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Interest rates, saving and

investment: Evidence from IndiaPremaChandra Athukorala

a

aDepartment of Economics and Australia South Asia Research

Centre, Research School of Pacific and Asian Studies,Australian National University, Canberra, ACT, 0200, Australia

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To cite this article: PremaChandra Athukorala (1998): Interest rates, saving and investment:

Evidence from India, Oxford Agrarian Studies, 26:2, 153-169

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Oxford Development Studies, Vol. 26, No. 2, 1998 153

Interest Rates, Saving and Investment:Evidence from India

PR E MA -C H A N D R A A T H U K O R A L A

A B S T R A C T The role of interest rates in the process of economic development is examinedthrough an empirical inquiry into the interest rate-saving-investment nexus in the Indianeconomy during the period 1955-95. The results aregenerally in support of the financialliberalization school of thought. Higher real interest rates seem topromote both financial andtotal savings, andstimulate private investment. On the investment side, thecombined salutaryeffect of interest rate increases operating through increased debt intermediation and self-financed capital accumulation outweighs thedirect cost effect oninvestment. Overall, the studycasts doubt on the robustness of results coming from the vast cross-country literature on thesubject and calls for systematic time-series analyses covering a variety of country situations toinform the on-going policy debate.

1. IntroductionThe role of interest rates in theprocess of economic development haslong been at thecentre of the debate on economic policy reforms in developing countries. Until the1970s interest rate policy inthese co untries was primarily guided bythe Keynesian viewthat interest rates should bekept low inorder topromote investment. Ashot across thebows of this view came from McKinnon (1973) and Shaw (1973), who forcefullyargued that in the context of the typical developing economy, financial repressionretards (rather than promotes) growth, byboth reducing thevolume of investible fundsand interfering with theefficient allocation of such funds. Despite counter argumentsby economists of Keynesian persuasion and structuralist development economists(Davidson, 1986;Thirlwall, 1976;Taylor, 1983), the McKinnon-Shaw "financialrepression paradigm" soon became the neworthodoxy of financial sector reforms indeveloping countries. In response to this shift in conventional wisdom, many develop-ing countries have experimented with financial liberalization reforms since the late1970s, mostly as part of IMF-World Bank sponsored liberalization/stabilization pro-grammes.

In line with this policy emphasis, a sizeable body of empirical literature hasat tempted to quantify theimpact of interest rates onsaving andinvestment. Given theP. Athukorala, Department ofEconomics and Australia South Asia Research Centre, Research School of Pacificand Asian Studies, Australian National University, Canberra ACT0200, Australia.I have received valuable comments from many individuals, comprising a list that is too long to acknowledgehere. However, special thanks go to MaxCorden, George Fane, Ross McLeod, Jayant Menon, James Riedel,Kunal Sen, Tony Thirlwall, Pan-Long Tsai and an anonymous referee who provided comments that materiallyimproved this work. Of course, anyerrors are mine.

1360-0818/98/020153-17 1998 International Development Ce ntre, Oxford

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154 P. Athukoralanature of data availability, many of these studies have been conducted using cross-country time-series data, focusing largely on the total domestic saving-interest ratenexus.1 The results of these studies are at best ambiguous; while some studies comedown against any interest rate effect on savings, others find a positive but rather smallinterest elasticity (around 0.1) which is "not large enough to warrant much policysignificance" (Fry, 199 5, p . 42 ). Overall, at face value, the available evidence seemto suggest that "the financial repression paradigm is ... a kernel of truth and a vastexaggeration" (D ornbus ch & Reynoso, 1989, p. 205). However, many of these studies"do not reach econometric standards that would allow the reader to take their resultsat face value" (D eato n, 1989; Gibso n & Tsak alotos , 19 94). Quite apart from generalmethodological flaws relating to model specification and econometric procedure,there are two fundamental limitations that make results from any cross-countrystudy on this subject rather dubious. First, cross-country regression analysis is basedon the implicit assumption of "homogeneity" in the observed relationship acrosscountries. This is a very restrictive assumption. It is common knowledge that thereare considerable variations among developing countries in relation to various struc-tural features and institutional aspects which have a direct bearing upon the impactof financial factors on the growdi process. Second, given vast differences amongcountries with respect to the nature and quality of data, cross-country comparisonis fraught with danger (Deaton, 1989; Srinivasan, 1994). Not only the statisticalprocedures for measuring saving and investment, but also the magnitude of errorsin data in the implementation of these procedures, vary significantly amongcountries.

These considerations point to the need for undertaking econometric analysis ofsaving/investment behaviour within individual countries over time in order to build asound empirical foundation for informing the policy debate. Unfortunately, systematiccountry studies of this nature are few and far between.2 Interestingly, some of theavailable country studies have, in fact, produced large and quite significant interestelasticities with respect to saving and investment. At the same time there aresignificant variations in the magnitude of measured elasticities among countries, sug-gesting that data should not be pooled without considerable caution.This article seeks to add to the literature by systematically tracing out the interestrate-saving-investment nexus in the Indian economy using data for the period 1955-95. India seems to be a very appropriate case study of the subject at hand for thefollowing reasons. First, the Indian saving and investment database is consideredrelatively good by developing country standards, and data are available on a compar-able basis for a period of time which is adequate for systematic econometric investiga-tion (Srinivasan, 1994). Second, saving/investment performance is a key emphasis inthe policy debate in India following the structural adjustment reforms initiated in 1991(EP W RF , 1995; Athukorala & S en, 1995); bu t, there is no ha rd empirical evidence toinform this policy debate.3The empirical analysis of the present paper focuses separately on total savings,private savings and financial savings. An attempt is also made to test explicitly thenexus of financial saving and investment by separate estimation of credit supply andinvestment functions. The estimation procedure draws upon recent advances in time-series econom etrics, in an attem pt to derive robu st coefficient estimates while gu ardingagainst the possibility of uncovering spurious relations. The paper is structured in thefollowing manner. Section 2 sets out the analytical framework, drawing upon thetheoretical debate on the role of interest rate policy in developing countries. Thedatabase and the econometric procedure are described in Section 3. Results are

