Discussion Papers In Economics
And Business

Testing the Effectiveness of Market-Based Controls:
Evidence from the Experience of Japan with Short-Term Capital Flows in the 1970s

Taro Esaka and Shinji Takagi

Discussion Paper 12-03

Graduate School of Economics andOsaka School of International Public Policy (OSIPP)
Osaka University, Toyonaka, Osaka 560-0043, JAPANTesting the Effectiveness of Market-Based Controls:
Evidence from the Experience of Japan with Short-Term Capital Flows in the 1970s

Taro Esaka and Shinji Takagi

Discussion Paper 12-03

March 2012

Graduate School of Economics andOsaka School of International Public Policy (OSIPP)
Osaka University, Toyonaka, Osaka 560-0043, JAPANTesting the Effectiveness of Market-Based Controls:
Evidence from the Experience of Japan with Short-Term Capital Flows in the 1970s*
Taro Esaka† and Shinji Takagi††

ABSTRACTThe paper tests the effectiveness of marginal reserve requirements employed by the Japanese
authorities in the 1970s to influence short-term capital flows, thereby contributing to the ongoingdebate on the use of capital controls—market- or price-based ones in particular. While the casefor using market-based controls almost entirely relies on the mixed evidence from the experienceof Chile with unremunerated reserve requirements in the 1990s, testing for their effectiveness on
the volume of inflows is hampered by the endogeneity of such a measure, which is typically
imposed or intensified when inflows surge. We address this problem by applying the method of
propensity score matching and find that an increase in marginal reserve requirements modestly
reduced the volume of short-term capital inflows through non-resident free-yen accounts. The
impact was not statistically significant, however, implying that the price elasticity of short-term
capital flows was small. We conclude that market-based controls must be nearly prohibitive,
perhaps combined with administrative measures, to be effective in a meaningful way.
Keywords: market-based capital controls; price-based capital controls; effectiveness of capital
controls; Japanese capital controls; propensity score matching
JEL classification codes: F32, N25*The authors thank Dongsoo Kang and other seminar participants for useful comments on an
earlier draft of this paper. All remaining errors are theirs.
†

Esaka: Kobe City University of Foreign Studies, 9-1, Gakuenhigashi-machi, Nishi-ku, Kobe,
651-2187, Japan; tesaka@inst.kobe-cufs.ac.jp††

Takagi (corresponding author): Graduate School of Economics, Osaka University, 1-7
Machikaneyama, Toyonaka, Osaka 560-0043, Japan; takagi@econ.osaka-u.ac.jp

I. INTRODUCTION
This paper applies the method of propensity score matching to test the effectiveness ofmarginal reserve requirements employed by the Japanese authorities in the 1970s to influenceshort-term capital flows, thereby contributing to the ongoing debate on the use of capital
controls—market- or price-based ones in particular. Use of capital controls, once condemned by
the mainstream thinking of the economics profession, has assumed greater respectability inrecent years (IEO 2004). Following the global financial crisis of 2007-08, moreover, it has
become increasingly accepted as a legitimate tool of protecting domestic financial markets fromthe vicissitude of cross-border capital flows (Ostry et al. 2011); the Group of Twenty (G20)
finance ministers and central bank governors in October 2011, calling capital controls“residency-based capital flow management measures (CFMs),” stated that these and other CFMs
“may constitute part of a broader approach to protect economies from shocks” and “cancomplement and be employed alongside, rather than substitute for, appropriate monetary,exchange rate, foreign reserve management and prudential policies.” 1
There is a long-standing debate on the effectiveness of capital controls especially as
countries liberalize international payments (Dooley 1996; Edwards 1999; Magud and Reinhart
2006; Habermeier et al. 2011). An open current account allows disguised capital flows throughleads and lags in current payments and receipts as well as through under- and over-invoicing. As
the capital account is progressively liberalized, additional room is created for substitution
between different types of capital flows, so that a measure to control one type of flows is to someextent offset by an increase in other types. As long as a country is committed to the principle of1“G20 Coherent Conclusions for the Management of Capital Flows Drawing on Country Experiences,” as endorsedby G20 Finance Ministers and Central Bank Governors, dated 15 October 2011.
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free trade, there is a natural limit to how much it can feasibly control external payments andreceipts, making it necessary to exempt certain types of capital transactions, such as trade credits
or foreign direct investment (FDI). Inevitably, these would create a channel through which
otherwise controlled capital inflows could take place. The longer controls are in place, the morecreative economic agents will become in finding ways to circumvent them.
The philosophical question of whether capital controls should or could be used undercertain circumstances is different from the practical issue of whether they are effective and, if so,whether the benefits outweigh the costs. Some, arguing against the use of capital controls, havebased their case on the idea that administrative controls must be progressively tightened toremain effective over time and, in view of the potential for substitution, will become increasingly
distortionary (e.g, Mathieson and Rojas-Suarez 1993). It is in this context that so-called marketor price-based controls have become the favorite device of emerging market economies. Suchcontrols, instead of resorting to administrative interventions, attempt to influence capital flows
through the incentives of market participants and are thus considered less distortionary. In 1991,
Chile became the first country in recent history to introduce such a mechanism, called
unremunerated reserve requirements (URR), which required a percentage of relevant inflows to
be deposited in a non-interest-bearing account at the central bank for a certain period of time.
The countries that have followed Chile in introducing URRs or other similar devices in recent
years include, among others, Colombia (1993, 2007), Slovenia (1995), Croatia (2004), Russia(2004), Thailand (2006) and, in the aftermath of the global crisis, Brazil (2009) and Korea (2011).
The generally favorable view of market-based controls within the policymaking
community had been evident even before the use of capital controls became widely accepted.
This came about from around 1996 when a substantial empirical literature began to emerge on-2-the effectiveness of the Chilean URR. Although a review of the literature by Nadal-De Simone
and Sorsa (1999) cautioned against prematurely accepting the Chilean experience as supportiveof the effectiveness of such controls, most had interpreted the body of empirical evidence tobelieve that the URR had lengthened the maturity of capital inflows (if not reduced the total
volume). Subsequent studies tended to be more supportive of the volume effect as well (Gallego
et al. 2002; Le Fort and Lehman 2003). Fischer (2001), speaking on behalf of the International
Monetary Fund (IMF), expressed support for “the use of market-based capital inflow controls:
for example, those that impose a tax on capital inflows.”
The case for market-based controls, however, rests almost entirely on a particularinterpretation of the experience of the Chilean URR, the effectiveness of which “has often been
exaggerated” (Edwards 1999). The available evidence on other countries is far from favorable.
For example, Concha et al. (2011) suggest that the Colombia URR had no impact on either thevolume or the composition of inflows for 1998-2008, a period that covers both the URR of 1993
and that of 2007 (though Cardenas and Barrera (1997) is more favorable to the compositional
impact). Likewise, Clements and Kamil (2009), covering only the URR of 2007, show that the
Colombian measure, while reducing external borrowing, had no impact on the much moresignificant, overall volume of non-FDI inflows. Coelho and Gallagher (2010) and Jittrapanum
and Prasartset (2009) find evidence of a negative impact of the Thai URR on inflows, but the
impact was either economically small or statistically not significant; moreover, the reduction in
net inflows may have been achieved, not by limiting inflows, but through capital flight. 2 Brazil’s

