Article

The Influence of Social Security on Household Savings: A Cross-Country Investigation

Author(s):
International Monetary Fund. Research Dept.
Published Date:
January 1980
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The importance of social security revenue and expenditure as instruments of fiscal policy is on the rise in a number of developed and developing countries. Pressures are mounting to expand protection for old age and against current risks, especially through liberalized health-care and income-maintenance programs, often in the face of inadequate growth in revenue raised chiefly through payroll taxation. These pressures are brought about by the emergence of a more equity-oriented social philosophy, as well as by various demographic and economic trends, the main ones being the increase in the proportion of retired to employed persons and accelerated inflation. In many countries, the ensuing rise in the need for resources cannot be met with relatively inelastic social security revenue, given regressive payroll taxes and nearly full tax coverage of the working population.

To cope with these developments, governments must consider a number of policy options regarding changes in the structure and size of social security benefits, on the one hand, and in the sources of financing, on the other hand. In weighing the options, they have to take into account the potential repercussions on income distribution, stability, and the allocation of resources. With respect to the latter, a major question relates to the choice by households—in their dual capacity as beneficiaries and taxpayers—between savings and consumption in response to social security. But, despite the sizable literature on savings and consumption behavior, the answer cannot be provided through theoretical deduction alone. Rather, it is necessary to ascertain empirically whether social security schemes tend to encourage, discourage, or leave unaltered the savings propensity of households. In view of conflicting evidence reported in earlier research on the subject, this paper attempts to shed further light on the impact of social security on variations among countries in savings propensity of households, by introducing certain methodological refinements and using richer and more recent data.

To begin with, a model of long-run household savings consistent with the principal hypotheses of savings behavior is formulated. Then, appropriate social security variables are incorporated in the model. The theoretical sections are followed by a survey of past research on the cross-country relationship between social security and savings. In the remainder of the paper, the influence of social security benefits and taxes on household savings is estimated separately across industrial and developing countries for the period 1969-71. The concluding section presents a summary and discussion of some policy implications.

I. A Model of Household Savings

For any meaningful test of the influence of social security schemes on intercountry differences in savings propensities, it is necessary to adopt a widely accepted theoretical explanation of household savings activity. Clearly, the explanation must be consistent with the major hypotheses of long-run savings behavior, be able to capture international variations in such behavior, and be amenable to estimation on existing statistical information. Whereas adequacy under the last two criteria will become apparent in later sections, the affinity of this model with the accepted hypotheses is discussed next.

A model of household savings that appears to meet these characteristics can be stated in the structural form

where for any given country

SH=real household savings per capita
YH=real household disposable income per capita
G=growth rate of labor productivity
NRR=aftertax real rate of return on household savings
WH=real stock of nonhuman wealth per capita
RE=real flow of corporate retained earnings per capita
INQ=index of income inequality
LPA=labor-force participation of the aged
DA=dependency ratio for the aged
DM=dependency ratio for minors
LEA=life expectancy of the aged

Equation (1) depicts, in the absence of social security schemes, the average propensity of households to save (by country, over a number of years) as a function of economic, social, and demographic variables. It indicates that the savings rate may rise across countries above a constant value (1 > α0 ≥ 0) in response to an increase in such economic determinants as per capita income (α1 ≤ 0), the growth rate of productivity (α2 ≥ 0), or the rate of interest (α3 ≥ 0), and to a decline in wealth (α4 ≤ 0) or in corporate retentions (α5 ≤ 0). As for socioeconomic variables, an increase in the inequality of income or a contraction in the labor-force participation of the aged may lead to a higher savings rate (α6 ≥ 0 and α7 ≤ 0, respectively). Further, the savings propensity is likely to fall as a result of demographic forces in the form of an increase in the ratio of the aged or of minors to working-age population (α8 ≤ 0 and α9 ≤ 0) or of a decline in life expectancy at retirement age (α10 ≥ 0). Accordingly, the model can accommodate any of three prototypes for which we express no preference a priori: the absolute-income hypothesis, the permanent-income hypothesis, and the life-cycle hypothesis.

The absolute-income hypothesis, derived from the Keynesian consumption function, states that savings is determined by the level of income. Moreover, household savings is negative at zero income (α1 < 0), and the marginal propensity to save is always positive and larger than the average propensity to save (1 > α0, > SH/YH) at any income level. The other determinants given in equation (1) are omitted from the original form (so that α2, …, α10 = 0), although the hypothesis does assign a minor role to the interest rate as well as to certain demographic variables.1

According to the permanent-income hypothesis, developed by Friedman partly on the basis of the Fisherian theory of interest, long-run savings is a constant proportion of long-run income. This proportion is in turn determined by the interest rate, the ratio of nonhuman wealth to income, and various social and demographic factors.2 Thus, in terms of equation (1), the savings ratio is affected negatively by the wealth ratio (α4 < 0) and the dependency ratios (α8 < 0, α9 < 0), and positively by the interest rate (α3 > 0)3 and the life expectancy of the aged (α10 > 0). The influence of other characteristics of households may be reflected in the intercept, but the remaining variables shown in equation (1) are excluded from the hypothesis (α1, α2, α5, α6, α7, = 0).

The life-cycle hypothesis, introduced and refined by a number of authors, also in the Fisherian tradition, provides perhaps the most versatile microeconomic explanation of savings behavior.4 In essence, it is based on the assumption that an individual accumulates wealth in his working years to compensate for the lack of earnings during retirement and thus to maintain a stable stream of consumption. Hence, under static conditions and barring intergenerational’ transfers, the net lifetime (long-run) sayings of the individual is zero; likewise, there is no net aggregate savings (α0 = 0).5 However, in a growing economy, the long-run aggregate savings/income ratio becomes positive as each successive generation enjoys a higher income from which to save. Thus, in a dynamic context, the savings ratio is affected by the rate of growth of productivity, as well as by variations in the distribution of lifetime between work and retirement and in the distribution of the population between workers and dependents. In terms of equation (1), the average propensity to save increases owing to a rise in the growth rate of productivity (α2 > 0) or the life expectancy of the aged (α10 > 0), or owing to a fall in the stock of wealth (α4 < 0), the labor supply of the aged (α7 < 0), or dependency ratios (α8 < 0, α9 < 0).6

Other less frequently listed determinants of the household savings ratio, under the life-cycle hypothesis, are the interest rate, corporate savings, and income distribution.7 The positive effect of a rise in the interest rate (α3 > 0) reflects the substitution of future for present consumption, dampened by any increase in corporate savings taken into account by shareholders in their personal savings decisions (α5 < 0). A similar positive response to a rise in the unevenness of the distribution of income (α6 > 0) is based on the trade-off between equity and efficiency, as relatively high savings propensities may tend to be associated with high levels of disposable income.8

In the life-cycle approach, there is no place for the absolute level of income as an explanatory variable (α1 = 0). Yet in an extension of the hypothesis, it has been suggested that the savings ratio is influenced indirectly by income through the labor-force participation of the aged.9 According to this view, individuals will tend to lengthen their work life if their real income is relatively low, so as to generate sufficient purchasing power to complement leisure with an adequate level of consumption during the retirement years. Also, higher life expectancy at normal retirement age will prompt them to prolong participation in the work force, everything else being equal. Thus, labor-force participation of the aged is explained by the relationship

indicating the impact of income (β1 > 0) and of life expectancy of the aged (β2 > 0), above a constant rate (β0 > 0), whereby the labor-force participation variable is to be treated endogenously in the savings equation (1). In sum, equations (1) and (5) comprise the analytical framework adopted to examine the link between social security and savings behavior.

