Knowledge that Transforms

In a recent article of this REVIEW, Sato and Hoffman (1968) derive an explicit form of production function with variable elasticity of factor substitution (hereafter VES production function) and give some estimates of VES functions for United States and Japan. The authors contend that the overall impression (of regression results.) is that VES functions are more realistic than CES (C-D) (Sato and Hoffman: p. 457). The purpose of this note is to show that their results both for United States and Japan, which they report to be most useful ones among other estimates under alternative forms of VES functions, are not valid, primarily because their estimates are not logically consistent with their a priori estimates used for estimating VES functions.

In a recent note in this Review,' Jene K. Kwon and Richard Thornton argue that their empirical research . . supports the implication of our model; to wit, that Federal Home Loan Bank (FHLB) bonds are a substitute for Savings and Loan (S and L) (p. 98). It is the contention of this note that no such distinctive implication can be drawn. Rather, what Kwon and Thornton have shown is that credit market instruments are substitutes for S and L deposits and that FHLB bonds are viewed by the market as one category of credit market instruments. This is not a new finding and is not a sufficient basis on which their policy recommendations can stand. Kwon and Thornton use two measures of savings flows to FSLIC-insured S and L's, net new savings (NNS) and changes in gross savings (AS). They conclude that NNS performs better on the basis of the R2 S2 (SE ?), Durbin-Watson and t statistics (p. 98). We will therefore confine our comments to this form of their model. Kwon and Thornton use a linear version of a stock adjustment model to compare various interest rate S and L deposit rate differentials in a model containing disposable income, lagged NNS and seasonal dummy variables. In all, six equations are estimated (p. 98). Their results indicate that when changes in the rate of return on FHLB bonds minus the rate on S and L deposits is estimated (equation 1.1, p. 98) the regression coefficient is significant at the 1 per cent level using the t test. However, it is also the case that when the commercial paper rate minus the rate on S and L deposits is estimated in the next equation, the regression coefficient is also significant at the 1 per cent level (equation 1.2, p. 98). Furthermore, the three-month bill-S and L differential (equation 1.3, p. 98) and the nineto twelve-month bill-S and L differential (equation 1.4, p. 98) are statistically significant at the 5 per cent level. Aaa corporate S and L rate differential (equation 1.5, p. 98) and the time deposit S and L rate differential (equation 1.6, p. 98) are the only interest rate differentials which are not statistically significant. Examining those equations in which the interest rate differentials are statistically significant, the implications of the analysis seem clear. There is no distinctive implication that FHLB bonds are substitutes for S and L deposits, rather the implication is that open market instruments are substitutes for S and L deposits and that as the rate spread between instruments and S and L deposits increase, the net new savings to S and L's decrease. This is certainly not a new finding and while interesting, this hardly shows that FHLB bonds are materially different from any other open market instrument in their effect on S and L deposits. This is to be expected as it is to be doubted that FHLB bonds are somehow special when viewed by the market along with other market instruments. This note has demonstrated that the implications drawn by Kwon and Thornton from their work are too strong. important unanswered questions that need to be addressed are: (1) how important an effect does the competition of FHLB bonds have on S and L deposit flows?, and the related question, (2) given that FHLB bonds do compete in the market with deposits at S and L's, what is the actual net impact on S and L's of an increase in FHLB bond flotations for the purpose of making advances to S and L's? 2 That the Kwon-Thornton work does not allow us to draw any conclusion about either of these questions is clear from regressing NNS on NNS lagged one period (one of the independent variables in their regression equations) and the seasonal dummy variables. In their equation 1.1 which contains the FHLB bond S and L rate differential, disposable income, and the seasonal dummy variables in addition to the lagged NNS variable, the adjusted coefficient of determination (R2) is 0.630. When only the lagged value of the dependent variReceived for publication May 4, 1972. Revision accepted for publication October 18, 1972. * opinions expressed in this note do not necessarily reflect those of the Federal National Mortgage Association. I wish to thank Harry S. Schwartz, Raymond Lombra and Richard Marcis for making useful comments on an earlier draft of this note. 1Jene K. Kwon and Richard M. Thornton, The Federal Home Loan Bank and Savings and Loan Associations: Examination of the Financing of Federal Home Loan Bank Advances, this REVIEW, LIV (Feb. 1972), 97-100. 2Kwon and Thornton make a somewhat similar statement in An Evaluation of the Competitive Effect of FHLB Advances by Savings and Loan Associations, Journal of Finance, XXVI (June 1971), 699-712.

