these monetary aggregates. No cointegrating relationship was found in equations (4) and (6). For equations (4) (and (6)), the coefficient on π was positive (and barely negative), as expected, reflecting the advantages of holding dollar-denominated money in the presence of pound-based inflation.48

While the coefficients on the variables in the cointegrating regression have the appropriate signs, the results for M2 and FCD$ indicate that the relevant monetary aggregate and y (or y$), P (or P*), π, and D do not form a cointegrating vector—there is no stable long-run relationship among these variables. In particular, the result for M2 can be explained by the wide variance in the holdings of dollar-denominated deposits and dollar currency by the Lebanese during 1964–93, and by the fact that data on dollar currency in circulation are unavailable, which increases the probability of measurement errors.

Nominal interest rates were found not to be important in any of the five cointegrating vectors and were consequently excluded. This result is not unexpected, given that government and financial market regulation constrained interest rates below the market rate for much of the sample period. Similarly, foreign interest rates (adjusted for expected depreciation of the exchange rate) also proved to be unsatisfactory. The fact that the income velocity of money is relatively unaffected by interest rate movements should also be expected given that deposits in broad money aggregates earn interest, which will presumably adjust in line with returns on nonbroad money assets.49

As to the question of which real monetary aggregate exhibits the more robust long-run relationship with the explanatory variables, Banerjee and Dolado (1986) show that although the estimates of the cointegrating regression are consistent, they are subject to a finite sample bias, and a general guide is that the bias is inversely related to the R2 statistic of any given equation. Accordingly, under this rule, M2LL would be ranked ahead of BM and M1 in terms of a stable long-run money demand function.

The results for those monetary aggregates where the assumption of a unit elasticity on price and income was accepted (BM, M2LL, and FCD$) are shown below, with (*) again denoting that the Z(α) test statistic of Phillips and Ouliaris (1990) is significant:

BM-P-y=-1.30-1.25π+0.80D+0.003T(7)R2=0.780,Z(α)=-28.73*M2LL-P-y=-0.20-1.34π+0.89D+0.002T(8)R2=0.795,Z(α)=-24.37*FCD$-P*-y$=-2.72+0.41π+0.53D+0.068T(9)R2=0.928,Z(α)=-22.15*

For BM and M2LL, there is evidence of a long-run relationship between (M - P - y), π, D, and T as the above joint restriction on the cointegrating vector was not rejected by the Phillips-Ouliaris test statistic (the Z(α) statistic is significant at the 0.05 and 0.10 percent level, respectively, in equations (7) and (8)). In equation (9), FCD$ also evidences a long-run relationship with the variables in its cointegrating vector, as again the joint restriction of a unit elasticity on P* and y$ was not rejected by the Z(α) test statistic, which was significant at the 0.125 percent level.

There is evidence of a long-run proportional relationship between these nominal monetary aggregates and nominal income, with the R2 statistic again indicating that the strength of this relationship (among pound-denominated monies) is relatively greater for M2LL than for BM. It should be noted that while the trend rate of growth of demand for BM and M2LL is quite low at 0.3 percent and 0.2 percent a year, respectively, for FCD$ there is a trend growth in demand of a sizable 6.8 percent a year. These reflect the replacement of the Lebanese pound by the U.S. dollar as the preferred form of money in Lebanon over the sample period. Accordingly, for BM, M2LL, and FCD$, the adjusted income velocity of money is stationary over the 1964–93 period.

The use of a deterministic trend term in equations (7) to (9) controls for the effects of financial innovation on money demand in Lebanon.50 In this sense, the “adjusted” income velocity of, for example, FCD$ is stationary around a trend. It is “adjusted” in that, after controlling for the effects of the war, expected inflation, and financial innovation on money demand, FCD$ velocity is stationary. In other words, the income velocity of money is not constant but is stable around a trend and will revert to that trend following a shock to money demand. The rationale for the inclusion of a trend term in explaining the demand for FCD$ and the other money aggregates is quite strong, given the wholesale changes in the process of Lebanese financial intermediation that accompanied the country's abandonment of the Lebanese pound and the embracing of the U.S. dollar as the dominant medium of exchange, store of value, and even unit of account.51