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Interest Rates, Saving and Investment 155presented and analysed in Section 4. The final section provides a summary andconcluding remarks.2 . Analytical FrameworkFor the purpose of empirical analysis, we formulate behavioural relationships fordomestic saving (DS), private investment (PRIV) and bank lending behaviour (BCP).The saving function is estimated separately for total domestic saving (TS)S privatesaving (PS) and financial saving (FS). Most of the previous studies of saving behaviourin developing countries have focused solely on total saving as the dependent variable.This is unsatisfactory because the positive response of private savings to interest ratesstipulated by the financial repressionist theory may be obscured in the typical develop-ing country context where public sector saving is a significant element in total saving.To shed light on this possible aggregation bias, we chose to estimate functionsseparately for total and private savings. Financial saving is treated separately for tworeasons. First, a comparison of results for the interest rate variable in the financialsavings and total private savings functions would shed light on possible substitutionbetween financial assets and total assets in response to interest rate changes. Second,the financial saving function, together with the function for bank credit to the privatesector, provide a test of the link between deposit interest rates and private investmentthat operates via credit supply. Previous analyses of the impact of interest rate oninvestment have paid little attention to the intermediate link between financial savingand bank lending,4 on the implicit assumption that the former is automatically trans-lated into investment. This is an incomplete approach because, as the post-Keynesiansremind us, the availability of financial saving does not necessarily imply increased banklending. If the banks can create credit without having to rely on initial deposits,an increase in financial saving through higher real deposit rates makes no differenceto the amount of total credit given to the private sector (Davidson, 1986; Thirwall,1976).

Domestic saving (TS, PS or FS) is specified as a function of real interest rate onbank deposits (RID), expected rate of inflation (PE), income (YR), terms of trade(TOT), the share of agriculture in GDP (AGS), population per bank branch ("bankdensity", BDN) and real remittance transfers by expatriate Indians (TRN). Accordingto the Keynesian view on saving behaviour, the income and substitution effects of achange in the real interest rate may work in opposite direction s, rendering th e net effectambiguous. Thus, the interest rate effect on saving is considered "secondary andrelatively unim porta nt, except where unusually large changes are in qu estion" (Keynes,1936, p . 94). Th e financial repression paradigm of M cK inno n (1973) and S haw(1973), however, postulates the impact of a change in real interest rate on saving in thetypical developing economy to be positive. The underlying reasoning is the following.In the typical developing economy where portfolio choices are rather limited, the savingprocess tends to be highly money intensive. Given this peculiarity of saving behaviour,plus the fact that the bulk of saving comes from small savers, the substitutions effectgenerally tends to be much larger than the income effect of an interest rate change. Onthese grounds we expect the regression coefficient of RID to be positive.In most of the existing studies, PE is assumed to impact on saving only through itsrole in the determination of RID. This model formulation assumes inflation neutrality;the absence of money illusion or real balance effect. There are, however, good reasonsfor doubting the validity of this assumption. Quite apart from its influence on realreturns to saving, inflation may influence savings through its impact on real wealth.

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156 P. AthukoralaIf consumers attempt to maintain a target rate of consumption to wealth or of liquidassets to income, saving will rise with anticipated inflation. Also, consumer response tounexpected changes in inflation can result in involuntary saving (Deaton, 1977). On theother hand, in an inflationary situation, a switch from financial assets to real assets canresults in a decline in savings. For these reasons, we include PE as an additionalvariable and assume its sign to be either positive or negative. The third variable, YR, isbased on the Keynesian absolute income hypothesis.T he TOT is included to capture the impact of real income losses or gains broughtabo ut by change in the price of foreign goods in terms of dom estic good s.5 Convention-ally, the sign of its coefficient is assumed to be positive. However, when we assumeforward-looking consumption smoothing behaviour on the part of private agents(households) in the face of volatile and unpredictable income, the sign can go eitherway, depending on whether movements in TOT are perceived to be temporary orperm anen t (Frankel & Razin, 1 992). A terms of trade deterioration th at is perceived tobe temporary may lead to an increase in absorption (that is, an increase in expendituremeasured in terms of domestic goods) as consumers attempt to offset the decrease inpurchasing power of domestic goods so as to keep real expenditure constant. In otherwords, deterioration of the terms of trade induces domestic residents to reduce theirsavings in order to sustain their real standard of living. By the same reasoning,households increase savings when faced with lower future real income as a result of along-run terms of trade deterioration.The remaining three variablesTRN, AGS and BDNare chosen in the light ofthe Indian debate on the de termination of dome stic savings. S ince the mid-1 970s, therehas been a significant increase in inward remittances by expatriate Indians, mostly theIndian workers who migrated to the Gtf countries in response to the oil boom.6 As itis unrealistic to assume that they are current transfers intended solely for consumption,remittances can be considered a positive influence on domestic savings Qoshi & Little,1994).7 Quite apart from this general impact, some studies have treated remittances asa special influence on financial savings (RB I, 1982, Appendix VI ; EPW RF , 1 995). Th etheory behind this relationship has not been stated explicitly, but the underlyingassumption seems to be that, as the bulk of remittances is transferred through bankingchannels, these tend to raise the money holdings of households. However, as Rakshitforcefully argues, "the fact that [remittances] initially raise money balances is of noimportance in explaining why households find it advantageous to hold a large fractionof their savings in the form of currency and bank deposits" (Rakshit, 1983, p. 761).In the debate on the causes of the rapid increase in the saving rate in India in thesecond half of the 1970s (the high saving phase of the Indian economy), one of theunderlying causes considered was the significant decline in the share of agriculture intotal G D P , partly as a result of a significant w orsening of the internal term s of trade andpartly as an outcome of the on-going process of structural transformation in theeconomy (Rakshit, 1983, p . 76; RB I, 1982, p. 46; Joshi & Little, 1994, p. 297). Th isview was based on the hypothesis that agricultural households have a greater marginalpropensity to consume compared to non-agricultural (mostly urban) households; ahypothesis which has not yet been supported by firm statistical evidence. In fact, thepermanent income hypothesis (which postulates a higher marginal propensity to saveout of transitory income) would lead one to expect marginal propensity to save to behigher for agricultural households than for non-agricultural households.A notable development in the Indian financial system following the nationalizationof commercial banks in 1969 has been the rapid expansion of bank branches in thecountry. Po pulation pe r bank bra nch declined persistently from over 90 000 in the