II. The Role of Social Security

Individuals are associated with social security schemes as recipients of benefits, on the one hand, and as contributors, or rather as taxpayers, on the other hand. For purposes of this inquiry, namely, to assess the impact of these schemes on household savings, the benefits can be classified in three basic categories: old-age transfers, transfers related to current contingencies, and loans. Although normally the beneficiaries must be enrolled under the scheme, in some countries certain benefits are provided universally or upon meeting certain requirements.

Old-age transfers consist predominantly of pensions disbursed periodically to eligible retirees—in some countries qualifying automatically at a specified age, and in others subject to a retirement test, an income test, or a means test—in an amount more or less related to covered preretirement wages. Alternatively, in a few developing countries, a lump-sum payment is made to each retiree from a provident fund (normally operated on a funded rather than a pay-as-you-go basis), in an amount equivalent to past tax contributions into the fund plus accrued interest. Additional cash benefits are provided under old-age programs upon disability or death of eligible workers or retirees.

Other social security transfers to eligible households cover (fully or partially) medical contingencies, unemployment, work injury, and/or indigence. These benefits, in cash or in kind, are provided either to the entire population, or to enrolled workers and their dependents, or to individuals who qualify under a means test.

The third category of benefits, available in certain developing countries, involves the extension of loans to enrolled individuals—in some instances, limited on the basis of seniority or a means test. These loans are provided for housing or for other personal needs (such as expenditure on education, marriage, or certain durable goods).

Among the various forms of financing social security schemes, individuals are directly aware of payroll tax payments, which alone in many countries confer eligibility for benefits—fostering the popular belief that payroll taxes for social security are tantamount to insurance premiums or pension fund contributions. Other less common sources of financing, such as general revenue or revenue from earmarked taxes, are provided by taxpayers at large.

Social security schemes may influence the household savings ratio via three effects: the income effect, the wealth effect, and the retirement effect.10 In the framework of equations (1) and (5), any change in benefits or payroll taxes has an income effect on the savings ratio by altering the average level of disposable income and the degree of income inequality. This impact is contingent on the extent to which, over the long run, benefits raise disposable income while payroll taxes reduce it, from the level that would obtain otherwise, for households in different income brackets. In the event that parameters α1, α4, α5, α6, and β1 turn out to be zero, the income effect vanishes. Also, insofar as benefits move closely with tax payments (as tax proceeds are used to finance benefits in unfunded or pay-as-you-go schemes) and as they have similar incidence on households across income categories, there is no income effect even in the presence of nonzero parameters.

The wealth effect indicates the direct savings response of individuals to expected future benefits. It is negative if individuals feel that they need to accumulate less wealth for retirement and current contingencies (such as sickness and unemployment) and thus reduce their desired ratio of savings to income, as a result of the protection afforded by social security. Alternatively, the effect is positive if the introduction of social security educates individuals as to the need to insure themselves for old age and against current risks, thereby raising their desired savings ratio.11 The net wealth effect may be either negative or positive, depending on the relative strength of the individual’s perception of the substitutability between expected social security benefits and savings, and of his effort to complement such benefits with additional savings because of the educational impact.

In addition, an indirect repercussion of social security benefits on the savings ratio may take place through the retirement effect. The prospect of receiving an old-age pension or lump-sum transfer may encourage individuals to retire sooner than they would in its absence. Hence, the retirement effect on savings is expected to be positive, reflecting the increased savings ratio brought about by the benefit-induced reduction in the labor supply of the aged and by the consequent need to provide for the longer retirement period.

The wealth and retirement effects are incorporated into the savings model by modifying equations (1) and (5), respectively, as follows:

and

where the new variables are

SSP=social security pension per aged person
SSF=social security lump-sum transfer (from provident fund) per aged person
SSO=other social security transfers per capita
SSL=social security loans per capita
AGE=age of social security system

Parameters α11 through α15 reflect the net wealth effect, and β3 through β5, multiplied by α7, indicate the retirement effect. As discussed earlier, the net wealth effect of the rate of each major type of benefit can be positive or negative (α11, α12, α13, α14, ≷ 0), whereas the retirement effect of the old-age benefit rates cannot be negative (α7β3, α7β4 ≥ 0) on the savings ratio—where benefit rates are defined as the proportion of per capita benefit to per capita disposable income. But these effects are qualified by variations in the age of social security schemes across countries. The educational impact of social security on savings habits may register with a lag; analogously, individuals may delay the reduction of savings or the acceleration of retirement induced by increased benefits. In contrast, it can also be argued that the wealth and retirement effects may wear off as the scheme matures, if individuals no longer believe that they will receive an adequate old-age benefit stream upon retirement—especially in countries where the size of benefits is not corrected sufficiently for inflation or where the number of taxpayers supporting the scheme shrinks in relation to the number of beneficiaries. Thus, the possible direction of the relationship between the age of the system and either the savings ratio or the labor-force participation of the aged is ambiguous (α15 ≷ 0, α7β5 ≷ 0).

Alternatively, savers may adjust their desired wealth/income ratio to the social security tax liability that they actually incur (that is, payroll tax payments of employers and employees) rather than to the social security benefits that they are entitled to receive in case of a contingency. To take into account this view of the wealth effect, equation (6) must be replaced by

where the effective payroll tax rate is calculated from the ratio of tax payments, SST, to gross wages and salaries, WS. Similarly, equation (7) is to be changed if individuals near retirement age are more apt to base their choice between work and leisure on the tax proxy for the expected stream of old-age benefits, so that

The parameters of the effective tax rate represent the wealth effect 11 ≷ 0)12 and the retirement effect (α7β3 ≥ 0), qualified by the age of the system (α12 ≷ 0, α7β4 ≷ 0).

A criticism that can be leveled at the foregoing analysis of wealth and retirement effects of social security is that it ignores the role of the extended family as a social security institution in certain countries. Gainfully employed individuals provide financial support to the retired, invalid, and sick members of the family in the expectation that they in turn will be similarly protected upon reaching old age or becoming invalid. This arrangement, which operates much like an unfunded social security scheme, may explain in part the relatively depressed savings ratio in a number of developing countries, implicitly casting doubt on the general applicability of the life-cycle hypothesis. However, notwithstanding the diminished importance of the negative wealth effect and the retirement effect in the presence of an extended family structure, the positive wealth effect reflecting the educational factor may still hold in these countries.

III. Review of Past Studies

There have been relatively few studies on the relationship between social security and savings behavior across countries—in contrast to the more extensive research based on U. S. time series.13 In the first attempt to analyze such a relationship, Aaron (1967) adopted a household savings function broadly in accord with the life-cycle hypothesis. Although he was concerned primarily with explaining social security expenditure (in terms of the age of the system and household savings, inter alia), he also explored the effect of the ratio of such expenditure to national income, the age of the system, the old-age benefit rate, and the growth rate of national income on the household savings ratio. Ordinary least-squares estimates on a cross section of 19 countries (most of them developed) for 1957 yielded a significant positive coefficient for the age of the system and a negative one for the ratio of social security outlays. But re-estimation by Pechman, Aaron, and Taussig (1968, Appendix D) on 1960 data for practically the same sample failed to confirm the significance of the inverse relationship between the social security expenditure ratio and the household savings ratio. In addition, they estimated, also on 1960 data, the labor-force participation of the aged as a function of the proportion of the population over retirement age, per capita national income, and the ratio of old-age benefits per aged person to per capita wages in manufacturing, all of which displayed the expected significant negative coefficients.