2

Apparently, international investors punished Thailand for introducing the URR. On 18 December 2006, Thailand
imposed a 30 percent URR on all portfolio inflows with maturities of less than one year but, faced with a sharp
decline in equity prices, was compelled to lift the measure on the following day for equity flows.

-3-foreign exchange tax had little effect on the volume or composition of inflows (Hebermeier et al.
2011).
Aside from the potential endogeneity of capital controls (i.e., controls are strengthened
when inflows surge), the ambiguity of the empirical literature on the effectiveness of marketbased controls is hardly surprising. After all the effectiveness of any measure depends, among
other things, on the administrative capacity of the country concerned, the willingness of market
participants to comply with the requirements, and the scope for substitution and circumvention,
as determined by the coverage of the measure as well as country-specific conditions (Hebermeier
et al. 2011). These conditions, for example, include the restrictiveness of the capital accountregime, the sophistication of financial markets, including the extent to which derivative
instruments are available, transactions costs involving a structure of fees and legal procedures,and the strength of incentives provided by interest rate differentials favoring domestic
instruments and the prospect of economic growth over the medium term. In order to understandwhat measure works under what circumstances, and how it works when it does, we must sort out
these factors and isolate the impact of each. This is where the experience of Japan in the 1970s
becomes helpful.
Japan’s use of a market-based control in the 1970s predates the Chilean URR by two
decades or more. At that time, Japan had a highly closed capital account, with all capitaltransactions in principle subject to approval, though the process of capital account liberalization
meant that the presumption of approval was applied to an increasing number of transactions. The
exception was short-term capital flows through non-resident free-yen accounts (FYAs) held at asmall number of commercial banks that the Ministry of Finance had authorized to do foreignexchange business. Foreign investors were freely allowed to move funds into and out of Japan-4-

through FYAs without any formality. Free-yen deposits (FYDs) provided foreign investors witha vehicle of investing in yen assets (though purchasing domestic securities with the funds would
have required approval), while Japanese banks used them as an additional source of funding.
Only when the pace of capital inflows surged did authorities introduce or raise reserve
requirements on FYDs. The measure was rescinded when capital inflows abated.
This experience of Japan provides a kind of controlled experiment in which to isolate theeffectiveness of a market-based capital control by removing the influences of factors other thanchanges in the price incentive itself. Japan’s closed capital account, administrative capacity inthe environment of a strong regulatory culture, and high degree of legal compliance (especiallygiven the fact that the control applied to a small number of tightly regulated banks asintermediaries) meant that there was little room for substitution among different types of capitaltransactions. For foreign investors, FYDs were the only vehicle of freely investing in theJapanese financial market. If the authorities imposed or raised reserve requirements, banks
became less willing to accept new deposits. But foreign investors, who were thus cut out of thedeposit market, had no other means of entering the Japanese market. We believe that, by
analyzing how changes in the balance of FYDs responded to changes in the reserve requirements,
we gain insight into how market-based controls do or do not work in practice and the conditions
required to make them work.
The rest of the paper is organized as follows. Section II reviews the origin and otherspecifics of non-resident free-yen accounts and explains the institutional arrangement underwhich short-term capital flows in general were regulated in Japan prior to the de jure opening ofJapan’s capital account in 1980. Section III presents a preliminary analysis of the impact of
capital control measures on FYD flows, by identifying capita control episodes and using an event-5-