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Interest Rates, Saving and Investment 157mid-195 0s to around 14 000 in the early 1990s (see Table 2 later). T he rapidexpansion of bank branches would have contributed to an increase in financial savingsin the economy, both by improving the accessibility to banking facilities of the generalpublic and by reducing the cost of banking transactions (through reduced transportcost) (Joshi & Little, 1994). Thus, a negative relationship can be assumed betweenpopulation per bank branch (bank density) and financial saving. However, whetherincreased financial intermediation itself significantly raises the overall propensity to savedepends also on the degree of substitution between financial saving and other items inthe house hold asset portfolio. Th us , the expected sign of this relationship in private andtotal saving functions is ambiguous.Common practice in the empirical literature on the determination of savings hasbeen to specify the dependent variable as the ratio of saving to income.8 A majorlimitation of this approach is that it implicitly imposes the "income homogeneityassumption" (that is, "savings are proportional to income, other things equal") oncoefficient estimates. In practice, income and saving may not always move in tandemand hence the arbitrary imposition of income homogeneity assumption may distortcoefficient estimates. The procedure followed in this study is therefore to start with anunres tricted equatio n (which contains the depe nde nt variable in level form an d the levelof real GDP as an additional explanatory variable) and then to impose and test thecoefficient restriction as part of the estimation process.Bank credit to the private sector is specified as a ftinction of FS, BDN and PRIV.The nationalization of banks in India (in 1969) and the subsequent expansion of bankbranches were largely motivated by the objectives of promoting financial savings andmaking bank credit accessible to a wider segment of the economy, particularly those inthe rural areas of the country as well as those engaged in small-scale business. Weinclude BDN in the credit supply function on these grounds and expect its sign to benegative. PRIV is included in the light of the Keynesian proposition that, in a fractionalreserve banking system, investment (demand for credit) can induce credit supplyindependently of changes in financial savings.The explanatory variables in the investment function are bank credit to the privatesector (BCP), the change in income (A YR), public sector fixed investment (PBIV) witha 1-year lag, implicit real rental cost of capital (RRCC) and real interest rate on bankdeposits (RID). BCP constitutes the key link between financial savings and privateinvestment. The inclusion of A YR as an explanatory variable implies an accelerator-type relationship betw een the level of dom estic econo mic activity and capital formation.Previous studies on the determinants of private investment in developing countries havecommonly used real lending rate to capture cost of investment. This practice issusceptible to specification error as the cost of credit is obviously only one element ofthe investor's profitability calculations. Mindful of this limitation, we use an overallindex of RRCC to represent the cost of investment in the investment function. RRCCcaptures the combined effect of changes in the cost of credit (bank lending rate), netof the exp ected rate of change of the price of investment good s, the rate of d epreciationand the current price of investment good, relative to the price of output (general pricelevel, PG).9 It is constructed as,

where NIL is the nom inal interest rate on bank lending, PK is the p rice of capital goodsproxied by the implicit price deflator of fixed investment (1980= 1.00), PG is the


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158 P. Athukoralageneral price level, measured by the GDP deflator (1980 = 1.00), PEK is the expectedrate of change in price of capital goods proxied by the rate of change in PK with a1-year lag and DR is the rate of depreciation.RID is included in the investment function to test McKinnon's "complementarityhypotheses" which postulates that higher deposit interest rates promote investment,independently of their impact on investment via greater credit availability (McKinnon,1973, pp. 60-61). In an economy with a rudimentary capital market, self-financing isa major source of investment funds. Given the fact that investment projects are lumpy,investors must accumulate funds in the form of deposits until the required principalis reached. This process of self-finance within enterprises is impaired by financialrepression because low real yield on deposits increases the cost of accumulating thenecessary money balances in preparation for making future investments. Thus, themore attractive the return on deposits, the more willing investors are to engage in theaccumulation process If this complementarity is at work over and above the cost effectof interest rate changes on investment, then the regression coefficient attached to RIDshould attain statistical significance with a positive sign.The choice of lagged PBIV is guided by the prevalent view in the Indian economicliterature that public investment plays an important complementary role in promotingprivate investment (B ardhan, 1984, Ch apter 4 ). Th is comp lementarity is expected towork on both supply and demand sides. On the supply side, the private sector relieson public investment for most of the infrastructure, because this is either a natural ora legal monopoly of the government. Thus, public investment in infrastructure andprivate investment should be complementary. On the demand side, in theory, therelationship is ambiguous. If there is some slack in the economy one would expect achange in public investment to push private investment in the same direction. Other-wise, some private investment will probably have to be "crowded out". This ambiguitynotwithstanding, given the dominant role played by the government in the provision ofinfrastructure and in key intermediate and investment goods producing industries, it isgenerally assumed that "the stimulation effect of public investment on private invest-ment tends to dominate any possible negative effect through competing for investiblefunds" (Bardhan, 1984, p. 25).Apart from the explanatory variables discussed above, an intercept dummy vari-ables ("liberalization dummy", LBD) is included in all equations, in order to testwhether the market-oriented policy reforms initiated in 1991 per se has had a specificinfluence (over and above their impact operating through the other variables explicitlyallowed for in regression specifications) on saving/investment behaviour in the Indianeconomy. There has been a slowing down of the annual increase in domestic saving/investment (and therefore a decline in the saving and investment rates) in India inthe aftermath of the reforms. Two schools of thoughts have emerged to explain thisphen om enon (EP W RF , 1995; Athukorala & S en, 1995). T he "official" view10 takesthe position that the decline is simply a reflection of lags involved in the adjustmentof the economy to the new policy environment, and the improved incentives forprivate sector performance should eventually lead to higher saving/investment levels.The critics of reforms, however, interpret the phenomenon as an invariable adverseoutcome of the reform process. According to this interpretation the reform process,by encouraging conspicuous consumption among the rich and middle classes, haspaved the way for a long-term deterioration in saving/investment performance in theeconomy.11For ease of reference, the complete model with the variable definitions is listed inTable 1.

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Interest Rates, Saving and Investment 159Table 1 . The model

DS',=f(YR RID PETOT TRN AGS BDN LBD)DS ,= TS DS FS,BCP, =f(DFS BDN PRIV,, LBD)PRIV,=f(BCP GYR RRCC ,, PBIV,-, RID,, LBD)Dependent variablesDS dome stic saving8TS total domestic saving3PS private saving8FS financial saving11BCP bank credit to the private sector3PRIV private fixed investment3Explanatory variables^YR ( + ) domestic income (proxied by G DP )3RID ( + ) NID - PE, real interest rate on bank depositsNID nominal interest rate on bank depositsPE ( + or ) expected rate of inflation proxied by the actual rate of inflation with a 1 -year lagTOT{+ or - ) = [PX/PM]* 100, terms of tradePX price index of exports (1980 = 1.00)PM price index of impo rts (1980 = 1.00)AGS ( + or - ) share of agriculture in total G D PTRN ( + ) remittances by Indian expatriates3BDN ( ) population per bank branch ("bank density")GYR{ + ) growth rate YR 3RRCC ( ) real rental cost of capitalNIL nominal interest rate on bank lendingPK price of capital goods proxied by the implicit price deflator of fixed investm ent (198 0 = 1.00)PG general price level, measu red by the G D P deflator (1980 = 1.00)PEK expected rate of change in price of capital goods proxied by the rate of change in PK with a1-year lagDR rate of depreciationPBIV ( + or ) public sector fixed investment3LB D (+ o r ) a dumm y variable (which takes value one for years 1991-9 5 and zero for other years) torepresent the post-trade liberalization period"Expressed in real terms.The sign expected for the regression coefficient is given in parentheses.