On the basis of the most comprehensive version of the life-cycle hypothesis, Feldstein (1977) tested the impact of old-age benefits and other characteristics of social security schemes with a two-equation model—which to a large extent has inspired our model. The two endogenous variables—the private savings ratio (including corporate savings) and the labor-force participation of the aged—are determined by the following exogenous variables: growth rate of real private national income, dependency ratios, real per capita national income, life expectancy of the aged, old-age social security benefit per aged person as a ratio of per capita income, age of the scheme, and the ratio of corporate retained earnings to national income. Alternatively, in an effort to identify more precisely the effects of old-age benefits, the benefit rate per aged person was replaced by the benefit rate per old-age pensioner, the coverage rate of aged persons, and the presence of a retirement test. Feldstein (1977) used averages from national income statistics for the years 1954-60 and social security data for 1958-60 for a sample of 12 developed and 3 developing countries to measure the variables. The model was estimated by applying single-equation as well as simultaneous-equation techniques: weighted ordinary least squares on the structural and reduced form equations, and weighted two-stage least squares—the weights consisting of the respective country’s population.

On the whole, Feldstein’s (1977) results gave support to the life-cycle hypothesis. The coefficients of the following exogenous variables of the savings function were significant (although indicating sensitivity to the inclusion or omission of particular variables) and carried the indicated signs: positive for the growth rate, the life expectancy of the aged, and the retirement test; and negative for the dependency ratios, various measures of the benefit rate, the per capita income level, and its inverse. As to the equation for the labor-force participation, varying degrees of significance were obtained with positive coefficients for per capita income and its inverse, and with negative coefficients for the components of the benefit ratio and, in some instances, the retirement test. The significance of the reciprocal of the income variable declines markedly (while that of other variables falls slightly) upon removal of the absolute level of income from either equation.14 The coefficients of the age of the social security system and the corporate retention ratio are insignificant in virtually all regressions.

In another recent study, Barro and MacDonald (1979) tried to explain the ratio of consumer expenditure to gross domestic product (GDP) with a pooled sample of annual observations for the period 1951-60 on 16 developed Western countries. They employed a one-equation model, also based essentially on the life-cycle hypothesis, in which the consumption ratio is determined by the growth rate in real per capita GDP, the reciprocal of real per capita GDP, the ratio of government expenditure to GDP, the proportion of old-age persons in total population, the old-age benefit rate, the unemployment rate, and the ratio of last year’s to the current year’s per capita GDP. Aside from the variables already discussed, the last two are supposed to take into account cyclical fluctuations, while the parameter of the government expenditure ratio is meant to capture the household’s perception of a proportion of government outlays as constituting an equivalent amount of imputed private consumption or saving.

Barro and MacDonald (1979) presented four sets of ordinary least-squares estimates. In one instance, the consumption function was weighted by the square root of population, in another it was not so weighted; further, each of these regressions was run alternatively with a constant term or with country dummy variables. In the equations with the intercept, the coefficients of all variables (except the ratio of past to current GDP) were at least marginally significant: carrying a negative sign for the government expenditure ratio, the growth rate, and the old-age benefit rate; and a positive sign for the proportion of elderly in the population, the unemployment rate, and the reciprocal of per capita GDP. Not surprisingly, the overall performance of the equation improved sharply upon introduction of country dummies (and exclusion of the growth rate), retaining the significance of the aforementioned variables although with reversal to a negative sign for the proportion of elderly and to a positive sign for the old-age benefit rate. The presence of dummies also contributed to a reduction in the serial correlation of residuals, which seems to be particularly severe otherwise (as revealed by the low value of the Durbin-Watson statistic). As for differences between weighted and unweighted regressions, the latter provided worse fits than the former, as expected.

Before interpreting the results, one should consider several problems associated with the measurement of variables and the estimation of the foregoing models. Measurement difficulties are likely to arise with respect to savings, per capita real income, and social security variables. The savings variable should be limited to the household sector,15 without including the corporate sector, so as to test rather than assume the transparence of the corporate veil in the personal choice between savings and consumption, and to acknowledge the possible distinction in the motives to save of households and corporations. Instead of translating per capita real income into U. S. dollars at the official exchange rate, as has been done in all the preceding models, the correct approach would be to adjust real income for differences in the purchasing power of currencies. Although one may criticize Aaron (1967) for using a crude social security variable, namely, the global social security expenditure ratio, it may also be questionable to break down, as Feldstein does, the benefit rate on the basis of rather soft data on the coverage ratio and the retirement test (obtained from statutory information that vaguely reflects actual practices), while ignoring the income and means tests used in some countries.

The estimation problems involve identification, specification, heteroscedasticity, distinction between long-run and short-run effects, and sample homogeneity. The first three are econometric issues, whereas the last two may affect only the interpretation of results. Estimation of most savings functions, including the ones discussed here, ignores the possible bias introduced by the difficulty of identifying such a function as distinct from the investment function, which also contains the growth rate and the interest rate as explanatory variables. This bias can be corrected through simultaneous equation estimation, but outside evidence suggests that the extent of the bias is not likely to be important.16

Another source of bias present in these models is the possible specification error brought about by the omission of certain key variables, notably, the interest rate, income distribution, the stock of wealth, and social security variables other than the old-age benefit rate. This error can result also from the linear form of savings function, given the generality of the underlying theory. Incidentally, the severe serial correlation reported by Barro and MacDonald (1979) may be symptomatic of the exclusion of certain exogenous variables or the choice of an incorrect functional form.

The weighting procedure employed by some authors to enhance the relative importance of larger countries in the regressions, by population size, may lead to heteroscedastic disturbances that (much like serial correlation) tend to inflate the significance of the estimated coefficients upon application of conventional tests of statistical significance.

Several studies fail to isolate long-run from short-run effects of social security. The Aaron (1967) and Pechman-Aaron-Taussig (1968) estimates, based on single-year cross sections, are undoubtedly influenced by short-run forces. Despite the explicit inclusion of certain cyclical variables, the Barro-MacDonald (1979) estimates do reflect short-run behavior, particularly in the presence of dummies that serve to eliminate intercountry variations (or shifts) among ten-year strings of country observations.

Most regressions were run on cross-sectional samples containing countries at different stages of development (although the majority are developed), under the assumption that they are drawn from the same population. A more realistic approach would have been to recognize that developed and developing countries may belong to different populations, by allowing differences in behavior to be reflected in differences in the values of corresponding parameter estimates.

Having discussed possible empirical shortcomings of previous research, one can summarize and interpret the findings on the impact of social security. In their estimates of the labor-force participation function, Pechman, Aaron, and Taussig (1968), as well as Feldstein (1977), found support for the postulated retirement effect: an increase in the old-age benefit rate may lead to a rise in household or private savings, via reduction in the labor supply of the aged. However, Aaron (1967) and Feldstein (1977) also produced evidence indicating that on balance the net negative wealth effect swamps the retirement effect, so that social security has a depressing impact on the propensity to save. Further, according to Aaron’s (1967) estimates, as the system matures, the negative influence of social security expenditure wears off. A plausible explanation is that individuals raise the savings ratio as they increasingly feel that the expected benefits will not be sufficient to provide protection against various contingencies, including old age.

Against these results, Barro and MacDonald (1979) found that increased old-age benefit rates are related to a rising savings propensity, suggesting that the retirement effect and/or the positive wealth effect more than offset the negative wealth effect. In the absence of a direct estimate of the labor-force participation equation, it is not clear whether indeed the retirement effect or a positive wealth effect, representing the educational role of social security, is the dominant influence. In sum, the conflicting findings of past studies, as well as the underlying empirical problems, warrant a fresh attempt to solve some of these problems and thus to provide more reliable evidence on the influence, if any, of social security on intercountry differences in savings propensities.

IV. Empirical Estimates of the Model

The theoretical model presented earlier is a two-equation system where the household savings ratio and the labor-force participation of the aged are endogenous. Having ruled out the application of either ordinary least squares to each structural equation17 or two-stage least squares to the savings equation,18 a third alternative was adopted to break the simultaneity. Specifically, the labor-force participation equation was substituted in the structural savings equation, and then ordinary least-squares estimation was applied to the resulting reduced form savings equation and the original labor-force participation equation. This approach is equivalent to indirect least-squares estimation, although without the advantage of being able to derive structural parameters from the reduced form coefficients.