study methodology. In testing how the behavior of FYD flows differed between the capital
control and the non-capital control episodes, the section uses as the counterfactual the sample
average rate of increase in FYDs and the rate of increase obtained from a simple prediction
model. Section IV estimates the average treatment effect of what we identify as intensive capital
controls by the method of propensity score matching. In particular, we estimate a probit model ofthe probability of intensive controls and, in identifying the counterfactual for each capital control
episode, use the probit score to find the closest match. Finally, section V presents concluding
remarks.
II. NON-RESIDENT FREE-YEN ACCOUNTS AS A CHANNEL OF SHORT-TERM CAPITAL FLOWS
The system of non-resident free-yen accounts was established on 1 July 1960 in order toprovide non-residents with a means to keep the proceeds from current exports to Japan in yen.
Along with the authorization to use the yen as a currency of trade settlement, this marked themodest beginning of the yen as an international currency. The non-resident holders of FYAs at
authorized foreign exchange banks were freely allowed to take the deposits out of the country at
any time. Thus, this also represented the establishment of external convertibility for the yen.
These decisions came about against the background of foreign pressure to liberalize Japan’s
trade and payment system as the Japanese economy had recovered from the devastation of WorldWar II and had recorded a double-digit growth year after year. Pressure intensified after majorWest European countries had restored external convertibility for their currencies at the end of
1958. The establishment of free-yen accounts and the authorization for the yen to be used for
trade settlement purposes were the very first measures implemented under the Trade andPayments Liberalization Plan announced by the Japanese government in June 1960 (JMOF 1999).

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The establishment of FYAs had far-reaching implications. Because accounts could be
opened not only with yen proceeds from current exports to Japan but also with the sale of foreignexchange to onshore banks, this meant that non-residents could freely bring short-term capital
into Japan through the banking system. In this sense, Japan’s capital account liberalization
followed a non-standard sequence, by liberalizing short-term capital flows before FDI and longterm portfolio flows; 3 the liberalization of short-term capital flows even predated the full
opening of the current account in 1964 (Takagi 1997). Free-yen deposit rates were subject tomaximum ceilings stipulated by law, which could be higher or lower than the prevailing worldinterest rates. 4 Overseas branches of Japanese banks also participated in this market, bytransferring Eurodollar deposits, converting them into yen, and depositing the proceeds with thehead offices.
FYDs were placed under the overall framework of regulatory supervision over short-term
capital flows during the 1960s. First, from 1962 to 1972, the authorities applied liquidity
requirements to short-term external liabilities, including non-resident free-yen deposits. Under
this system banks were required to keep a certain percentage of short-term external liabilities(such as foreign currency deposits, FYDs, and unsecured borrowing from foreign banks) inliquid foreign assets, such as cash, deposits, call loans, and foreign government securities, andbankers’ acceptances. Initially, the liquidity requirements were set at 10 percent, which wereraised to 25 percent in 1964 before being lowered to 15 percent in 1966; from 1963 to 1965,
marginal liquidity requirements of 35 percent were additionally imposed (on an increase in3

Ishii et al. (2002) discuss what has been widely accepted as the conventional wisdom on the pace and sequencing
of capital account liberalization. Japan’s liberalization of FDI proceeded in several stages from 1969 to 1978, while
the liberalization of portfolio investment was not completed until the mid-1980s, at least on the outflow side.
4
For example, the ceilings were 5.84 percent for time deposits with maturities of more than three months in the caseof non-resident financial institutions; in the case of non-financial firms and individuals, 4.3 percent for three months,
5.5 percent for six months, and 6 percent for a year.

-7-liabilities) in order to discourage inflows. The primary purpose of the liquidity requirements,
however, was prudential. According to Inuta (2000), they were for the most part not binding on
the behavior of foreign exchange banks, especially after 1966 (when the rate was reduced to 15
percent), because most banks for internal risk management purposes kept a sufficiently highshare of their external liabilities in liquid foreign currency assets in any case.
Second, from 1964 to 1972, the Japanese authorities applied outright quantitative controlson the amount of new short-term capital inflows, including through FYAs. This required, foreach bank, the ratio of external liabilities to external assets to remain within the ratio actuallyachieved during a particular period in the past. Basically, what this intended was to allow anincrease in external liabilities, only to the extent that the banks were willing to increase external
lending, thereby limiting the net inflow of short-term funds. Third, in 1968, the authorities
introduced a limit on the amount of so-called “yen conversion,” defined as the amount obtainedby subtracting the balance of relevant foreign assets from the balance of relevant externalliabilities.5 The amount of yen conversion essentially refers to the amount of foreign borrowingthat could be lent out to domestic borrowers. Under this regulation, which was to be terminatedunder the Japan-US Yen-Dollar agreement in 1984 (Frankel 1984), the authorities determined
the yen conversion limit for banks on a monthly average basis (initially 103 percent of the5Consider the following simplified representation of a commercial bank balance sheet:
AssetsForeign currency assets (A)Liabilities
Foreign currency liabilities (B)Non-resident free-yen deposits (C)The amount of yen conversion is equivalent to B + C – A. In December 1973, free-yen deposits were excluded fromthe liabilities side except for the claims to overseas branches of Japanese banks.

-8-average balance during December 1967 and January 1968). The yen conversion limit was
tightened over time as speculation built up on the sustainability of the Bretton Woods parities. 6
Our assessment of market-based capital controls in Japan covers the period 1971-79, andis based not on the foreign currency liquidity requirements on all short-term inflows (whichchanged only three times during 1963-1966) but on the average and marginal reserve
requirements on FYDs, which replaced the liquidity requirements in June 1972. Not only do the
newly introduced reserve requirements provide a richer set of data for our purpose (as they werechanged no less than 9 times in a variety of circumstances over a 7-year period), but the period
1971-79 is one in which a transparent legal framework replaced the opaque “window guidance”
of the 1960s. 7 This means that any regulatory change in the 1970s was a legal measure binding
on foreign exchange banks, and that any response of banks was a result of their own economic
decision, and not of arm-twisting by the authorities. In 1979 the marginal reserve requirements
were reduced to zero, with the system of non-resident free-yen deposits itself abolishedaltogether in December 1980, when the new Foreign Exchange and Trade Control Law cameinto effect making the distinction between residents and non-residents all but meaningless.
Except at the very beginning, non-resident free-yen deposits during most of our sample
period were subject to two sets of regulations: (i) yen conversion limits and (ii) statutory reserve
requirements. The application of these regulations changed in response to the tightening or
easing of pressure on the exchange rate as well as the size and direction of net capital inflows.
For example, following the announcement on 15 August 1971 by the United States government
to suspend the gold convertibility for the dollar, the authorities “requested” the foreign exchange
6For example, the yen conversion limit for domestic banks was set equal to zero after February 1970.
Whereas the authorities had administered the regulation of banks with respect to short-term capital flows throughadministrative fiat, in the immediate aftermath of the currency crisis of August 1971, they placed it within theframework of a ministerial ordinance in September.
7-9-