3 . Data and Econometric ProcedureThe model formulated in the previous section is estimated over the sample period1955-95 using annual data.12 All variables, except interest rates and inflation (which aremeasured in proportional form), are measured in natural logarithms. Data sources arelisted and methods of data transformation adopted and key limitations of the data arediscussed in the Appendix. The data series are summarized in Table 2, in order to aidthe interpretation of the results.In line with standard practice in modern time-series econometrics, we begin theestimation process by testing the time-series properties of the data. Two tests for unitroots are used: the augm ented D icky-Fuller (ADF) test and the K wiatkowiski-Phillips-Schmidt -Sh in (KPSS) test. Th e latter tests the null of a unit root against the alternativeof stationarity, w hile the form er tests the nu ll of stationarity against the alternative of aunit root. T h e choice of the KPSS test to supplem ent th e widely used ADF test is basedon evidence that tests designed on the basis of the null that a series is 7(1) have low


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ONOTable 2 . Summary data on variables used in econometric analysis 3

D e p e n d e n t v a r i a b l e sTS

PS

FS

BC P

PRIV

E x p l a n a t o r y a r i a b l e sYRRI DRI LBDN (x 000)TO TTR N

PBIV

AG SPRCC

1956-60

7185(12.35)6136

(10.55)1703(2.93)648

(1.11)5534(9.52)

581560.702.88

89.013525 0

(0.43)3854(6.63)49.43.46

1961-65

9583(13.23)7305

(10.09)2140(2.96)1012(1.40)6363(8.79)

72412-0.361.38

81.8144146

(0.20)6070(8.38)45.84.26

1966-70

12300(14.62)10178(12.05)2320(2.76)1422(1.69)9227

(10.96)

84150-2.30-0.9065.1145242

(0.29)6661(7.92)47.13.46

1971-75

17294(17.11)14203

(14.05)4007(3.96)2419(2.39)10518

(10.40)

101093-1.081.92

37.7154206

(0.20)7342(7.26)45.34.68

1976-80

26577(21.40)20969

(16.88)7100(5.72)4160(3.35)12565

(10.12)

1241901.468.56

23.6116977

(0.79)10224(8.23)40.08.14

1981-85

29414(19.40)23847(15.73)10307(6.80)5627(3.71)14593(9.63)

151589-2.765.8416.8

1192059(1.36)14135(9.32)37.98.94

1986-90

41275(20.94)36668(18.60)15111(7.67)6899(3.50)20890(10.60)

1970860.988.6814.1

1171957

(0.99)19156(9.72)31.913.48

1991-95

58059(23.01)54623(21.65)25945(10.28)

8370(3.32)33081(13.11)

2522750.527.1214.3

1093313(1.31)21903(8.68)31.110.50

1955-95

25211(19.38)21741(16.71)8579(6.59)3820(2.94)14096(10.83)

130119-0.283.94

42.8129.8

1144(0.88)11168(8.58)41.17.12

S352*.

"Value series are in crores of Indian rupees at constant (1981) prices (crore = 10 million). Figures in parentheses ar e percentage of G D P .bAnnual averages.Source an d methods: Data sources an d methods of data compilation are explained in the Appendix.Downlo

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Interest Rates, Saving and Investment 161power in rejecting the null. Reversing the null and alternative hypotheses is helpful inovercoming this problem (Kwiatkowiski et al, 1992).According to the test results the key variables of interest do not have the same orderof integration; in particular, the three saving series and the interest rate series arestationary.13 The two alternative tests yield remarkably similar results. Thus, now-fashionable econometric procedures that are appropriate for 7(1) variables are notapplicable in our case. However, given the presence of non-stationary variables, it isnecessary to guard against the possibility of estimating spurious relationships. Thetime-series econometrician's prescription in this type of situation is to difference thenon-stationary variables (to achieve stationarity) and use them in that transformed formtogether with the other (stationary) variables. This procedure, while statistically accept-able, has the disadvantage of ignoring long-run relations. W e therefore opted to use theunrestricted error correction modelling (JJECM) procedure of Hendry, which mini-mizes the possibility of estimating spurious relations while retaining long-run infor-mation (Hendry, 1995). This approach also has the added advantage that it providesfor estimating lag effects without arbitrarily constraining the lag structure at the outset.Under the UECM method, the long-run relationship being investigated is embeddedwithin a sufficiently complex dynamic specification, including lagged dependent andindependent variables, in order to minimize the possibility of estimating spuriousrelationships. The estimation procedure starts with an autoregressive distributed lag(ADL) specification of an appropriate lag order:

m mY, = a + 2 AiY t- X BiX,-i + p, (1)1=1 i"=0where a is a vector of constants, Y t is an (XI) vector of endogenous variables, X, isa (k X 1) vector of explanatory variables, A and B are (n X ri ) and ( X k) matrices ofparameters, and x tis a stochastic error term.Equation (1) can be reparameterized in terms of differences and lagged levels so asto separate the short-run and long-run multipliers of the system.

m 1 m iAY,= a+ X AY,-i+ BAX,-i+C0Yt-m + C1X,-m + it (2)1=1 =0where

C , = Im \= S B ,V=o '

and where the long-run multipliers of the system are given by Co lC\.Equation (2) constitutes the "maintained hypothesis" of our specification search.This general model is "tested down" (using ordinary least squares, OLS) by droppingstatistically insignificant lag terms and imposing data-acceptable restrictions on theregression parameters. The testing procedure continues until a parsimonious errorcorrection representation is obtained which retains the a priori theoretical model as itslong-run solution. To be acceptable, the final equation must satisfy various diagnosticchecking proced ures. In applying the UEC M proc edure, we set the initial lag length onall variables in the general ADL equation at two periods. Th is is the established practicein modelling with annual data. After undertaking a preliminary specification searchusing OLS, the equation for PRIV (which contains a contemporaneous terms of ajointly determined variable (BCP) among explanatory variables) was estimated usingtwo-stage least squares (2SLS). The remaining equations were estimated using OLS.


