Our paper describes in some detail the rather remarkable financial development that took place in the United States in its earliest decades, starting with the Federalist financial revolution of the 1790s.
In the four decades thereafter, and starting virtually from scratch–although there were, to be sure, colonial precedents that had to be abandoned as much as amplified (see Perkins, 1994, for a discussion of these)–the United States built an articulated, innovative, and modern financial system that equaled that of any other country. One consequence of financial development and innovation was that the United States became history's most successful emerging market, attracting the capital of investors of older nations seeking higher returns. Another was that a wide range of American entrepreneurs, business enterprises, and governments enjoyed more access to financing, domestic and foreign, than did those of other countries. These developments, it seems, placed the United States of the early 19th century on a trajectory of economic growth higher than that of other nations. The die was cast early, perhaps earlier than most economic historians have thought. As its financial system continued to develop (with occasional setbacks) and promote growth over the course of the 19th century, the United States became the world's largest and, in many dimensions, most advanced economy.
We do not merely state this case and document it with traditional historical evidence. We also treat it as a hypothesis and subject it to the sorts of statistical tests that macro economists now apply to rich contemporary, multinational data sets in order to make similar arguments about the effects of financial development on economic growth (e.g., King and Levine 1993b; Levine and Zervos 1998;
Rousseau and Wachtel 2000). Our historical data sets are far more limited than contemporary ones.
But our test results are consistent with modern findings on the key role of financial development in accounting for differences in the economic growth performance of today's countries. The remarkable economic growth of the United States, we think, may very well have been "finance led." Judging by U.S. history, then, the widespread contemporary interest in developing and improving financial systems within and between countries to foster economic growth is not misplaced.
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Appendix A. Construction of Series for the Money Stock, 1790-1820
This appendix explains the methods used to construct a series for the stock of money over the period from 1790 to 1820 and presents the new figures. We begin with the estimates for 1820-1850 that Peter Temin compiled for his 1969 book The Jacksonian Economy, and then replicate his
technique as closely as possible to extend the estimates backward to 1790. Our measure of the broad stock of money sums the total liabilities of banks to the public and specie in the hands of the public. At the time, the banking sector included a central bank (the BUS 1790-1811 and the Second BUS 1817- 1839), banks chartered by individual states, as well as private (non-chartered) banks. Total liabilities for the latter are generally unavailable before 1850 but considered to be small. We discuss each of these components below.
A.1 Obligations of Chartered Banks to the Public
These obligations are estimated by summing notes in circulation, deposits, and amounts due to other banks, and then subtracting amounts due from other banks and holdings of notes of other banks.
Fenstermaker (1965a, Table 10, pp. 66-7) includes these items from 1819 to 1837 for a reasonably broad subset of banks, and also provides the total number of banks in each year. Temin used the ratio of total state banks to those reporting in each year to “blow up” the bank liabilities for Fenstermaker's subset to reflect the nation as a whole. We compute state bank obligations to the public for 1803-1820 by applying this technique to balance sheet items from Fenstermaker (1965b). This source is less representative. In 1818, coverage includes 43 percent of the state banks, but this falls to 29 percent by 1810, 24 percent by 1805, and only 13 percent by 1803. For 1803-1804, the sample includes only state banks in Massachusetts. Other states enter the sample as follows: Virginia 1805; Mississippi 1811; Pennsylvania, South Carolina, the District of Columbia, and New Hampshire 1814; New York 1817; Kentucky 1818. The article does not break down the balance sheet items or the number of banks
by state. We again “blow up” these figures using the actual number of chartered banks to estimate obligations.
We then use the amount of bills discounted at the Massachusetts First National Bank of Boston (from Gras, 1937, pp. 650-651) to extend bank liabilities and specie inside banks backward through 1790. Under the strong assumption that the amount of bills discounted by the few banks that had obtained charters over this period fluctuated similarly to the Massachusetts Bank’s obligations to the public and specie, we then use the ratio of the capital of the Massachusetts Banks and total banking capital in the US in each year to “blow up” bills and discounts to a national level. We then splice this item onto both the obligations and specie series in 1803. This places much emphasis on fluctuations in a single bank's balance sheet, yet these records are the only ones available prior to 1803. It is also important to note that only three banks had obtained state charters as late as 1790, and that the obligations of state banks thus account for only a small portion of the money stock at this time.
A.2. Obligations of the First and Second Banks of the United States to the Public
The 1985 reprint of James Wettereau's Statistics of the First Bank of the United States includes observations for specie holdings, notes in circulation, individual and government deposits, notes of other banks, and balances due to and from other banks at (or around) the end of each year from 1792 to 1801 from balance sheets of the BUS. This allows computation of total BUS obligations to the public until 1802. Records for 1802-1807 do not appear to have survived, and when observations again become available in 1808 they include only specie, notes in circulation, and deposits. It so happens that the sum of individual and government deposits in 1801 is the same as that for total deposits in 1808.
We fill in the intervening years by repeating the 1801 estimates for 1802-1807.
Notes in circulation are 5.08 in 1801 and 4.5 in 1808. Blodget (1806) includes a series for total bank notes for 1801-1807. One way to back out figures for 1801-1807 is to treat BUS notes as a
residual obtained by subtracting state bank notes from total notes. Here, data on notes in circulation from Fenstermaker (1965a, 1984) can be used to estimate those notes attributable to state banks, and the residual should approximate BUS notes. Since there is an overlapping observation in 1801, it can be used to splice the constructed series (1802-1807) to Wettereau's figure for 1801. The manipulation of Fenstermaker's data is similar to that described above for total obligations of state banks. Since notes of other banks and balances due to and from other banks are not available for the BUS after 1801, these cannot be included in the total BUS obligations. Rather, we compute obligations without them from 1802-1808 and ratio splice the result to the total for 1801.
