Knowledge that Transforms

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.

T HIS research provides an estimate of the demand for educational attainment across states within a framework of optimal community choice. The communi,ty is envisioned as having the ability to choose a level of educational attainment for its students which will maximize its net benefits subject to the prevailing technical relationship. The technical relationship specified in this paper considers separately the impact of school inputs, pupil inputs, and social characteristics on educational attainment. Most previous studies of the educational industry have failed to specify a structural model of educational attainment which simultaneously accounts for supply and demand factors. Those which have attempted to measure the effects of inputs on educational attainment have failed to standardize for demand conditions. Expenditure studies have either ignored supply conditions or have resulted in reduced form equations in which the structural parameters cannot be identified. McMahon (1970) estimated the relationship between expenditures and inputs using state data. His conceptual framework did not, however, permit the estimation of the price and income elasticities of demand.' Estimates of income elasticities have been made by Hirsch (1960), Brazer (1959), and Pryor (1968). Their estimates utilized a single equation expenditure function approach which failed to take account of variations in price and the simultaneous aspects of the determination of price and quantity. With the exception of a recent paper by Barlow (1970) estimates of price elasticities have not been published in studies of educational expenditures.2 While Barlow's demand function is similar in form to our own, his single equation estimation procedure ignored the effects of supply changes and thereby introduced the possibility of simultaneous equation bias. The approach utilized in this research is to take account explicitly of the simultaneous nature of demand and supply. Our conceptual framework permits the identification of both price and income elasticities as well as output elasticities for the inputs. Our price and income elasticities are statistically significant and consistent with theoretical expectations. We find that school inputs, pupil inputs, and community characteristics all have important impacts on educational attainment.

N this paper, we first re-estimate the structural equations of the Puerto Rican Model (Dutta and Su, 1969) by seven alternative methods of estimation, and obtain predictions on endogenous variables by using seven sets of alternative estimates. We have obtained three different types of predictions -pure ex-post, pure ex-ante and partial ex-ante. We then compare the predictive power of the various estimators in terms of several descriptive criteria. The specification of the model remains as before (Dutta and Su, 1969). The model consists of 36 jointly determined variables and 36 equations. Twenty-three of these equations are stochastic, and the rest are definitional. Of the 23 behavioral equations, equations (1) to (6) explain consumption expenditures on food, services, other nondurable goods, housing, automobiles, and other durable goods; equations (7) and (8) describe private investment expenditures and changes in inventory investment; equations (9) to (16) cover imports including food, nondurable goods, automobiles, other durable goods, capital goods, raw material, payments on service account, and payments on all other service account; equations (17) to (19) determine exports, namely traditional exports, nontraditional exports, and exports of services; equations (20) to (22) deal with the production sector and explain gross product of three sectors, namely: agriculture, manufacturing, and all other industries and finally equation (23) explains wage share. The relations (24) through (36) are definitional. II Estimation

T HIS paper presents an empirical approach to the aggregation problem. An aggregate dependent variable is properly thought to depend on a long string of disaggregated explanatory variables. There are two extreme ways of analyzing an equation such as this. We may use all of the explanatory variables individually, or we may summarize them in an index and use only the index as an explanatory variable.1 Both of these approaches have serious shortcomings and the purpose of this paper is to consider methods that lie between these extremes. The difficulties with the two extreme approaches are well known. On the one hand data limitations in the form of degrees of freedom inadequacies or multicollinearity usually completely rule out the purely empirical approach. On the other hand, the use of indexes as explanatory variables implies a specification error with unhappy implications worked out by Theil (1954 and 1971), Klein (1946), Allen (1965) and others. It is useful to consider the role of prior judgement in these two approaches. The one analysis excludes prior information about the coefficients and estimates would depend entirely on data evidence if that were possible. The other approach is based on the assumption that the coefficients are known with certainty (up to a factor of proportionality) and no data evidence can alter that constraint. In this paper we will pursue an intermediate course that involves weakening the assumption that the index weights are known with certainty but not to the point that the weights are totally unknown. More formally, this means formulating a proper prior distribution on the coefficients in the unconstrained equation and updating that prior via Bayes Rule. We will analyze import demand functions. One phenomenon specific to import functions is that the domestic indexes that are typically used have weights that are quite unlikely to be adequate for analyzing competition between imports and domestic goods. For example, services substitute little if any with imports and should have low weight in the domestic price index. We hope from our analysis to identify the prices and incomes that are most critical in determining imports and also to obtain better estimates of the aggregate elasticities.

SINCE World War II there has been a tremendous increase in the number of men and women attending college, an increase in the attachment of women to the labor force, and more recently, widespread concern with the economic position of women. Despite these developments, there has been little investigation of the impact of higher education on women's earnings.' This study undertakes such an investigation. Its scope will be limited to considering only monetary costs and benefits associated with education since only these can be measured directly.