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162 P. AthukoralaPE and AGS variables in the three saving equations and TOT in the financial savingequation had statistically insignificant coefficients in all experimental runs and weretherefore d ropped in the final equations. BDN was omitted from the equation for bankcredit as it turned out to be highly correlated with DFN. In all cases the specificationchoice was statistically acceptable in terms of the joint variable deletion tes ts against th e

maintained hypothesis. As intercept du m m y variable, DIP, was included in the equationfor total domestic saving to allow for an "unexplained" dip in the saving rate between1982 and 1987.14 Likewise, an intercept dummy variable, >58, was added to theequ ation for bank credit equ ation t o allow for an outline observ ation for 1 958 in the BCPseries. In both cases, without these intercept dummies the equation does not pass thetests for normality (JBN) and heteroscedasticity (ARCH). However, the coefficientestimates for the oth er variables are remarkab ly resilient to the addition of these variables.4. ResultsThe final parsimonious estimated equations, together with a set of commonly useddiagnostic statistics, are reported in Table 3. Long-run elasticities relating to the keyexplanatory variables computed from the long-run (steady-state) solutions to theestimated equations are given in Table 4. All equations are statistically significant atthe 1% level (in terms of the standard F-test) and they perform well by all diagnostictests. In terms of the Chow test for parameter stability conducted by splitting the totalsample period into 19 56-79 and 1980 -95, 15 ther e is no evidence of para m eter instability.On reestimating for the sub-period 1956-90, all equations pass the Chow test ofout-of-sample forecasting ability (Chow's prediction failure test) for the post-reformperiod (1991-95). Apart from these tests, a residual correlogram of up to 6 years wasestimated for each equation, with no evidence of significant serial correlation. Theequations also comfortably passed the C U S U M test on the recursive residuals and theC U S U M S Q t e s t .T o com m ent first on the three saving functions,16 the long-run income hom ogeneityassumption (unit coefficient restriction on the real income variable) is statisticallyacceptable in all three equations. Thus, other things remaining equal, a 1% change inreal GDP is associated with a similar change in savings, resulting in a constant ratio ofeach component of saving to GDP in the long run. The results for the RID variablesuggest that saving behaviour in India depends significantly on the real rate of return onbank deposits. A 1% increase in RID is associated with a 1.50% increase in real financialsaving, 1.72% increase in real total private saving and 1.1% poin t increase in total privatesaving. Considering the fact that during the study period financial savings on averageaccounted for over 45% of total private savings (Table 2), there is no evidence ofsubstitution of other assets for financial assets in response to changes in the real depositrate. Overall, the Indian experience provides ample support for the McKinnon-Shawproposition that, in an economy where the saving behaviour is highly intensive infinancial assets, high real interest rates promote both financial and total saving in theeconomy. The interest elasticity of total private saving (1.71) is greater in magnitudecompared to that of total domestic saving (1.11) and the difference is statisticallysignificant at the 1% level. These results support our postulate that use of total saving(which include public sector saving) as the measure of saving performance, as has beendon e in mo st of previous stud ies on this subject, is likely to infuse a down wa rd bias intoestimates of interest elasticity of saving.

Among the remaining variables, BDN stands out as a highly significant variable inexplaining both total and financial savings. Thus, there is strong support for the view

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Interest Rates, Saving and Investment 163Table 3. Determinants of saving and investment in India: regression results

1. Total domestic saving (TS)ATS,= - 1.04+1.31AGDP 1 + 0.69ARID, + 0.lOATRN, + 0.77RID,-,-OAOBDN,-1-0.23BDN,-1(2.13)** (3.28)*** (2.33) (2.56)*** (2.02)** (1.56)* (4.06)***

0.12TOT,-,- 0 . 6 9 ( 7 - 5 - GDP),-,- 0.12D/P(1.10) (4.14)*** (2.91)***i? 2 = 0 .65 F (9,28) = 5 .70 S = 0 . 07 L /H -F ( l ,2 9 ) = 0 .07L M l - F ( l , 2 7 ) = 0 .9 3 LM 2 - F ( 2 , 2 6 ) = 1.00 RESET- F(l,27) = 0.32 JBN-f(2) = 0.83ARCm - F(l,27) = 0.27 ARCH2 - F(2,26) = 1.16 CHOW- F(9,20) = 1.33 PRF- F(5,23) = 0.832. Total private saving (PS)ADS,= - l.62+l.29AGDP , + 0.90ARID t + 0.l3ATRN ,+ 1.07RIDt-,-0.20BDN,-,(2.57)*** (3.33)*** (2.49)*** (2.62)*** (2.20)** (3.58)***

0 . 2 2 7 O T , - , - 0 . 6 2 ( D S - GDP),-,- 0.10DLB(1.62)* (3.78)*** (1.87)**^ = 0.57 F(8,29) = 4.69 SE= 0.08 LIH- F(l,28) = 1.80LAf l ( l ) - F ( l , 2 8 ) = 0 . 0 3 LM 2 -F ( 2 ,2 7) = 0 .07 RESET-F(l,28) = 0.20 JBN- *2(2) = 0.85/AC/1 - F(l ,28) = 0.03 ARCH2 - F(2,27) = 0.06 CHOW- F(7,24) = 2.09 PRF - F(5,26) = 1.42

3 . Financial saving (FS)J K S , = - 0.95 + 1.20iGDP,+ 1.31/D,- 0A9BDN,-, - 0 . 8 7 ( F 5 - GDP ),-, + 0.15F5,-2 + 0.26DL(3.96)*** (1.34)* (2.18)** (5.16)*** (5.59)*** (1.47)* (2.55)***i? 2 = 0.55 F(6 ,28) = 6 .64 S = 0 .1 6 UH- F(l,32) = 0.62LMl -F(l,32) = 0.02 L A 2- F( 2,3 1) = 0.27 RESET-F(.l,?>2) = 4.81 JBN~x2(2) = 0.130 .35 ^ i ? C H 2-F (2 , 31 ) = 1.23 CHOW- F(14,20) = 0 .73 PR F- F( 5, 29 ) = 1 .634. Bank credit to the private sector (BCJ?)ABCR, = - 2.77 + 0.37AFS,+ 0A7FS,-1 + 0.59PRIV,-, + 0 .84B CR, -1 - 0 .46 DL B - 2 .11D58(2.04)** (2.06)** (3.40)*** (2.32)*** (7.75)*** (2.51)*** (8.36)***R2 = 0.84 F(6,33) = 30.47 SE = 0.22 M l - F ( l , 3 2 ) = 2 .0 5 LA 2 - F ( 2 , 3 1 ) = 1.23 RESET-F(.\,32) = 1.29 JBN-y?(2) = 0.03y 4 i ? C Hl - F ( l , 3 2 ) = 0 .0 3 ^ i ? CH 2 - F ( 2 , 3 1 ) = 0 .5 3 CHOW-F(5,30) = 1.79 F i F -F (5 ,3 0) = 1.245. Private investmentJ P R / F , = 0.73 + 0.07BCP, + 0.81A 2YR, + 0.89AYR,-t + 0.16PBIV,-2(1.58)* (2.15)** (2.38)** (1.56)* (2.22)**