Senate Document 128 (35th Congress, 2nd Session, 1838, pp. 208-211) includes all items required to compute obligations of the second BUS and its branches to the public for 1817-39.
A.3. Specie in the Hands of the Public
Temin obtains official estimates of specie in the hands of the public from 1830 onward, and uses net specie flows to extrapolate the series backward to 1820. Temin’s estimate of the specie stock for 1820 (at $41 mil.) is 64 percent higher than that provided by Ezra Seaman for September of that year. We use North's net specie flows to extend Temin's estimates back to 1807, where Blodget's continuous series (1792-1807) ends. By then, the constructed estimates are 101.5 percent higher than those of Blodget. Lacking a more acceptable technique for estimating specie from 1808 to 1820, however, we apply it back to 1808, and then ratio splice Blodget’s series to the result. This probably overstates the level of specie, but the series exhibits the fluctuations of Blodget's series for 1792-1807. Note that continuing to use specie flows before 1807 produces a series which is much less variable than Blodget's series and the later estimates. From this series for specie stock we subtract specie holdings by state banks (constructed with Fenstermaker's data as above) and specie in the first BUS to obtain an estimate of specie in the hands of the public.
Table A.1
The Stock of Money, 1790-1820 (millions of current dollars) 1790 29.1 1806 62.2 1791 26.4 1807 60.1 1792 35.6 1808 62.8 1793 41.4 1809 56.6 1794 46.0 1810 64.1 1795 42.7 1811 45.9 1796 39.1 1812 63.4 1797 40.8 1813 61.8 1798 39.6 1814 81.9 1799 49.0 1815 75.1 1800 55.5 1816 86.4 1801 53.0 1817 94.1 1802 52.0 1818 84.4 1803 45.0 1819 86.0 1804 70.0 1820 85.0 1805 54.2
Appendix B. Time Series Properties of Data Used in the Empirical Analysis
This section presents results of tests for unit roots and cointegration for the series and systems used in the analysis. We begin with a set of augmented Dickey-Fuller (ADF) tests, which take the presence of a unit root as the null hypothesis. If ADF tests are unable to reject the unit root for a series in levels, yet reject after differencing, there is some justification for treating the series as I(1) in subsequent modeling. The univariate representations for the ADF tests include four lags, which conforms with the
“Akaike plus two” criterion. The trending nature of all series make both constant and trend terms necessary in the levels specifications, while a constant-only regression is used for the first differences.
The log transformation is applied to all series prior to testing. Table B.1 reports test statistics and significance levels. The test rejects the null of a unit root for the data in levels and fails to reject for first differences.
We next test for cointegration in each of the ten VAR systems that we consider. A system with non-stationary variables is classified as cointegrated if a linear combination exists which yields a
)xt ' µ % j
k&1
i'1
'i)xt&i % Axt&k % et,
stationary series when applied to the data. In the tri-variate case, a single cointegrating vector also implies that the error terms of the system are stationary. The technique developed in Johansen (1991) provides a regression-based test for determining both the presence of cointegration and the number of linear stationary combinations which span the space. Each system is modeled as a VAR of the form
(B.1) where xt is a vector containing the three potentially endogenous variables and k is adequately large both to capture the short-run dynamics of the underlying VAR and to generate residuals that approximate the normal distribution. The lag order for each system is chosen with a series of nested likelihood ratio tests. The presence of trends in the data suggest the inclusion of an unrestricted intercept. The
Johansen methodology tests whether the AA matrix in (B.1) is of less than full rank via the trace and maximum eigenvalue statistics.
Table B.2 reports these test statistics for the relevant tri-variate combinations. Rejection of the null hypothesis of no cointegration ® = 0) coupled with a failure to reject the null of at most one
cointegrating vector ® = 1) provides evidence of a single long-run relationship in a given system. This result obtains in all but the first two systems, with rejections of the r = 0 hypothesis at the 10% level or less.
The Johansen regressions also facilitate a test for stationarity of the variables in those systems for which the number of cointegrating vectors is clear. These tests (not shown here) are able to reject the null of stationarity at the 1 percent level for the series which comprise the eight systems that can be characterized by a single cointegrating vector. Since ADF tests are unable to reject the unit root for a given series and the Johansen test rejects stationarity in these cases, our decision to treat all variables as I(1) processes is a reasonable one.
Table B.1
Augmented Dickey-Fuller Test Statistics for Macroeconomic Indicators and Measures of Market Development, 1790-1850
Levels 1st Difference
Macroeconomic Indicators Real GNP Per Capita (GNP)
-3.13 -3.73**
Real Private Domestic Investment Per Capita (INV)
-2.50 -4.32**
Real Imports Per Capita (IMP)
-1.83 -3.95**
Cumulative Business Incorporations (INC)
-2.39 -2.39
Cumulative Non-Financial
Business Incorporations (NFINC)
-2.81 -2.73*
Financial Variables
Real Per Capita Money Stock (MSTOCK)
-2.38 -3.65**
Number of State Chartered Banks (NOB)
-1.44 -3.67**
Number of Listed Securities, Three Markets (3MKT)
-1.58 -3.61**
Number of Listed Securities Per Capita, Three Markets (3MKTP)
-1.55 -3.63**
Number of Listed Securities, Four Markets (4MKT)
-0.97 -2.95**
Number of Listed Securities Per Capita, Four Markets (4MKTP)
-0.89 -2.96**
All variables are in logs. The test regressions use four lags, and include constant and trend for the levels variables and constant only for first differences. * and ** denote rejection of the unit root hypothesis at the 10% and 5% levels respectively, using finite sample critical values are from Fuller (1976), Table 8.5.2.