The equations were estimated on cross sections of country averages of 1969-71 annual observations, grouped in samples of 14 industrial countries and 40 developing countries, the size of the samples being dictated chiefly by the availability of data.19 Other refinements over previous research include the measurement of per capita real income (taking into account intercountry differences in purchasing power) and the rate of growth of labor productivity (i.e., the difference between the growth rates of output and of the labor force). Further, several additional explanatory variables were introduced: various social security benefit and tax rates; the aftertax long-term interest rate on financial savings less inflation (for developing countries the inflation rate alone was used because of the lack of adequate data on interest and tax rates); and an index of income inequality consisting of estimates of the Gini coefficient. However, no attempt was made to deal with the savings/investment simultaneity and to quantify the wealth variable.

With regard to social security variables, special care was taken in selecting the appropriate economic bases to calculate effective benefit and tax rates. Four benefit rates were measured by the ratios to household disposable income of (1) old-age pensions, (2) old-age lump-sum payments (from provident funds), (3) other transfers, and (4) loans. Old-age transfers, in addition, were normalized for the proportion of the elderly in the total population. The tax rate was given by payroll taxes for social security as a ratio of compensation of employees plus entrepreneurial income. Because of insufficient data, for developing countries the tax rate could not be computed, while the benefit rates were taken as ratios to private disposable income.

Albeit unweighted and population-weighted regression estimates are shown in Tables 1 through 4, the following discussion focuses primarily on the unweighted estimates. Weighted estimates (obtained by multiplying each observation by population size) substantially raised the goodness-of-fit of the regressions, but at the cost of introducing a high degree of collinearity among several variables for all samples, and some heteroscedasticity for developing countries. Nonetheless, both sets of results have essentially the same implications regarding the effects of social security.

Table 1.Industrial Countries: Unweighted Regression Results1
Dependent Variable
Independent VariablesSH/YH (1)SH/YH (2)SH/YH (3)LPA (4)LPA (5)
Constant53.776

(46.748)
20.558

(7.481)
36.656

(10.045)
-11.786

(57.412)
17.828

(49.901)
1/YH2.294

(5.364)
4.654

(5.614)
5.362

(5.370)
G0.033

(1.900)
1.794

(0.507)
0.790

(0.516)
NRR1.486

(1.142)
0.700

(0.583)
1.215

(0.540)
RE/YH0.284

(0.438)
INQ19.800

(24.578)
DA-0.712

(0.494)
-0.601

(0.267)
DM-0.242

(0.207)
-0.157

(0.093)
-0.270

(0.102)
LEA-2.208

(2.940)
2.081

(3.360)
0.094

(2.888)
SSP/YH0.092

(0.039)
-0.178

(0.090)
SSO/YH-0.406

(0.202)
SST/CN0.292

(0.107)
-0.522

(0.225)
AGE-0.118

(0.053)
-0.086

(0.047)
-0.115

(0.127)
-0.220

(0.101)
Summary Statistics 2
R¯20.2100.7250.7590.3170.386
SEE¯4.3802.5842.4186.2425.921
X¯11.37511.37511.37512.46412.464
Table 2.Developing Countries: Unweighted Regression Results1
Dependent Variable
Independent VariablesSP/YP

(1)
SP/YP

(2)
SP/YP

(3)
LPA

(4)
Constant23.784

(5.770)
22.035

(8.821)
16.642

(2.925)
21.894

(3.730)
1/YP-0.422

(0.166)
-0.423

(0.206)
-0.363

(0.171)
0.866

(0.267)
G0.409

(0.360)
0.496

(0.428)
0.508

(0.316)
INF-0.259

(0.145)
-0.232

(0.175)
-0.290

(0.139)
INQ2.181

(9.223)
2.247

(10.631)
DA-0.014

(0.033)
-0.022

(0.037)
DM-0.047

(0.031)
-0.040

(0.042)
SSP/YP-0.004

(0.074)
-0.257

(0.095)
SSF/YP0.151

(0.168)
-0.199

(0.257)
SSO/YP0.302

(0.394)
0.426

(0.272)
SSL/YP-0.179

(1.725)
AGE-0.041

(0.092)
0.250

(0.133)
Summary Statistics 2
R¯20.3030.2260.3410.326
SEE¯4.7605.0164.6278.125
X¯15.58415.58415.58429.594

Table 1 presents the unweighted results for industrial countries. In regression (1), fitted on the basic household savings function (without the social security variables) the coefficients of the growth rate of productivity, interest rate, income inequality, and dependency ratios carry the expected signs, whereas those of the reciprocal of per capita real income and retained corporate earnings are insignificant and have incorrect signs. Regressions (2) and (3), incorporating alternatively the social security benefit and tax rates-along with the age of the system-while excluding the weakest basic variables,20 indicate a significant positive relationship between savings and old-age transfers or payroll taxation, and a negative relationship between savings and other social security transfers. Labor-force participation regressions (4) and (5) show the expected positive although insignificant coefficients for the reciprocal of income and life expectancy, and negative significant ones for the old-age transfers or payroll taxes and the age of the system.

Table 3.Industrial Countries: Weighted Regression Results1
Dependent Variable
Independent VariablesSH/YH

(1)
SH/YH

(2)
SH/YH

(3)
LPA

(4)
LPA

(5)
Constant64.085

(47.800)
12.154

(6.205)
31.335

(5.823)
198.837

(43.629)
238.263

(49.306)
1/YH-3.353

(2.140)
-1.910

(2.261)
-2.287

(2.884)
G1.552

(1.618)
2.486

(0.308)
1.037

(0.262)
NRR1.681

(0.778)
0.288

(0.395)
0.868

(0.372)
REIYH0.181

(0.233)
INQ19.514

(23.625)
DA-0.677

(0.582)
-0.470

(0.200)
DM-0.386

(0.230)
-0.136

(0.062)
-0.239

(0.066)
LEA-1.335

(2.739)
-10.568

(2.681)
-13.127

(2.955)
SSP/YH0.106

(0.023)
-0.099

(0.069)
SSO/YH-0.295

(0.146)
SST/CN0.338

(0.084)
0.072

(0.309)
AGE-0.079

(0.040)
-0.092

(0.049)
-0.427

(0.102)
-0.565

(0.095)
Summary Statistics 2
R¯20.9930.9980.9970.9860.984
SEE¯53.90031.80035.420139.500151.600
X¯511.364511.364511.364710.218710.218
Table 4.Developing Countries: Weighted Regression Results1
Dependent Variable
Independent VariablesSP/YP

(1)
SP/YP

(2)
SP/YP

(3)
LPA

(4)
Constant19.026

(6.237)
26.170

(13.364)
23.246

(1.708)
17.018

(5.784)
1/YP-0.313

(0.106)
-0.551

(0.286)
-0.438

(0.073)
1.117

(0.311)
G0.479

(0.279)
-0.049

(0.521)
INF-0.201

(0.124)
-0.255

(0.124)
-0.268

(0.091)
INQ0.754

(7.844)
5.384

(9.952)
DA-0.127

(0.169)
0.010

(0.163)
DM-0.003

(0.038)
-0.027

(0.073)
SSP/YP0.164

(0.070)
0.126

(0.039)
-0.328

(0.075)
SSF/YP0.202

(0.138)
-0.789

(0.299)
SSO/YP-1.447

(0.771)
-1.515

(0.406)
SSL/YP2.773

(1.790)
AGE-0.141

(0.140)
0.657

(0.196)
Summary Statistics 2
R¯20.9960.9970.9970.998
SEE¯67.24059.74057.760170.000
X¯367.507367.507367.5071,002.914

The response of household savings to independent variables that exhibit significant coefficients is summarized in Table 5. The first column shows the elasticity of the household savings ratio with respect to these variables in industrial countries, based on unweighted regressions (2) and (3). Estimates of the elasticity with respect to the growth rate, interest rate, and dependency ratios are broadly consistent with those reported by other authors.21 As to the social security variables, it appears that a rise of 1 per cent in the rate of old-age pensions or of payroll taxes leads to an increase of nearly ½ of 1 per cent in the savings ratio, whereas a rise of 1 per cent in the rate of other benefits depresses the savings ratio by ½ of 1 per cent. The size of the negative influence of the age of the system depends on the estimated regression.