- 0A5RRC C-1 + 0.56RID,-1 - 0.28PRIV,- + 0.23APRIV,-2 + 0.10DLB(1.37)* (3.03)*** (3.35)*** (1.95)** (1.74)**/? 2 = 0.62 F(9 ,29) = 5 .30 5B =0 .0 6 SPEC-^(3) = 3.35L M 1 ( 1 ) - Z 2 (1) = 1.24 L M 2- x 2 (2) = 1.52 RESET- / ( I ) = 0 . 7 6 JBN - z 2 (2) = 1.05A R C m ( l ) - F ( l , 2 8 ) = 0.42 A RCH2 - F(2,27) = 0.75 CH OH 7- F(14,21) = 1.43 P F- F( 5, 25 ) = 1.32Notes: t-ratios of regression coefficients are given in parentheses. Approximate critical values for the f-ratios areas follows: 10 % = 1.30 (*), 5% = 1.69 (**) and 1% = 2.42 (***).Test statistics: LIH = F-test for the long-run incom e homogen eity restriction; SPEC = S argan's test for the correctspecification of instrum ents; LM = Lagrange mu ltiplier test of residual serial correlation; RESET = Ramsey testfor functional form mis-specification; JB N = Jarque-B era test for the normality of residuals; ARCH= Engle'sautoregressive conditional heteroscedasticity test; CHOW= Chow test for parameter stability conducted bysplitting the sample period into 1956-79 and 1980-95; PRF= Chow's test for prediction failure (theout-of-sample forecasting ability) conducted to test the ability of the equation reestimated for the pre-reformperiod (1956-90) to forecast the dependent variable for the post-reform years (1991-95).


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Table 4 . Estima tes of long-run elasticities SQ2Explanatory variableRI D1.11(1.81)**1.71(1.92)**1.50(2.02)**1.94(2.17)**

GD P1.00(4.14)***1.00(3.78)***1.00(5.58)***

BD N- 0 . 3 3(9.58)***- 0 . 3 2(7.15)***- 0 . 5 6(12.11)***

TO T0.17(1.01)0.35(1.46)*

FS

0.56(3.71)***

BC P

0.23(2.32)**

PRIV

0.69(2.61)***

PBIV

0.55(2.59)***

RRCC

- 1 . 5 5(1.32)*

1. Total domestic saving (TS)2. Total p rivate saving (PS)3. Financial saving (FS)4. Bank credit to the private sector (BCP)5. Private investment (PRIV)

Note: f-ratios of regression coefficients are given in parentheses. Approximate critical values for the f-ratios are as follows: 10%= 1.30 (*), 5%= 1.68 (**) and1% = 2.42 (***). Not applicable.Source: Computed from the long-run (steady-state) solutions to the estimated models reported in Table 3.Downlo

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Interest Rates, Saving and Investment 165that the expansion of banking facilities since the 1970s seems to have contributedsignificantly to im provem ent in saving propen sity in the econom y. Th ere is no evidenceto suggest that the expansion of bank branc hes gen erated g reater financial savings in theform of bank deposits at the expense of the accumulation of other assets. Thecoefficient of T O 7* is positive a nd statistically significant in bo th total and private savingfunctions, suggesting that real income gains (losses) associated with terms of tradeimprovements lead to higher (lower) saving. Thus, Indian households seem to considerterms of trade movem ents as temporary shocks. Remittances (TRN) seem to impact onyear-to-year fluctuations in total and private saving, but do not seem to yield astatistically significant long-run impact. As noted, AGS failed to attain statisticalsignificance in all saving functions. T hu s, the re is no empirical sup port for the view tha tthe decline in the share of agriculture in domestic production (and the accompaniedincreased urbanization of the economy) has contributed to the growth in domesticsaving.Among the developing countries of similar income status, India has shown aremarkably high propensity to save, particularly since the early 1970s (Joshi & Little,1994). Our results clearly point to the progressively important role financial intermedi-ation has played in the continued buoyancy in domestic saving. Throughout the periodunde r study up to 1991 , the nom inal interest rate was an adm inistered price, changedat infrequent intervals. However, there were no persistent adverse movements in realdeposit rates; macroeconomic policy had an anti-inflationary stance, and high inflationand sharply negative real deposit rates w ere not allowed to persist for long. At the sametime, the spread of banking facilities played a useful supplementary role in increasingfinancial (and hence total) saving.

The equation for bank credit to the private sector suggests that a 1% increase infinancial saving is translated into a 0.56% increase in bank credit to the private sector.PRIV is also highly significant, suggesting that investment demand is a significantdeterm inant of bank lend ing. In the investmen t function, th e coefficient of the real bankcredit variable (BCP) is significant and suggests that a 1% point change in the volumeof real bank credit is reflected in a 0 .23 % change in the level of real private investm ent.This result, taken together with the estimates of interest elasticity of financial saving(1.5) and elasticity of bank credit with respect to financial saving (0.56), suggest thata 1% increase in real deposit rate brings about a 0.20% increase in private investmentthrou gh t he supply of real ban k cred it. T he re is som e statistical evidence (significant atthe 10% level) of a negative effect of the bank lending rate (as part of total cost ofinvestm ent) on the level of inve stm ent; a 1% increase in the former results in a 1.50%decline in the latter. At the same time, there is strong statistical support forMcKinnon's complementarity hypothesis that high real deposit rates promote privatesector capital formation by facilitating the accumulation of finance necessary to makethe investment. After allowing for other influences, a 1% increase in the real depositrate is associated with over 2 % increase in private investmen t.17 Overall, the cumu lativenet impact of the real deposit rate on investment operating through financial intermedi-ation and through the direct complementarity effect outweighs the cost effects ofinterest rate changes on investment.There is also evidence of a significant positive effect of lagged government invest-ment on private investment. This result supports the view that in the Indian economypublic investment plays an important complementary role in promoting private invest-ment. The standard accelerator mechanism also appears important in explainingprivate investment. Finally, the results for the LBD do not lend support to thepessimistic view that the m arket-oriented policies initiated in 1991 have had an adverse

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166 P. Athukoralaeffect on saving/investment performance in the Indian economy. On the contrary, thecoefficient of LBD is statistically significant with the positive sign in private saving,financial saving and private investment functions. This result suggests that, onceallowance is made for the other influences, liberalization has in fact enhanced saving/investment performance in the economy. In the bank credit equation the coefficient ofLBD is negative and statistically significant. This result seems to reflect stringentrestrictions on bank lending as part of the stabilization policy in the aftermath of thereforms.