banks to improve their compliance with the yen conversion limits; following the temporary
floating of the yen, in early September 1971, the authorities further tightened the conversion
limits by changing the application from monthly average balance to daily balance basis. The
conversion limits were then eased as the yen began to face depreciation pressure with thequadrupling of oil prices in November 1973. In December 1973, free-yen deposits were removed
from the application of conversion limits, except for those held by the overseas branches ofdomestic banks.
The adjustment of reserve requirements followed a similar pattern during the capital flow
cycle of 1971-73 (except that the system itself was not introduced until June 1972); from 1974
on, reserve requirements became the only instrument, which the authorities used to control shortterm capital flows through FYDs (Figure 1). When reserve requirements were introduced, they
were made to apply only to an increase over the 21 April-20 May 1972 average balance, at therate of 25 percent, which was then raised to 50 percent a month later. 8 With the oil crisis, therequirements were reduced to 10 percent in December 1973 and further to zero in September
1974. Average reserve requirements (on outstanding balances) were introduced, at the very low
rate of 0.25 percent, only in June 1977 when the yen began to experience appreciating pressure
again. This was followed by the reinstatement of marginal reserve requirements, at the rate of 50
percent in November 1977 and the requirements became prohibitive in March 1978. The
requirements were then reduced to 50 percent in January 1979 and further to zero a month later.
It should be noted that these regulatory actions with respect to FYDs took place within abroader policy context. For example, as appreciation pressure on the yen mounted in the earlysummer of 1971, the government eased restrictions on the holding of foreign currency assets by
8This decision, made on 1 July, followed the reopening of the Tokyo foreign exchange market which had beenclosed for four days after the floating of the British pound.

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residents. Following the suspension of gold convertibility for the dollar, it introduced farreaching restrictions on foreign exchange transactions, including the requirement that authorized
foreign exchange banks limit the growth in foreign liabilities beyond the August 18 level. As theyen began to float permanently on 14 February 1973, the yen was under selling pressure, which
intensified with the October oil crisis. The authorities intervened in the foreign exchange market
by selling dollars ($2.44 billion in November; $1.45 billion in December). At the same time, they
took measures, from November 1973 to February 1974, to promote capital inflows anddiscourage capital outflows, including the abolishment of all restrictions on purchases by nonresidents of domestic securities. From June to August 1974, they further tightened exchange andcapital controls, including the raising of the yen conversion limit for resident foreign banks andthe liberalization of purchases by non-residents of short-term government securities.
III. THE IMPACT OF CONTROL MEASURES ON FYD FLOWS: AN EVENT STUDY
We have grouped these and other control measures taken with respect to free-yen
deposits over 1971-79 into nine “episodes,” which are summarized with dates and additional
details, where relevant, in Table 1. It was necessary to cluster two or more measures into a single
event because they were taken very close to each other, making it impossible to isolate theimpact of one from that of the other(s). The first two episodes concerned (the tightening or
easing of) the application of yen-conversion limits and a quantitative restriction. The next two
involved a market-based measure (a change in average or marginal reserve requirements)
combined with a change in the application of yen conversion limits, while the remaining fiveepisodes entailed only a market-based measure. Strictly speaking, the concurrent use by theJapanese authorities of non-market-based controls may to some extent diminishes the integrity ofour tests to measure the impact of changes in reserve requirements on FYD flows, but we believe- 11 -

that the problem is not so serious especially if we use monthly data as we do. Except for theimposition or elimination of a quantitative ceiling, the non-market-based measures in all othercases only involved a change in the way the benchmark balance was calculated, whether daily ormonthly, and no change was made in the yen conversion limit itself. In a later section, when we
estimate the impact of controls on FYDs, we exclude the first two episodes in one specification
to focus on the impact of market-based controls alone.
In assessing the impact of these measures on FYD flows, we must note that the FYD
balance steadily rose over the period, increasing from a mere 350 billion yen in January 1971 toover 7.5 trillion yen in November 1980; the rate of increase seldom turned negative (Figure 2).
The steady rise in the balance of FYDs could be explained not only by the increasing desire offoreign investors to participate in the growing Japanese financial market with attractive returns,
but also by the expanding volume of Japanese trade, to the extent that a part of FYDs reflected
the volume of trade settlement. Even when the controls became binding or prohibitive, thebalance of FYDs could still increase as banks had no choice but to accept deposits from theirforeign corresponding banks received for trade settlement purposes or from their long-standing
foreign customers who desired to keep their current receipts in yen. Thus, in measuring theimpact of controls, a change in the FYD balance must be compared to the hypothetical change in
the absence of such controls, both of which could be positive. The key is finding the rightcounterfactual.
Obviously, the simplest counterfactual would be the average rate of increase in FYDs
obtained from the sample. An event study was performed on this basis (Table 2). In view of thelimited number of events, we perform no formal test except to note whether each event was“successful” or “not successful” relative to the counterfactual. In particular, we judge a- 12 -