Table 5.Elasticity of Savings with Respect to Selected Independent Variables1
Unweighted RegressionsWeighted Regressions
Independent variablesIndustrial countries 2(SH/YH)Developing countries 3(SP/YP)Industrial countries 2(SH/YH)Developing countries 3(SP/YP)
1/YP-0.19-0.44
G0.00/0.560.34/0.82
NRR0.00/0.070.00/0.09
INF-0.10-0.13
DA-1.07/0.00-0.76/0.00
DM-1.41/0.00-1.27/-0.72
SSP/YH0.470.43
SSP/YP0.000.08
SSO/YH-0.49-0.28
SSO/YP0.00-0.24
SST/CN0.340.34
AGE-0.56/0.00-0.32/0.00

These results confirm the extended life-cycle hypothesis, but they do not support the view that income inequality promotes savings propensities and that households count on corporate resources to meet their demand for wealth. The absence of a significant relationship between savings and inequality might be attributed chiefly to the crudeness of data on income distribution.22 However, the lack of a negative impact of corporate savings underscores the households’ perception that large, widely held corporations in industrial countries are independent of shareholder control, whereby households do not take into account corporate distribution policy in their decision to save.23

More important, the coefficients of the social security variables, and the corresponding elasticity estimates, suggest that in industrial countries the retirement effect overwhelms the net wealth effect of old-age transfers, while other transfers tend to exercise a negative wealth effect on savings, reflecting asset substitution. If the influence of social security is tested with the payroll tax rate instead of the benefit rates, it seems that, on balance, the schemes have a positive impact on household savings—presumably for the most part owing to the retirement effect. Yet, as the social security system grows older, the inducement to save is apparently eroded as enrolled households step up asset substitution, perhaps because of increasing belief in the adequacy of future benefits to provide protection for old age and against current contingencies.

The statistical performance of the model in the unweighted form, shown in Table 2, is on the whole weaker for developing countries, which is not surprising in light of the less reliable underlying data.24 Given the limited number of countries for which national income accounts are broken down between the household and corporate components of the private sector, it was necessary to use data on private instead of household savings and income. By the same token, it was not possible to test the relationship between the household savings ratio and the corporate retention ratio. This impediment, in itself, should not constitute a major drawback for two reasons: first, the corporate share in private savings or income is smaller in developing countries (i.e., for the dozen or so countries where the breakdown is available) than in industrial countries; second, much of the corporate activity is carried out by family-owned enterprises whose savings activity may well be regarded as an extension of household savings behavior.

The coefficients of all the basic exogenous variables display the expected signs, but those of the reciprocal of income, the growth rate, and the inflation rate alone have some significance. Among the social security variables, which—besides the old-age pension and other transfer rates—also include the rates of old-age lump-sum transfers and of loans, only the negative coefficient of the old-age pension rate and the positive coefficient of the age of the system are significant in the unweighted run (4) of the labor-force participation equation.

Reflecting to a greater extent the behavior of more populated developing countries, the coefficients of several social security variables in Table 4 are highly significant: the rates of old-age pensions and of other transfers with positive and negative signs, respectively, in weighted regression (3); the rates of all old-age transfers with a negative sign, and the age of the system with a positive sign, in regression (4).

Accordingly, the elasticity estimates (based on weighted regression results) for developing countries, shown in the fourth column of Table 5, are larger than the estimates in the second column (based on unweighted results). In either case, the elasticities of the private savings ratio with respect to social security benefit rates are substantially smaller than the corresponding estimates for industrial countries. The weighted estimates for developing countries indicate that an increase of 1 per cent in the old-age pension rate would lead to a rise of less than 110 of 1 per cent in the savings ratio, and an equivalent increase in the rate of other transfers would result in a drop of ¼ of 1 per cent in the savings ratio. No savings response to changes in benefit rates is detected in the unweighted estimates.

In contrast to the results for industrial countries, where per capita income has virtually no impact on the propensity to save, in developing countries an increase in income depresses the labor supply of the aged and raises the savings ratio. Although the latter is in line with the extended life-cycle hypothesis, it can also be interpreted as a vindication of the Keynesian view. In any event, the difference in behavior between developed and developing countries may be ascribed to the extended family structure and to the greater variance in per capita income in developing regions. Moreover, in developing countries, the inflation rate performs fairly well as an inverse proxy for the real return on financial savings, the reason being that in many developing countries nominal interest rates—often subject to discretionary ceilings—fail to catch up with rapidly rising prices, so that real interest rates decline and may become negative.

With regard to the wealth and retirement effects of social security on private savings, the unweighted regressions lend support to the null hypothesis—notwithstanding the possible income effect suggested by the significance of the income variable. At the same time, estimates of the labor-force equation indicate a potentially strong retirement effect of old-age pensions, dampened by the age of the system. The insignificant coefficient of old-age benefits in the savings equation may be taken to mean that the retirement effect is fully offset by an equally powerful negative net wealth effect.25 But perhaps a more plausible explanation is that neither effect operates, given the social security function of the extended family in many developing countries. In contrast, the weighted-regression results reveal some asset substitution in response to other than old-age transfers in countries with large populations.

All in all, the estimates of the savings function for developing countries are probably subject to considerable measurement error and to some misspecification.26 The problem stems, at least in part, from the heterogeneous nature of the sample countries, which possess diverse cultural, social, and institutional characteristics not captured by the model.27 More explicit account should also be taken of differences in economic environment among developing countries, particularly those involving variations in uncertainty and risk, which are reflected to a limited extent by the rate of inflation.

V. Summary and Conclusion

A model intended to explain differences in savings propensities of households among countries has been formulated in conformity with the absolute-income hypothesis, the permanent-income hypothesis, and an extended version of the life-cycle hypothesis. Comprised of two structural equations—the dependent variables being the average propensity to save of households and the labor-force participation of the aged—this eclectic model is especially useful in that it incorporates the effects of social security schemes. Such schemes provide protection for old age, through periodic pension payments or lump-sum disbursements from provident funds, and against current contingencies, through payments for medical expenses, unemployment, work injury, indigence, and, in some countries, the extension of loans. Although initially many social security schemes were funded, by now most of them are virtually unfunded, so that payroll taxes form the main source of revenue to finance benefits.

Social security institutions in their present form can have three distinct effects on the household savings ratio. First, the income effect consists of the possible repercussion of changes in benefit or tax rates transmitted through the average level of disposable income and the degree of income inequality. Second, the wealth effect represents the direct savings response of households to expected future benefits. Households may react either by cutting back savings if they perceive such benefits to be an adequate substitute for personal savings, or by raising their savings propensity if social security schemes enhance their awareness of the need to accumulate additional wealth as protection against various contingencies. Third, the retirement effect indicates the positive indirect relationship between old-age benefits and the savings ratio, via changes in the labor supply of the aged; an increase in old-age benefits is likely to induce earlier retirement, which in turn may lead to a higher propensity to save for the longer retirement period. Of these consequences of social security, particularly the wealth and retirement effects have attracted attention and generated controversy in the literature (as opposed to the income effect, which is likely to vanish over the long run, if benefits and taxes have similar incidence on households), and they are also the focus of this analysis.