5. ConclusionIn this paper, we have examined the interest rate-saving-investment interrelations inthe Indian economy using annual d ata for the period 19 55-9 5. The results are generallyin support of the propositions of the M cK inno n-S haw financial repression paradigm. Inthe Indian context higher real interest rates seem to promote both financial and totalsaving. There is no evidence of a significant asset substitution effect resulting fromchanges in interest rate as postulated by the neo-structuralist development economists.In line with Shaw's debt-intermediation view, financial saving in turn promotes invest-ment through the provision of bank lending to the private sector. There is also strongstatistical support for McKinnon's complementarity hypothesis that higher depositrates promote investment through facilitating self-financed capital accumulation. Thehigher level of real interest rates does seem to have a negative impact on the level ofinvestment, but the cumulative net effect of interest rates on investment is unambigu-ously positive.It is of course not possible to generalize from a single country case. However, ourresults do cast doubt on the robustness of results coming from the vast cross-countryliterature on this subject. Thus, further systematic empirical investigations covering avariety of individual country situations are needed before valid generalizations on thepolicy relevance of the financial repression paradigm can be made.

Notes1. Fo r recent surveys of this literature, see Fry (1 995 ), Gibson & Tsakalotos (1994) and D eaton(1989).2. T h e available coun try studies include, Athukorala & Jayasuriya (199 4), Onis & Riedel(1993), Collins (1994), Warman & Thirlwall (1994) and de Melo & Tybout (1986).3. The most recent Indian studies in this subject area are Krishnamurty et al. (1987), Ketkar& Ketkar (1992) and Laum as (1990 ). Thes e are both dated and partial in subject coverage.Krishnamurty et al. (1987) examine the determinants of the saving rate using data from1954/55 to 1981/82. Lau ma 's (1990) test of M cK inno n's com plementarity hypothesis relatesto the period 1954/55 -1974/75 . Ketkar & Ketkar (1992) examine the impact of bank branchexpansion on financial saving, private investment and GDP growth covering the period1952/53-1984/85.4. One notable exception to this general practice is the study by Warman & Thirlwall (1994),who explicitly test the link between financial saving and investment through the estimationof a bivariate credit supply function.5. In theory, terms of trade changes are already a part of real GDP. However, in practice theprice deflators used in national accounting generally allow only for changes in the generallevel of prices and fail to capture price structural effects on the level and growth of realincome such as those due to changes in the terms of trade. Thus, changes in TOT can beexpected to have an additional effect on that of changes in GDP on savings (Ady, 1976,p . 110).

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Interest Rates, Saving and Investment 1676. Accord ing to balance of paym ents recor ds, private transfers (w hich pred om inan tly consist ofremittances) increased from 0.2% of GDP in the early 1970s to 1.5% of GDP in the early1990s (Table 2).7. In this connection, it is impo rtant to note that, according to the me thod of estimation ofhousehold saving in India, the contribution of migrant transfers (or any other form of incometransfers from the rest of the world such as interest and dividend transfers) is captured inreported saving estimates (Little & Joshi, 1994).8. This specification is usually adopted purely for practical reasons: to conserve degrees offreedom, to minimize the problem of collinearity among the explanatory variables and tocircumvent the problem of lack of appropriate price deflators.9. I am grateful to G eorge Fan e for help in developing this measure of RRCC. On the theorybehind this measure, see Jorgenson (1967).10 . This view has been expressed in annual issues of The Economic Survey, Ministry of Finance,Delhi.11 . For details on this debate see Athukorala & Sen (1995).12 . All data series are on the basis of the Indian fiscal year, 1 April in the previous year to 31March of the given (stated) year.13. The test results are available on request from the author.14 . In these 6 years the rate of growth of domestic savings lagged behind that of GDP, resultingin a decline in the saving rate below the historical trend. The underlying causes of this dipin saving rate are still debated. See Roy & Sen (1991) and the work cited therein.15 . This time division is chosen to reflect the gradual pro-market shift in India's economic policyfrom about the early 1980s. We also conducted the Chow test for parameter stability byequal splitting of the data to yield the same test results.16 . Note that the estimated financial saving function (Table 3) has relatively low overallexplanatory power (in terms of lower R2 and higher SE) compared to the total and privatesaving functions. Perhaps this somewhat surprising result has something to do with themethod adopted by the Central Statistical Organisation (CSO) in compiling the financialsaving series. The CSO estimates show only the net increase in financial assets (as-

setsliabilities). Given the rapid growth of borrowing by the household sector from banksand other institutional sources since the early 1970s and the resulting additions to itsfinancial liabilities, it is likely that this series does not fully reflect the rate of growth offinancial saving in the economy (RBI, 1982, p. 35).17 . Contrary to what one would have anticipated a priori, the intercorrelation be tween RID andRRCC does not seem to have disturbed the robustness of this result. In the formulationadopted, the partial correlation coefficient of RID (0.49) is larger in magnitude than simplecorrelation between RID and RRCC (0.32).ReferencesAdy, P. (1976) Growth models for developing countries, in: A. Cairncross & M. Puri (Eds)Employment, Income Distribution and Developm ent Strategy: Problems of the Developing Countries(London, Macmillan), pp. 106-119.Athukorala, P. & Jayasuriya, S. (1994) Macroeconomic Policies, Crises, and Grow th in Sri Lanka,1969-90 (Washington, DC, World Bank).Athukorala, P. & Sen K. (1995) Economic reforms and the rate of saving in India: someobservations, Economic and PoliticalWeekly, 30, pp. 2184-2190.Bardhan, P. (1994) The Political Economy of Development in India (Oxford, B asil B lackwell).Collins, S.M. (1994) Saving, investment and external balance in South Korea, in: S. Haggard,R.N. Cooper, S.M. Collins, C. Kim & S. Ro (Eds) Macroeconomic Policies and Adjustment inKorea, 1970-1990 (Cambridge, MA, MIT Press) .Davidson, P. (1986) Finance, funding, saving, and investment, Journal of Post-Keynesian Eco-nomics, 9, pp. 101-110.de Melo, J. & Tybout J. (1986) The effect of financial liberalisation on savings and investment inUruguay, Economic Development and Cultural Change, 34, pp. 561-587.De aton, A . (1977) Involuntary saving through unanticipated inflation, American Economic Review,67, pp. 899-910.Deaton, A. (1989) Saving in developing countries: theory and review, Proceedings of the WorldBank Annual Conference on Development Economics, 1989 (Washington, DC, World Bank),pp . 61-108 .