tightening (easing) event to be “successful” if the increase in the balance of FYDs over the postevent window ( r post ) was smaller (larger) than the counterfactual ( r [ post ] ), which is calculated as
the average one-month and two-month changes in the sample that excludes the observations forthe event and three succeeding months. That is to say, the event is judged to be successful if r post
< r [ post ] in the case of tightening and r post > r [ post ] in the case of easing.
Compared to the sample average increases of 2.75 and 5.29 percent for one-month andtwo-month changes, respectively, four and two out of the nine events are judged to have beensuccessful. Only the 6th and 8th events are judged to have been successful in terms of both onemonth and two-month changes. The first of these two events involved only the introduction ofsmall average reserve requirements of 0.25 percent. This may explain why the balance of FYDs
increased over a two-month horizon, though the pace of increase (1.98) was smaller than thesample average of 5.29. In contrast, the latter event entailed a 50-percentage increase in marginalreserve requirements, from 50 percent to the prohibitive level of 100 percent. The balance of
FYDs declined over both one-month and two-month horizons.
Alternatively, following a model used by Fratzscher (2009) for exchange rate forecasting,
we can obtain the counterfactual from a simple linear prediction model, such as:

rt + k = α k + ∑ β k Fi ,t + ε tiWhere t is a time subscript; i refers to the ith factor; rt + k is the realized rate of change in thebalance of FYDs over the forthcoming k months; αk and βk are coefficients for the kth periodahead; ε is a while-noise error; Fi ,t , a set of factors (explanatory variables), is assumed toinclude: the 3-month moving average rate of change in FYD, 3-month moving average Japan- 13 -

U.S. interest rate differential favoring Japanese instruments, 9 and 3-month moving averageexchange market pressure index (EMPI). Following Kaminsky and Reinhart (1999), we define
EMPI as a weighted average of monthly changes in the yen-US dollar exchange rate and foreignexchange reserves (an increase in EMPI means greater depreciation pressure for the yen), where
the weights are given by the respective variables’ standard deviations. All data, except for FYDs
(obtained from the Bank of Japan, Economic Statistics Monthly), come from the International
Monetary Fund, International Financial Statistics, CD-ROM.
The model is estimated separately for each time horizon, by using the sample that
excludes the observations for the event and three succeeding months. The results of an event
study based on the predicted value of FYDs as the counterfactual are summarized in Table 3.
Compared to the results based on the sample mean rates of increase as the counterfactual, thesuccess rate for one-month changes declines (from 4 to 3) while that for three-month changes
rises (from 2 to 4). Given the fact that the first event represents perhaps the tightest measure(enforced by a quantitative ceiling), this result in which it was found successful for both onemonth and two-month changes appears more plausible. Likewise, the 7th event also represents anintensively tight control measure (an introduction of marginal reserve requirements at 50
percent). It makes therefore intuitive sense that, with the counterfactual obtained from a
prediction model, the 7th event becomes successful though only for the two-month horizon. For
the one-month horizon, the event (the second event) that was found successful with the sample
mean rate of change but ceased to be successful with the counterfactual calculated from theprediction model, involved an easing of the controls in place. The positive assessment of the 6th

9

The interest rates are the Treasury bill rate for the US and its equivalent for Japan.

- 14 -

and 8th episodes of controls does not change whether we use the predicted value or the sample
average as the counterfactual.

IV. ESTIMATING THE AVERAGE TREATMENT EFFECT BY PROPENSITY SCORE MATCHING
The event study methodology we have employed has three conceptual problems. First,
the number of events is too small to test the statistical significance formally. Second, the choiceof counterfactual is rather arbitrary. Finally, as a consequence, it is not capable of dealing withthe problem of endogeneity (as the incidence or tightening of a capital control tends to coincidewith increasing free-yen deposits). In econometric terms, this is the problem of self-selection. To
the extent that capital controls are introduced precisely because short-term capital inflows are
surging, a failure to account for this endogeneity is likely to bias our assessment against
concluding that the capital controls were effective.
To address this problem, we use the methodology of propensity score matching, whosebasic idea is to identify as the counterfactual a similar episode in which capital inflows surged
but no control was imposed. Propensity score matching was developed by Rosenbaum and Rubin(1983) and has been used widely in the program evaluation literature (Caliendo and Kopeinig2008). More recently, the methodology has also been applied successfully to monetary andinternational economics, for example, by Glick et al. (2006) (capital controls and currency crises),
Lin and Ye (2007) (inflation targeting) and Fatum and Hutchison (2010) (foreign exchange
market intervention). The advantage of this methodology is that it can formally control for thenon-random selection problem while avoiding need to specify the functional form because it isnonparametric (Heckman et al., 1998; Dehejia and Wahba, 2002; Imbens, 2004; Imbens andWooldridge, 2009).
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The empirical literature employing a variety of propensity score matching methods isoften called the treatment effect literature, because it involves testing the difference between“treated” observations (those for which there is a presumed impact of the event or measureconcerned) and “control” (non-treated but otherwise similar) observations (in order to avoid
confusion with “capital control,” we avoid using the term “control observations” in the rest of thepaper) In the terminology of this literature, therefore, what we attempt to do is to estimate theaverage treatment effect (ATE) of a market-based capital control measure on the volume of FYD
flows.
Our assessment of the effectiveness of capital controls is based on the average treatment
effect of what we call “intense” capital controls (which excludes the introduction of average
reserve requirements at 0.25 percent, which would remain in place until December 1980). These
are not events, as was the case previously, but periods or succession of months where thecontrols measures were in place. Because an easing of controls essentially represents the absenceof controls, an episode of easing can only be considered to be part of the non-treated sample.
Specifically, the “treated” observations (for which controls were applied intensively) include: (i)August-November 1971; (ii) May 1972-November 1973; and (iii) November 1977-January 1979.
These correspond to the periods when a maximum limit was imposed on the outstanding balanceof free-yen deposits, marginal reserve requirements on free-yen deposits were 25% or higher, or
the yen conversion limit was strengthened. Except for the first episode, the tightening of control
entailed a change in the price incentive, with any administrative measure playing a secondaryrole.
By observing how the balance of FYDs changed from month to month (Figure 3), we
find no clear pattern that the rate of increase was consistently negative during the months of- 16 -