Estimates of the model using cross sections of 1969-71 average observations, divided into samples of industrial and developing countries, suggest that social security schemes influence decisions on household savings, especially in industrial countries. In these countries, when allowance is made for a differential impact of various types of benefit, the (positive) retirement effect outweighs the (negative) wealth-substitution effect of old-age transfers, while the latter effect prevails for other transfers. An alternative test, performed under the assumption that households consider payroll taxes to be a proxy for future social security benefits, reveals that the positive retirement effect overrides the other effects. However, the results also indicate that the wealth-substitution effect strengthens as the social security system becomes older, apparently because of the gradual rise of household confidence in the system’s ability to deliver an adequate stream of benefits.

For developing countries, the statistical power of the estimates is generally lower and the nature of the findings is less conclusive. Despite the significant inverse impact of old-age transfers on the labor supply of the aged, the net effect of these transfers on savings is a draw, implying that the retirement effect and the wealth-substitution effect offset each other. But it is more probable that these effects, as well as the validity of the life-cycle hypothesis, are precluded by the extended family structure, which in many developing countries operates like an unfunded social security system. Nonetheless, transfers for other than old age seem to have a negative net wealth effect on the savings ratio in countries with a relatively large population.

From this evidence, it can be inferred that the expansion of the rate or coverage of old-age social security pensions and of payroll taxes does not have an adverse impact on household savings. If anything, such benefits and taxes are more likely to encourage savings in industrial countries—although this incentive may wear off as the social security institutions mature. There seems to be, however, substitution of other social security transfers for savings in industrial countries and more populated developing countries. Therefore, to some extent, social security can serve as an instrument of resource mobilization if the authorities invest the surplus of social security institutions (at least to the extent that they are funded) in projects with an adequate social marginal rate of return, without inhibiting private savings and investment activity.

Nevertheless, policy decisions with regard to social security benefits and taxation should be made in a broader setting. Investigation should also be undertaken of their consequences on vertical and horizontal equity (taking into account, for instance, the usual regressivity of the payroll tax across income brackets and its uneven coverage of various income sources), the country’s comparative advantage, and the choice of factor inputs, over the long run, as well as on internal and external stability over the short run.

APPENDIX

Data were collected for samples of 14 industrial and 40 developing countries from the following sources:28

The industrial countries included in the sample are Australia, Austria, Belgium, Canada, Denmark, the Federal Republic of Germany, France, Italy, Japan, the Netherlands, Sweden, Switzerland, the United Kingdom, and the United States. The developing countries are Brazil, Burma, Chile, Colombia, Costa Rica, Cyprus, the Dominican Republic, Ecuador, Egypt, El Salvador, Finland, Greece, Guatemala, Honduras, India, Ireland, Israel, Jamaica, Korea, Libyan Arab Jamahiriya, Malaysia, Malta, Mauritania, Mauritius, Mexico, Morocco, Nicaragua, Niger, Nigeria, Panama, the Philippines, Portugal, Senegal, Singapore, the Syrian Arab Republic, Thailand, Trinidad and Tobago, Tunisia, Turkey, and Venezuela.

All data refer to country observations for the period 1969-71 unless indicated with a subscript denoting the last two digits of the year. The variables used in the regressions are29

  • SH/YH = SH/(SH + CH)

  • SP/YP = SP/(SP + CH)

  • LPA = WP(+65)70/P(+65)70

  • 1/YH = [GDP/(SH + CH)]/IDP

  • 1/YP = [GDP/(SP + CH)]/IDP

  • G = In-1 [(In RDP73 - In RDP64)/9)] - In-1 [(In WP70 - In WP60)/10]

  • NRR = [LRI (1 - MYT72)] ΔCPI/CPI

  • INF = ΔCPI/CPI

  • RE/YH = (SP - SH)/(SH + CH)

  • INQ = GC

  • DA = P(+65)70/ P(20-65)70

  • DM = P(-20)70/P(20-65)70

  • LEA = LE(65)70

  • SSP/YH = [SSP/(SH + CH)] · P70/P(+65)70

  • SSP/YP = [SSP/(SP + CH)] · P70/P(+65)70

  • SSF/YP = [SSF/(SP + CH)] · P70/P(+65)70

  • SSO/YH = SSO/(SH + CH)

  • SSO/YP = SSO/(SP + CH)

  • SSL/YP = SSL/(SP + CH)

  • SST/CN = SST/(CE + EN)

  • AGE = 1970 - START

The underlying data are defined as follows (sources are shown in parentheses):

SH:household savings, in national currency units (J)
SP:private savings (including gross savings of corporations), in national currency units (A)
CH:personal consumption expenditure, in national currency units (J), (A)
GDP:gross domestic product, in national currency units (D)
RDP:real gross domestic product, in national currency units in 1975 prices (D)
IDP:index of real per capita GDP, U. S. value = 100 (E)30
P:total population, in number of persons (D)
P(+65):P(20-65): population between ages of 20 and 65, in number of persons (C)
P(20-65):population between ages of 20 and 65, in number of persons (C)
P(-20):population between ages of 20 and 65, in number of persons (C)
WP:population of age 20 or younger, in number of persons (C)
WP(+65):working population of age 65 or older, in number of persons (C)
LEA(65):life expectancy at age 65, in number of years (I)
LRI:nominal interest rate (net of withholding tax where applicable) on long-term bonds, or savings or time deposits (F)
MYT:marginal income tax as a proportion of gross earnings of a married couple with two children where husband earns an average production worker’s wages (G)
CPI:consumer price index, 1975 = 100 (D)
GC:Gini coefficient of income inequality (H)31
SSP:expenditure of social security institutions on old-age pension benefits, in national currency units (B)
SSF:expenditure of provident funds on old-age lump-sum benefits, in national currency units (B)
SSO:expenditure of social security institutions on other benefits (i.e., other than SSP and SSF), in national currency units (B)
SSL:gross flow of personal and housing loans extended by social security institutions, in national currency units (L)
SST:payroll tax revenue (from employee and employer contributions) of social security institutions, in national currency units (B)
CE:compensation of employees, in national currency units (J)
EN:entrepreneurial income, in national currency units (J)32
START:year in which basic social security legislation came into force (K).
REFERENCES

    Aaron, Henry,Social Security: International Comparisons,” in Studies in the Economics of Income Maintenance, ed. byOttoEckstein,The Brookings Institution (Washington, 1967), pp. 1348.

    Ando, Albert, and FrancoModigliani,The ‘Life Cycle’ Hypothesis of Saving: Aggregate Implications and Tests,American Economic Review, Vol. 53 (March1963), pp. 5584.

    Barro, Robert J., and Glenn M.MacDonald,Social Security and Consumer Spending in an International Cross Section,Journal of Public Economics, Vol. 11 (June1979), pp. 27589.

    Cagan, Phillip,The Effect of Pension Plans on Aggregate Savings: Evidence from a Sample Survey, Occasional Paper 95, National Bureau of Economic Research (Columbia University Press, 1965).

    Esposito, Louis,Effect of Social Security on Saving: Review of Studies Using U. S. Time-Series Data,Social Security Bulletin (May1978), pp. 917.

    Feldstein, Martin S. (1973), “Tax Incentives, Corporate Saving, and Capital Accumulation in the United States,Journal of Public Economics, Vol. 2 (April1973), pp. 15971.

    Feldstein, Martin S. (1977), “Social Security and Private Savings: International Evidence in an Extended Life-Cycle Model,” in The Economics of Public Services: Proceedings of a Conference Held by the International Economic Association at Turin, Italy, ed. byMartin S.Feldstein and Robert P.Inman (London, 1977), pp. 174205.