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168 P. AthukoralaDombusch, R. & Reynoso, A. (1989) Financial factors in economic development, AmericanEconom ic Review P apers and Proceedings, May, pp. 204-209.EPWRF (Economic and Political Weekly Research Foundation) (1995) Economic reforms andrate of saving, Econom ic and Political Weekly, 30, pp . 1021-1041.Frank el, J.A. & Razin A. (1992) Fiscal Policies and the World Economy, 2nd edn (Cambridge, MA,MIT Press) .Fry, M. (1995) Financial development in Asia: some analytical issues, Asian-Pacific EconomicLiterature, 9, pp . 40-5 7.Gibson, H.D. & Tsakalotos, E. (1994) The scope and limits of financial liberalisation indeveloping countries: a critical survey, Journal of Development Studies, 30, pp. 578-628.Hendry, D.F. (1995) Dynamic Econometrics (Oxford, Oxford University Press).Jorgenson, D.W. (1967) The theory of investment behavior, in: R. Ferber (Ed.) Determinants ofInvestment Behavior, Universities-National B ureau Conference S eries No . 18 (New York,Columbia University Press), pp. 129-155.Joshi, V.H. & Little, I.M.D. (1994) India: Macroeconomics and Political Econom y, 1964-1991(Washington, DC, World Bank).Ketkar, K.W. & Ketkar, S.L. (1992) Banking nationalisation, financial savings, and economic

development: a case study of India, Journal of Developing Areas, 27, pp. 69-84.Keynes, J.M. (1936) The General Theory of Employment, Interest and Money (New York, Harco urtBrace) .Krishnamurty, K., Krishnaswamy, K.S. & Sharma, P.D. (1987) Determinants of saving rates inIndia, Journal of Q uantitative Econom ics, 3, pp. 335-357,Kwiatkowiski, D., Phillips, P.C .B ., S chmidt, P . & S hin, Y. (1992) Testing the n ull hypothesis ofstationarity against the alternative of a unit root: how sure are we that economic time serieshave a unit root?, Journal of Econometrics, 54, pp. 159-179 .Lau ma s, P.S . (1990) M onetization, financial liberalization, and economic developm ent, EconomicDevelopm ent and C ultural Change, 38, pp . 377-390.McKinnon, R.I. (1973) Money and C apital in Economic Development (Washington, D C , B rookingInstitute).Onis, Z. & Riedel, J. (1993) EconomicCrises and Long-term Growth in Turkey (Washington, DC,World Bank).Rakshit, M. (1983) On assessment and interpretation of saving-investment estimates in India,Econom ic and Political Weekly, Annual Number, pp. 753-776.Roy, T. & Sen K. (1991) Changes in saving rate and its implications for growth, Economic andPolitical Weekly, 20, pp. 1055-1058.RBI (Reserve Bank of India) (1982) Capital Formation and saving in India: 1950-51 to 1979-80,Repo rt of the Wo rking Group on S aving, (Delhi, RB I).Shaw, E. (1973) Financial Deepening in Economic Development (Oxford, Oxford University Press).Srinivasan, T.N. (1994) Data base for development analysis: an overview, Journal of DevelopmentEconomics, 44, pp. 3-27.Taylor, L. (1983) Structuralist Macroeconomics: Applicable Models for the Third World (New York,Basic Books).Thirlwall, A.P. (1976) Financing Economic Development(London, Macmillan).W arm an, F . & Thirlwall, A.P. (199 4) Interest rates, saving, investment and growth in Mexico1960-90: tests of the financial liberalisation hypothesis, Journal of DevelopmentStudies, 30,p p . 6 2 9 - 6 4 9 .

AppendixThe data series used in this study were obtained directly or compiled from following publications:C S O , National AccountsStatistics, Delhi (various issues); Gov ernmen t of Indian (199 5), EconomicSurvey 1994-95, Delhi: Ministry of Finance; and Reserve Bank of Indian, Monthly Bulletin andReport on C urrency and Banking, Delhi (various issues).In the selection and transformation of most of the data series, we have simply followedestablished practice in this field of research. However, the choice of data series on private andfinancial savings, price deflators and interest rates, and the measurement of expected inflationneed some explanation.CSO data permit disaggregation of private savings into corporate savings and householdsavings. However, the line of demarcation between households and firms in these estimates is

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Interest Rates, Saving and Investment 169rather arbitrary. The "household sector", as denned by the CSO, is highly heterogeneous and isvirtually synonymous with the non-corporate sector of the economy. Apart from the "householdproper", it includes partnerships and unincorporated businesses, and there is evidence that therelative importance of the latter two has increased over time (RBI, 1982). Thus, the availablehousehold saving series is not a good measure of household saving behaviour as suggested bytheory. Moreover, as "household saving" is measured by deducting government saving andcorporate saving from total gross domestic savings, measurement errors in the latter variables areinvariably absorbed into the former (Srinivasan, 1994). For these reasons, we do not treathousehold saving as a separate category in this study.The deposit and lending rates used are the 1-year deposit rate (minimum) and the 1-yearlending rate of the S tate B ank of India. Ideally, the lending and deposit rate series should havebeen constructed as weighted averages of rates relating to deposits/loans of different termstructures using relative shares of respective deposits/loans. Unfortunately, information on thematurity structure of deposits is not readily available. There is, however, evidence that, as mostof the key series move in tandem, the choice of a particular series over the preferred weightedaverage does not make significant difference in empirical analysis (Laumas, 1990).The data series on saving (TS, PS, FS) and remittances (TKN) were deflated by the GDPdeflator (1981 = 1.00). The series on bank credit (BCF) and private and public fixed investment(PRIVand PBIV) were deflated by the implicit deflator for fixed capital formation (1981 = 100).Following the majority of studies in this subject area, the expected rate of inflation is proxied bythe rate of inflation (measured using the GDP delator) with a 1-year lag. Likewise, the expectedrate of change in capital goods price is measured as the rate of change of capital goods price(measured by the implicit deflator for gross domestic fixed capital formation) with a 1-year lag.The static inflationary expectations hypothesis that undergirds this variable choice is consideredappropriate for a low-inflation country like India, especially when working with annual data.For useful discussions on the nature and limitations of Indian data on saving and investment,see RB I (1 982), Rakshit (1983) and S rinivasan (1994). It is generally believed tha t, on the w hole,these data have a much firmer foundation (both in terms of coverage and intertemporal consist-ency) than those for any other country at the same stage of development.

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9-interest rates, saving and

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