The above findings for M2LL and FCD$ provide the basis for the use of these monetary aggregates as nominal anchors in support of the achievement of price stability and output growth targets in Lebanon. In any operational use of such aggregates, the monetary authorities could use M2LL to target growth in pound-denominated money and FCD$ to target growth in dollar-denominated money. Accordingly, the former monetary aggregate would be increased in line with the growth of Lebanese GDP valued in Lebanese pounds, the latter with growth in Lebanese GDP valued in dollars. For a constant LL/US$ exchange rate, growth in the two monetary aggregates would be the same. However, if the Lebanese pound appreciates (depreciates) against the U.S. dollar, then that implies a relatively slower (faster) rate of growth of the dollar component of Lebanese broad money when compared with the constant exchange rate case.52 An implication of this result for any Lebanese monetary program is that when the Lebanese pound appreciates against the U.S. dollar, growth of broad money, including foreign currency deposits in Lebanese pounds (M2), will be less than the growth of GDP in Lebanese pounds. When the Lebanese pound depreciates against the U.S. dollar, growth of M2 (in Lebanese pounds) will exceed the growth of GDP in Lebanese pounds.

Conclusions

Stationarity and cointegration analyses of Lebanese monetary aggregates suggest that, after controlling for the effects of the 1975–90 civil war, the long-run demand for several definitions of money is determined by real income, prices, and domestic inflation. This finding of a stationary long-run relationship between money, prices, and output lends support to the possibility of effective monetary targeting in Lebanon. The data also reveal that discretion in the rate of growth of monetary aggregates can play an important role in controlling movements in the Lebanese price level, as the hypothesis of proportionality (homogeneity restriction) was not rejected for three (BM, M1, and M2LL) of the five monetary aggregates analyzed.

Traditionally, monetary authorities wish to target a monetary aggregate that has a predictable relationship with a nominal macroeconomic variable (such as the price level or nominal output) and that shows little responsiveness to changes in interest rates or movements in other economic variables. While three (BM, M1, and M2LL) of the five monetary aggregates failed to reject the restriction of a unit elasticity on the price level, BM, FCD$, and M2LL failed to reject the restriction of a unit elasticity on prices and income. It was found that demeaned and detrended M2LL, BM, and FCD$ had stationary income velocities of money; following any given shock to money demand, the velocities revert to a constant mean and standard deviation around a trend component that accounts for financial innovations affecting money demand in Lebanon. Hence, the hypothesis of an absence of any long-run relationship between nominal money and nominal Lebanese output is rejected for M2LL, FCD$, and BM. This outcome provides scope for the Lebanese authorities to influence the path of the price level and output by establishing appropriate targets for each of the Lebanese pound and U.S. dollar components of broad money. In the context of monetary programming, the results also suggest that M2LL is preferred to BM as the monetary target for pound-denominated money, and FCD$ is preferred as the monetary target for dollar-denominated money.

36

Earlier estimates of Lebanese money demand were carried out by Short (1981) and Saïdi (1984b).

37

The movements in velocities depicted in Chart 9 follow the typical U-shaped pattern described by Bordo and Jonung (1987). In Lebanon, the pre-1982 downward path of most income velocities reflects both the introduction of modem banking and payments systems and the desire of Lebanese residents to hold their wealth in liquid forms in the early years of the civil war. After 1982, they began to economize on pound-denominated money balances (and to hold their wealth in dollar-denominated and nonmoney forms) in reaction to the likelihood of a protracted conflict and high domestic inflation.

38

The demand for FCD and M2 is expected to be positively and negatively related, respectively, to the Lebanese rate of inflation, as then individuals would switch their demand for money from pound-denominated currency and deposits, which bear the inflation lax, to their dollar-denominated equivalents, which do not.

39

The specification in equation (1) excludes consideration of capital flight and repatriation. To the extent that these are strongly related to inflation and exchange rate movements, their exclusion could lead to biased parameter estimates.

40

Stationarity of a series implies that graphs of a realization of a lime series over two equal-length time intervals should exhibit similar statistical characteristics. Stationary series have a tendency to return to their original value after a random shock; the mean and variance of such a series do not change with the passage of time.