T-HE theory of effective protection was initially developed under the assumption of fixed input coefficients.' Later, the theory was expanded to allow for factor substitution.2 In the process, the concept of effective protection has been changed from defining it as the percentage change in value added per unit of output to defining it as the percentage change in the price of the value-added product. The original definition is: Iv'-v fi = v

SINCE 1966, the Federal Reserve Board has experimented with the use of Regulation Q, the regulation which specifies the maximum interest rates banks are permitted to pay on time and savings deposits, as an active tool of monetary policy. Since September 1966, the Federal Home Loan Bank Board has been empowered to impose ceilings, on savings and loan associations and some mutual savings banks movements in these ceilings to be coordinated with movements in Regulation Q ceilings. In spite of our now-substantial experience with these ceilings, there appears to be no consensus on what purely macro-economic consequences emerge from the employment of deposit ceilings. We need answers to the following two questions: 1) Does the existence of effective deposit ceilings strengthen or weaken standard monetary policy actions? 2) Can the manipulation of deposit ceilings serve as an independent active tool of stabilization policy and, if so, what is the direction of its impact? 1 The official view of the Federal Reserve Board on question (2) appears to be that effective deposit ceilings (attention is typically focused on Q ceilings) are depressive while relaxation of those ceilings is expansionary.2 We know of no clear statement of their position on question (1). There is scant discussion of these issues in the professional literature. In the most general theoretical discussion of deposit ceilings, Tobin (1970) suggests (relying on some casual empirical evidence) that the sign of response of the level of economic activity to changes in interest rate ceilings may vary depending on whether an interest rate variable or a reserve variable is exogenous. He apparently believes that the response we are interested in (letting reserves be held constant) is typically inverse a rise in interest rate ceilings being restrictive. He does not consider question (1). Using a much more restrictive model, Warren Smith (1967) has argued that monetary policy is stronger with effective ceilings (only Q ceilings are considered) than without them. He does not consider question (2). We deal with both questions in this paper. Our analysis suggests that the Tobin and Smith conclusions (on questions (2) and (1), respectively) are incompatible. The model we employ in dealing with these questions differs from both the Smith and Tobin models.3 Our model builds on theirs by incorporating two intermediary claims (both commercial bank time deposits and nonbank intermediary claims), by including currency demand functions, and by considering the impact of incorporation of the price level in the model. Unlike Tobin, we employ the commonplace macro-economic assumption of a single marketable security.4

T HIS paper presents an inquiry into the consumption patterns of a commune. The unit studied is the Israeli kibbutz, which integrates a cooperative enterprise and an egalitarian commune a comprised of 100-200 families or more.' In spite of its unusual size the kibbutz resembles the conventional single-family household as it acts as one indivisible unit in determining the level of its consumption expenditures and outlining the composition of the bill of goods and services to be allocated, mostly in kind, among that household's members.2 Apart from its unusual size, the communal household differs from the conventional household in two respects: The commune's life span is indefinite and it does not have a life cycle. Thus, for instance, the household's size is subject to variation not associated with transformation from one phase to another in a predetermined life cycle. As a predetermined life-span is nonexistent, the commune may be looked upon as a continual household. In this paper an attempt is made to take advantage of the continuous. nature of the communal household in the pursuit of a test of time series and cross-sectional estimates of the consumption-income relationship.3 The first part of this paper focuses on the formulation of a proposition concerning the consumption-income relationship. It also considers alternative ways to estimate the consumption-income coefficients. The second part of this paper presents some empirical results. The generality of our interpretation of the empirical results is discussed in a concluding section.