intense capital controls (indicated by the shaded areas). During the first episode (when thecontrols were entirely administrative), the rate of increase did fall but the fall started prior to theintensification of controls and it started to rise before the tightened controls were eased. Nor isthere evidence to suggest that rate of increase was smaller during the periods of intensive
controls than the rate of increase during the other months. Rather, except for the latter part of thesecond episode, the rate of increase in FYDs appears to be generally larger when the controls
were intensive.
Formally, we estimate the average treatment effect on the treated (ATT) defined as:

ATT = E[Yi (1) | Ti = 1] − E[Yi (0) | Ti = 1] ,(1)where T is the dummy variable for treatment ( T = 1 when treated; T = 0 when not treated); Yi (1)denotes the outcome when observation i involves the treatment (i.e., intensive capital controls)and Yi (0) is the outcome that obtains in the absence of intensive capital controls. 10 The
fundamental problem in estimating equation (1) is that the second term on the right-hand side
( Yi (0) | Ti = 1 ) is not observable, that is to say, the key point is to identify the counterfactual forE[Yi (0) | Ti = 1] . If we assume that the treatment decision (i.e., the imposition of intensive
controls) is random, the sample mean of the outcome for the non-treated reference group( E[Yi (0) | Ti = 0] ) can be used to estimate the ATT (Table 4). Then, we cannot reject the
hypothesis that the imposition of intensive controls raised (rather than reduced) the rate ofincrease in FYDs at the 10 percent significance level. This is consistent with what we have

10

Both continuous and discrete scalar variables can be applied as the outcome variable in the estimation of theaverage treatment effect. In this analysis, only one of the potential outcomes is observed for each i, and the other isunobserved or missing. The unobserved outcome is called the counterfactual outcome (Caliendo and Kopeinig 2008).
The matching estimators impute the missing potential outcome by using average outcomes for observations withsimilar values of covariates.

- 17 -

previously gathered from looking at Figure 3. When all non-treated observations are used as thereference group, the capital controls had only a small impact, if any, on the volume of short-term
capital flows through FYDs.
The problem of this approach is obvious because the imposition of intensive controls is inpractice not random. If the treatment decision (i.e., the imposition of intensive controls) is
systematically correlated with a set of observable covariates that also affect the outcome, theestimate of the ATT will be biased (Dehejia and Wahba, 2002; Caliendo and Kopeinig, 2008).
To identify the second term on the right-hand side of equation (1), conditional independence
must first be assumed as follows:
Yi (0), Yi (1) ⊥ Ti | X i .

(2)

The conditional independence (or unconfoundedness) assumption implies that, given the pretreatment variables or covariates X, the treatment assignment is independent of the potential
outcomes. Therefore, for observations with similar values of X, the treatment assignment is
random with respect to the potential outcomes. We further assume the following so-called
common support (or overlap) condition: 11
0 < P(Ti = 1 | X i ) < 1 ,(3)where P(Ti = 1 | X i ) = P( X i ) is the conditional probability of adopting the treatment given theobserved covariates X. This probability is what is commonly called a “propensity score.” Thiscondition requires the existence of some comparable reference observations for each treatment
observation.

According to Heckman et al. (1998), we only need to assume that Yi (0) ⊥ Ti | X i (unconfoundedness for controls)and P(Ti = 1 | X i ) < 1 (weak overlap) to identify the ATT.

11

- 18 -

Under the conditional independence assumption and the common support condition, as
specified above, equation (1) can be rewritten as
ATT = E[Yi (1) | Ti = 1, X i ] − E[Yi (0) | Ti = 0, X i ] ,(4)where E[Yi (0) | Ti = 1, X i ] is replaced with E[Yi (0) | Ti = 0, X i ] for the non-treated observations.
One matching method, called covariate matching, pairs the treated observations with the nontreated observations with similar observed values of covariates X. This method, however, isdifficult to implement when the set of covariates X is large.
To overcome the high-dimensionality problem, Rosenbaum and Rubin (1983) propose
matching based on their propensity scores, which are the probabilities of treatment assignment
conditional on X and can be estimated using logit or probit models. This matching is called the“propensity score matching” method. It applies the conditional independence assumption on thepropensity score ( Yi (0), Yi (1) ⊥ Ti | P( X i ) ) and matches the treated observations to the nontreated observations on the basis of the one-dimensional variable P( X i ) , instead of themultidimensional vector X. Using propensity score matching, the ATT can then be estimated as
ATT = E[Yi (1) | Ti = 1, P ( X i )] − E[Yi (0) | Ti = 0, P( X i )] .