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    Shome, Parthasarathi, and Katrine AndersonSaito,The Impact of Social Security Institutions on Resource Mobilization and Allocation: The Asian Experience,presented at the Annual Meetings of the American Economic Association (unpublished, Chicago, August1978).

    Tanzi, Vito, and JosephAschheim,Saving, Investment, and Taxation in Underdeveloped Countries,Kyklos, Vol. 18 (No. 2, 1965), pp. 20526.

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SUMMARIES

Distributional Aspects of Stabilization Programs in Developing Countriesomotunde johnson and joanne salop (pages 1-23)

This paper reports the results of some preliminary research into the repercussions, for income distribution, of stabilization programs associated with the use of Fund resources in the upper credit tranches. In the first section, it explores the relationship between the balance of payments and the distribution of income from a theoretical perspective. Here the general concern is whether adjustment influences the distribution of income in some systematic manner; the particular concern is to delineate the conditions under which a decline in the real wage is necessary for adjustment actually to take place. Using neoclassical analysis, one finds that the ratio of the nominal wage to the price of exports must decline, but whether this involves a fall in the overall real wage depends on many variables, including the relative proportions of traded and nontraded goods in the consumer’s market basket. Second, it presents a qualitative analysis of the distributional effects of the measures that tend to be included in these programs, viz., ceilings on net credit expansion, currency depreciation, and the relaxation and simplification of exchange restrictions and controls. Finally, drawing on unpublished case studies, it discusses the distributional consequences of stabilization programs in Bolivia (1972-73), Ghana (1966-70), Indonesia (1966-74), and the Philippines (1970-76).

A Supply Framework for Exchange Reform in Developing Countries: The Experience of Sudankarim nashashibi (pages 24-79)

The paper examines the question of exchange rate determination in a developing country with a structural imbalance, quantitative restrictions on trade, and extensive public administration of prices and production. The paper argues that, under conditions of widespread price and cost distortions, the purchasing power parity theory approach to exchange reform coupled with supportive demand management policies is inappropriate in determining the extent of exchange rate adjustment needed and in reorienting the economy toward equilibrium. Drawing on Sudan’s experience over the past decade, the paper derives the cost structure of the country’s major exports and import substitutes for 1972/73 and 1976/77. Competitiveness—defined as net foreign exchange earned or saved per unit of domestic resources used in the production process after correcting for all price distortions and netting out taxes and subsidies—was derived for seven irrigated crops and three rainfed crops. The results showed that Sudan had a strong comparative advantage in cotton and groundnut cultivation in irrigated areas and in oilseeds in the rainfed areas. The cultivation of cereals in irrigated areas was found to be highly inefficient. The analysis showed that the devaluation in 1972 was necessary to restore the competitiveness of some export crops but that it had little effect on expediting the movement of exports and on correcting the structural imbalance in the economy. The devaluation was undermined by the reorientation of the cropping pattern toward inefficient crops and by expansionary fiscal and monetary policies that rapidly eroded the gains in competitiveness, as shown by the 1976/77 results. This prompted a further devaluation in 1978. A hypothetical supply curve of exportables as a function of the exchange rate was derived, and the implications of the exchange rate depreciation for income distribution in Sudan were drawn.

The paper argues that when a depreciation serves as an instrument of structural change, a reorientation of investment priorities must take place concurrently with the correction of distortions and the changes in relative prices to ensure the availability of all the supporting services that are the linchpin between an increase in production and an increase in exports. This has implications for demand management policies, which have to be applied selectively, allowing for a resurgence of investment and credit in the competitive external sector and a deflation of other sectors. Moreover, the envisaged adjustment period must be sufficient to allow for the shifts in production and consumption induced by the depreciation and its supportive policies to materialize fully.

The Optimal Basket in a World of Generalized Floatingleslie lipschitz and V. sundararajan (pages 80-100)

Recently, policymakers have been discussing the best basket peg for a country seeking stability in a world of generalized floating. This paper argues that the real exchange rate is the important policy variable and that, although exchange rate data are available daily, price data are available only after a lag. Consequently, continuous, discretionary fine tuning of the real exchange rate is impossible, and a rule is needed for fixing the nominal exchange rate so that the real exchange rate is stabilized. This rule, or optimal basket, is one that minimizes the variance of the real exchange rate about its equilibrium, while maintaining the average value of the real exchange rate close to its equilibrium over the reference period.

A solution for the optimal basket is derived. In general, the optimal weight of the currency in the basket will differ from the preassigned elasticity weight (which denotes the importance of the currency in the real exchange rate index) because the variances and covariances of exchange rates and relative prices have an important effect on the real exchange rate index. Weights in the optimal basket are chosen to maximize the contribution of this effect to stability. Various cases of the general solution are discussed, with particular emphasis on the conditions under which a single currency peg is optimal.

A numerical illustration of the solution is provided, and a comparison, in terms of relative stability, is made between an optimal basket peg and a basket peg in which simple elasticity weights are used. The question remains as to whether the variances and covariances required to derive the optimal weights are stable over time and therefore useful in setting an exchange rate rule for the future. Since these parameters are of the nature of regression coefficients, their stability can be tested and the results of such tests are presented. It is argued that insofar as these parameters may be estimated from historical data, and may be expected to be stable, the derivation of useful optimal basket weights is feasible.

Why Does the Current Account Matter?joanne salop and erich spitaller (pages 101-34)

This paper focuses on the current account as an indicator of the need for adjustment, its role in the adjustment process, and the implications of possible inconsistencies in countries’ desired current account for exchange rate surveillance. Fundamentally, the current account is the difference between the economy’s savings and investment. Hence, any criterion for determining the appropriateness of adjustment action, such as a deviation from sustainability or optimality, depends on a corresponding assessment of the levels of savings and investment that underlie the current account. While in the long run the scope for conflict over countries’ desired current account is likely to be small, mutual inconsistencies in countries’ current account targets are more likely to arise over the short run to medium-run during which the adjustment actually takes place. This is due to the relationship between the current account and the level of employment that may exist at that time. Accordingly, during a period of inadequate demand, a country may find it desirable to maintain its level of aggregate demand and employment through the current account by devaluation. The efficacy of pursuing this strategy—and, by extension, the importance of surveillance—turns on whether the Keynesian model is an appropriate characterization of the supply sector of the economy. Some of the more interesting aspects of the paper are (1) an analysis of the concept of sustainability; (2) the depiction of the supply sector in terms of a framework of rigid real wages versus rigid money wages and the implications for the need for surveillance and the desirability of global expansion; and (3) the inclusion of a survey of official views on the current account.

Estimation of the Timing Asymmetry in International Tradewilliam l. hemphill (pages 135-60)

Because of the time required for transport and for customs procedures, the exports of a country or region in a given period will not all be received and counted as imports by partners during the same period. This article presents a method for estimating the transport lag and for simulating the corresponding asymmetry in international accounts. As the timing asymmetry is shown to be a large and variable portion of the total merchandise asymmetry, the primary use of the results is to explain the discrepancy between global sums of exports and imports—especially in determining the discrepancy to be expected in a consistent set of global trade forecasts. In addition, the model provides an indirect estimate of the proportion of trade invoiced in U. S. dollars.

The availability of data imposes a choice, in estimating the lag, between disaggregation into smaller geographical regions and "disaggregation" into smaller units of time. Annual-data regressions yield plausible estimates for regional groupings of countries, but the time series are found to be extremely collinear; these results are rejected. Monthly data for the world as a whole indicate that about 3 per cent of yearly exports is not received and counted as imports until the following year—an average lag of about 0.6 month. As this estimate is rather shorter than the magnitude often assumed, some direct corroborative evidence is marshaled in its support. The global timing asymmetry is found to have varied between zero and $6 billion during the early 1970s. The proportion of trade not invoiced in U. S. dollars is estimated to be on the order of one half, but with a large margin of error—especially if the tentative finding of a decrease in vehicular use of the dollar, starting about 1974, is confirmed by future observations.