41

Integration is the representation of a process as a sum of past shocks. A process is said to be integrated of order d(I(d)) if after differencing d times the resulting process is stationary (denoted I(0)).

42

Nonstationary unit root processes are characterized by lime paths that exhibit trend movements, and the paths of such series would usually be expected to diverge from their original value over time. Variables with unit roots are characterized by fluctuations around a stochastic trend, with shocks leading to permanent movements in the series away from trend. However, if there are strong long-run linkages between a group of individual series so that a linear combination of them is stationary (stable), then the series are said to be cointegrated.

43

A problem with the Box-Jenkins approach of differencing the data to overcome the spurious regression problem is that such a procedure disregards the important long-run relationships among the levels of the series, which are commonly derived from economic theory.

44

Note that data with a unit root will always be nonstationary and are denoted I(1). Nominal GDP and other macroeconomic time series are typically not stationary, owing to the upward trend in the series.

45

OLS-based static cointegrating regressions involving I(1) variables are asymptotically consistent, regardless of measurement error or simultaneity bias. This is the “super-consistency” result of Stock (1987)—see also McDermott (1990).

46

The inclusion of a deterministic dummy variable in the cointegrating regression would be expected to alter the true critical values of the cointegration test. However, the civil war years of 1975–90 represent half of the 1964–93 sample period, a fraction that also coincides with the global minimum of the function that corrects the asymptotic critical values.

47

The Phillips-Ouliaris test is preferable to the augmented Dickey-Fuller tests and Durbin-Watson tests more commonly reported in the literature because it is robust to a wide variety of serial correlation and time-dependent heteroscedasticity. Similar advantages are found for the Phillips-Perron test for stationarity, when compared with alternative tests. The advantages of the Phillips-Ouliaris test are even greater when the researcher, from economic theory, has knowledge of the appropriate variables of the cointegrating relationship.

48

Both FCD$ and y$ are valued in dollars in equation (4), and at end-1993 about 70 percent of M2 was composed of dollar-denominated bank deposits.

49

The only domestic interest rate series available for the entire 1964–93 sample period is the Lebanese discount rate (DISRLBN), and the LIBOR on six-month deposits (LIBOR6A) was used to measure domestic returns on U.S. dollar deposits. Other rates of return were not used owing to a lack of data; International Financial Statistics data on rates of return on government bonds are available only from 1979, and interest rates on term deposits only from 1982. The variable RETFCD, which represents the log of the LIBOR6A rate less the depreciation of the LL/USS exchange rate, was also constructed and tested in all regressions. It was not accepted as an appropriate explanatory variable of money demand for any of the five monetary aggregates.

50

Work by Amu and others (1991) also recommends that examinations of money demand take account of the impact of innovations in the process of financial intermediation.

51

The Lebanese financial system between 1964 and 1993 has been characterized by growing dollar-based monetization; the BDL's introduction and regular use in the 1980s of treasury bills as a new type of borrowing instrument resulted in a shift in portfolio preferences; and a gradual shift to dollar-based lending for domestic investment occurred, each of which took time for the banking public to adapt to. Muscatelli and Papi (1990) argue that whether trend variables are necessary in money demand equations is essentially an empirical question—one that has been answered in the affirmative here by the Phillips-Ouliaris test.

52

The failure to find a stable income velocity of M2, along with findings of stationary adjusted income velocities for the components of Lebanese M2 (FCD$ and M2LL), indicates that dollars are not likely to be close substitutes for pounds, particularly if, as is likely, foreign currency deposits are used primarily as a store of wealth.

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Cover Economic Dislocation and Recovery in Lebanon
Economic Dislocation and Recovery in Lebanon
Author:
Mr. Jose Martelino
,
Mr. S. Nuri Erbas
,
Mr. Adnan Mazarei
,
Ms. Sena Eken
, and
Mr. Paul Cashin
ISBN:
9781557754592
ISSN:
0251-6365
Publication Date:
01 Dec 1995
Publisher:
International Monetary Fund
DOI:
https://doi.org/10.5089/9781557754592.084
Pages:
71
Chart 9

Velocity of Monetary Aggregates
Chart 10

Monetary Aggregates

(In millions of Lebanese pounds)
Chart 11

Composition of Total Liquidity

(In percent)

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