(5)

By pairing each treated observation to non-treated observations with similar propensity scores,
this methodology can eliminate sample selection bias to identify a credible counterfactual in theestimation of the ATT (Dehejia and Wahba, 2002; Caliendo and Kopeinig, 2008).
Caliendo and Kopeinig (2008) suggest that the probability of adopting a treatment shouldbe estimated by using observed pre-treatment variables (pre-treatment characteristics). To obtain
the probability of adopting a treatment at time t, we thus use a set of covariates X at time t-1.

- 19 -

This means that equation (5) can be represented (by replacing subscript i with as a time
subscript):
ATT = E[Yt (1) | Tt = 1, P( X t −1 )] − E[Yt (0) | Tt = 0, P( X t −1 )] .

(6)

In what follows, we proceed in the following sequence. First, we estimate the probability of an
intensive capital control by using a probit model, where the dependent variable takes the value ofone (when the control is intensified) or zero (otherwise). Second, we estimate the averagetreatment effect of intensive capital controls on FYD flows by calculating the mean difference inoutcomes between the treated and the non-treated observations based on propensity scorematching.
In finding the right non-treated matches for the treated observations, we employ nearestneighbor and radius matching methods. The nearest-neighbor method (with replacement) pairs
each treated observation with a non-treated observation with the closest propensity score. In
contrast, the radius method pairs a treated observation with all non-treated observations withestimated propensity scores falling within a specified radius r (Dehejia and Wahba 2002):
r=0.005, r=0.01, r=0.05 in our case. Use of four matching methodologies is intended as arobustness check, as no matching method is ideal for all situations (Imbens 2004; Caliendo andKopeinig 2008). The common support condition is imposed, however, in the application of bothtypes of matching methods, namely, we drop the non-treated observations whose estimatedpropensity scores are lower than the lowest score among the treated observations or higher thanthe highest score among the treated observations (Becker and Ichino 2002).
A probit model of the probability of an intensive capital control was estimated, with thefollowing explanatory variables: the 3-month moving average monthly rate of change in FYD attime t-1, the 3-month moving average exchange market pressure index (EMPI) at time t-1, the 3- 20 -

month moving average Japan-U.S. interest rate differential at time t-1, the 3-month moving
average rate of increase in industrial production at time t-1, and a dummy variable for the postSmithsonian floating exchange rate period (i.e., after February 1973). 12 EMPI is as definedpreviously. To ensure robustness, we estimated four different specifications of the model by the
maximum likelihood method, namely, the full model (denoted as [1] in the table), a specificationin which the average lagged rate of change in FYD is removed ([2]), one that excludes May
1972-November 1973 from the sample ([3]), and one that excludes the first episode (ofadministrative controls, August-November 1971), in order to focus on the impact of marketbased controls only (Table 5). The estimated coefficients of most explanatory variables are
significant at conventional significant levels, though not all have the expected signs. The probit
model fits the data quite well (pseudo R2 ranging from 0.30 to 0.52), with the accuracy of
predictions ranging from 77 to 88 percent of the time; the null hypothesis that all slopes are zerois rejected at the 1 percent level.
In all four specifications, the coefficients of EMPI is negative and significant, suggestingthat as the yen faced appreciation pressure the probability of introducing an intensive control
rose. Except for the third specification, the coefficient of industrial production is positive andsignificant, implying that as the economy expanded the probability of imposing a capital control
rose, presumably to mitigate pressure on the exchange rate. Likewise, except for the third
specification, the coefficient of the floating regime dummy is negative and significant. Thismeans that the authorities were more prone to use capital controls when the exchange rate was
fixed (this result does not obtain in the third specification because it excludes the fixed exchange

12

We have verified that no substantive results, including the ultimate assessment of the effectiveness of controls,change even if we alternatively use the one-period lags in place of the 3-month moving averages. The use of 3month averages, however, yields a better fit for the probit model.

- 21 -

rate period of May 1972-November 1973). On the other hand, the coefficient of the interest rate
differential is negative and significant. While this may be counterintuitive (if it means that anincrease in the relative rate of return on domestic assets led to a decline in the volume of inflows),
the overall results in all specifications are found to satisfy the critical balancing property (Becker
and Ichino 2002), namely that the distributional properties of the covariates are similar between
the treated and non-treated observations. 13
The ATT of intensive capital controls on FYD flows can then be estimated on the basis of
the matched samples (Table 6). Matching was carried out separately based on the estimatedpropensity scores from each version of the probit model, as reported in Table 5 ([1][2][3][4]),utilizing four matching methodologies: nearest-neighbor matching with replacement, 14 as well as
radius matching with the choice of r=0.005, r=0.01, r=0.05 (denoted in the table as A, B, C, D,
respectively). 15 Hence, there are 16 combinations ([1A] through [4D]). Except in one case ([2B],
the ATT is estimated to be negative. This means that, with the application of propensity scorematching, the overall results become more favorable to the effectiveness of capital controls (inTable 4, the ATT obtained without propensity score matching was 1.575 and nearly significant at
the 10 percent level). Even in the least favorable case ([2B]), ATT (0.49) is smaller.
None of the ATTs have turned out to be significant at conventional levels, however. The
numerically highest t-statistic is 1.42 ([3D]), which obtains when the radius matching method of
0.05 was used (hence allowing farthest apart observations to be included in the non-treatedsample) and the period May 1972-November 1973 (during which the earlier event study shows13

Lin and Ye (2007) and Caliendo and Kopeinig (2008) suggest that the purpose of the probit estimation is not to
explain accurately the probability of treatment choice but to control effectively for the covariates that affect thetreatment choice and the outcome.
14
In the case of matching with replacement, each non-treated observation can be matched more than once. Matchingwith replacement increases the average quality of matching by increasing the set of possible matches (Caliendo andKopeinig 2008).
15
This was done by using the Stata ado-file of Becker and Ichino (2002).