The Influence of Social Security on Household Savings: A Cross-Country Investigationgeorge kopits and padma gotur (pages 161-90)

This article examines the role of social security schemes in determining intercountry differences in the savings propensities of households. To this end, an eclectic model of long-run household savings behavior is constructed on the basis of the absolute-income, the permanent-income, and the life-cycle hypotheses, incorporating the effects of social security benefits (old-age pensions or lump-sum disbursements, transfers against current contingencies, and loans) and taxes (payroll contributions). At a theoretical level, three types of effect are identified. The income effect consists of the possible repercussion of changes in benefit or tax rates transmitted through the level of disposable income and the degree of income inequality. The wealth effect represents the decline in savings in response to a rise in benefits if households perceive such benefits to be an adequate substitute for personal savings, or the increase in savings if social security schemes enhance the households’ awareness of the need to accumulate additional wealth as protection against various contingencies. The retirement effect indicates the positive influence of old-age benefits on the savings ratio through changes in the labor supply of the aged; an increase in old-age benefits may induce earlier retirement, which in turn may raise the propensity to save for the longer retirement period.

The model is estimated on cross sections of 1969-71 average annual observations grouped in samples of industrial and developing countries. The results indicate that in industrial countries the retirement effect outweighs the wealth-substitution effect of old-age pensions, whereas the wealth-substitution effect prevails for other transfers. An alternative test, performed under the assumption that households consider payroll taxes to be a proxy for future benefits also reveals that the retirement effect overrides the other effects. However, the wealth-substitution effect appears to strengthen as the social security system becomes older. In developing countries, the net effect of the benefits is a draw, except in the more populated countries in this sample where transfers other than for old age seem to have a negative net wealth effect on the savings ratio. On the whole, the quality and scope of the data, as well as the statistical power of the regression estimates, are weaker for developing than for industrial countries.

From this evidence, it can be inferred that the expansion of the rate or coverage of old-age pensions and of payroll taxes does not have an adverse impact on household savings. If anything, such benefits and taxes are more likely to encourage savings in industrial countries—although this incentive may wear off as social security institutions mature. There seems to be, however, a substitution of other social security transfers for savings in industrial countries and in more populated developing countries.

RESUMES

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Mr. Kopits, Senior Economist in the Tax Policy Division of the Fiscal Affairs Department, holds degrees from Georgetown University. He is also professorial lecturer at the Johns Hopkins School of Advanced International Studies. Previously, he was on the staff of the Brookings Institution and the U.S. Treasury Department.

Ms. Gotur, economist in the Tax Policy Division of the Fiscal Affairs Department when this paper was prepared, is currently a doctoral candidate in economics at the George Washington University.

The authors are grateful to Harry Grubert and to several colleagues in the Fund for useful comments.

The fundamental Keynesian equation is given by

where S and Y denote aggregate (or per capita) savings and income, respectively. However, Keynes (1936, Chs. 8 and 9) discussed a number of cultural and demographic determinants that he called “subjective” factors, as distinct from the absolute level of income and the interest rate, regarded as “objective” factors.

For the theoretical underpinning of the model, see Fisher (1930, Part III). Although Friedman focused on the determination of short-run consumption or savings, he explained (1957, Ch. 2) the permanent component thereof with the general form

which can be expressed by the linear relationship

where S denotes permanent savings, R the interest rate, W nonhuman wealth, U social and demographic variables that determine the utility function of households (age, family composition, etc.), and Y permanent income. The parameter values are k0 > 0, k1 0, and k2 ≷ 0.

For a derivation and test of the substitution effect of the interest rate (at a constant level of permanent income), see Wright (1969).

The hypothesis was first proposed by Harrod (1948, Lecture 2). For derivations, see Ando and Modigliani (1963), Tobin (1967), Modigliani (1970), and the recent extension by Feldstein (1977).

In the event of a steady rise in intergenerational transfers through bequests, the parameter α, > 0 would obtain.

The role of the growth rate and dependency ratios has been analyzed and tested by Leff (1969), Modigliani (1970), and Feldstein (1977); yet, unlike others, Modigliani substituted the growth rate of labor productivity for the overall economic growth rate. In a recent elaboration of the hypothesis, Feldstein (1977) argued that, in addition to these determinants, individuals would raise their target wealth/income ratio (or equilibrium savings ratio) in view of accelerated retirement from the labor force and of extension in the retirement period owing to increased life expectancy.

Following Harrod (1948, Lecture 2), Modigliani (1970) and Feldstein (1973) highlighted the inverse relationship between corporate and household savings. Tobin (1967) emphasized the influence of the interest rate, and Modigliani (1970) tested the effect of the functional distribution of income in savings behavior, while acknowledging the problems associated with his measure of such distribution in terms of the ratio of nonwage income to total private income before taxes and transfers.

For a contrary view, namely, that a more equitable income distribution need not lead to a long-run reduction of savings (whereby α6 = 0) in an international context, see Tanzi and Aschheim (1965).

For a theoretical derivation of the wealth and retirement effects of old-age benefits, see Feldstein (1977).

Cagan (1965) observed that, similarly, pension plans may induce households to recognize the need to save more for the retirement years.

As against equation (6), where the wealth effect is allowed to differ by major type of benefit, equation (8) does not discriminate by type of benefit or contingency, since normally all benefits are financed from the same payroll tax.

See the recent survey by Esposito (1978). The only studies of other individual country experiences, by Shome and Saito (1978) and by Wallich (1978), relate to five Asian countries and to Chile, respectively.

Feldstein (1977) failed to provide sufficient justification for including both the per capita income and its reciprocal in the reduced form savings equation and the labor-force participation equation.

Ideally, in this context, savings should include additions to the stock of durable goods.

See the results of the experiment undertaken by Modigliani (1970, pp. 206-209).

Despite the appearance of a recursive system that would have justified the use of ordinary least squares, we found that the error terms of the savings equation and the labor-force participation equation, estimated by two-stage least squares, are highly correlated.

This approach had to be rejected because of the relatively small samples, particularly for industrial countries.

A description of the data is provided in the Appendix.

Namely, excluding variables whose coefficients are smaller than their corresponding standard errors, to minimize the loss in degrees of freedom.

See, for example, the elasticity estimates given in Leff (1969) and Wright (1969).

These data have been compiled by Jain (1975) from a large variety of primary sources.

Alternatively, part of the explanation may lie in the relatively smaller proportion of shareholders in most countries other than the United States.

For a survey of empirical evidence regarding savings behavior in developing countries, see Mikesell and Zinser (1973).

Support for this interpretation was provided by a two-stage least-squares estimate of the savings equation, in which a significant negative coefficient was obtained for the labor-force participation variable.

Indirect evidence of misspecification is provided by relatively low values of the Durbin-Watson statistic, oscillating between 1.4 and 1.7 for regressions (1) through (3) in Tables 2 and 4.

The introduction of four regional dummies (for Europe, Latin America, East Asia, and the remaining countries) improved the fits and raised the Durbin-Watson statistic, but without materially altering the results.

Complete information on the sources is provided in the References.

All the variables except INQ, LEA, and AGE have been premultiplied by 100.

As the original index refers to 1970, the 1969 and 1971 values were calculated with the growth rate of real per capita GDP for the years 1969-70 and 1970-71, respectively, from (D).

For several countries, for which coefficients were not available, the coefficients of comparable countries were used.

Includes withdrawals from the entrepreneurial income of quasi-corporate enterprises for Austria, Belgium, Chile, Denmark, the Federal Republic of Germany, Panama, the Philippines, and Portugal.

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