- 22 -

that the controls were not effective) was excluded. Focusing on the bottom of the table where theestimates from model [4] are reported, we find no evidence that the results are sensitive to the
exclusion or inclusion of the first episode of administrative controls from the sample. On thisbasis, we conclude that while the market-based controls had only a marginal, if any, impact on
the volume of short-term capital flows. This may mean that the price elasticity of short-term
capital inflow is small, so that market-based controls may need to become prohibitive, perhaps
accompanied by administrative measures, to be effective in a meaningful way.
V. CONCLUSION
We have tested the effectiveness of marginal reserve requirements used by Japan in the
1970s to influence short-term capital flows through free-yen deposits. Capital controls, marketbased ones in particular, are receiving increasing respectability in the policy making community
as a legitimate tool of protecting domestic financial markets from the vicissitude of cross-border
capital flows, but the case for them almost entirely relies on the mixed evidence from theexperience of Chile with unremunerated reserve requirements in the 1990s. We contribute to theongoing debate on the use of capital controls by providing evidence from the experience of Japanin the 1970s when the capital account was closed except for short-term capital flows throughFYDs.
Testing for the effectiveness of a capital control on the volume of inflows is notoriously
difficult because of the endogeneity of such a measure, which is typically imposed or intensified
precisely when inflows surge. Thus, unless we address this problem, any assessment ofeffectiveness is biased against concluding that the measure was effective. We applied the methodof propensity score matching to account for the endogeneity, and found that an increase in
marginal reserve requirements only modestly reduced the volume of short-term capital inflows- 23 -

through non-resident free-yen accounts. This was in contrast to the results that obtained when theendogeneity was not properly controlled for. Even so, the impact was not statistically significant,
implying that the price elasticity of short-term capital flows was small. We conclude that marketbased controls must be nearly prohibitive, perhaps combined with administrative measures, to beeffective in a meaningful way.
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Concha, Alvaro, Arturo Jose Galindo, and Diego Vasquez, 2011, “An Assessment of Another
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Shuppan.
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Figure 1
Average and Marginal Reserve Requirements on Free-Yen Deposits, 1972-79 (in percent)
0.4
0.35
0.3
0.25
0.2
0.15
0.1
0.05
0

Marginal (left scale)Jun 1979

Oct 1979

Feb 1979

Oct 1978Jun 1978Oct 1977

Feb 1978Jun 1977

Feb 1977

Oct 1976

Jun 1976

Feb 1976

Oct 1975

Jun 1975

Feb 1975

Jun 1974

Oct 1974

Feb 1974

Oct 1973

Jun 1973

Feb 1973

Jun 1972

Oct 1972

100
90
80
70
60
50
40
30
20
10
0

Average (right scale)

Source: Bank of Japan, Economic Statistics Monthly, March 1986

Figure 2
Free-yen Deposits, 1971-1980 (Left scale: outstanding monthly balance in billions of yen; Right scale:monthly rate of change in percent)
30
25
20
15
10
5
0
-5
-10
-15
1971M1
1971M5
1971M91972M1
1972M5
1972M9
1973M1
1973M5
1973M9
1974M1
1974M5
1974M9
1975M1
1975M5
1975M9
1976M1
1976M5
1976M9
1977M1
1977M5
1977M9
1978M1
1978M5
1978M9
1979M1
1979M5
1979M9
1980M1
1980M5
1980M9

8000
7000
6000
5000
4000
3000
2000
1000
0

Note: Non-resident free-yen deposits in all authorized foreign exchange banks.
Sources: Bank of Japan, Economic Statistics Monthly and Economic Statistics Annual, various issues.

- 28 -

Figure 3
Monthly Rate of Change in Free-yen Deposits during Intensive Control Periods, 1971-1980(In percent)
30

1

25
20
15
10
5
0
-5
-10
0
1971M1
1971M5
1971M91972M1
1972M5
1972M9
1973M1
1973M5
1973M9
1974M1
1974M5
1974M9
1975M1
1975M5
1975M9
1976M1
1976M5
1976M9
1977M1
1977M5
1977M9
1978M1
1978M5
1978M9
1979M1
1979M5
1979M9
1980M1
1980M5
1980M9-15

Notes: Non-resident free-yen deposits in all authorized foreign exchange banks; the shaded areas indicate themonths in which controls were intensified.
Sources: Bank of Japan, Economic Statistics Monthly and Economic Statistics Annual, various issues: authors’estimates.

- 29 -

Table 1:
Measures to Control Short-Term Capital Flows through Free-yen Accounts, 1971-1980
Date
1. Tightening,
August-September
1971
2. Easing, December
1971-January 1972
3. Tightening, MayJuly 1972

4. Easing, December
1973
5. Easing, September
1974
6. Tightening, June1977
7. Tightening,
November 1977
8. Tightening, March
1978

Measures, as applied to authorized foreign exchange banks
Requirements introduced to keep the outstanding balance of free-yen deposits at or below
18 August balance (19 August 1971); the base changed to 27 August balance (28 August1971); the application of the “yen conversion limit” changed from monthly average balance
to daily balance basis (7 September 1971)
The application of the “yen conversion limit” reverted back to average monthly basis (21December 1971); restrictions introduced in August 1971 abolished on the maximum
balance of free-yen deposits (6 January 1972)
The application of the yen conversion limit changed from monthly to daily (May 1972);
25% marginal reserve requirements imposed on free-yen deposits on the basis of 21 April20 May 1972 balance (1 June 1972); marginal reserve requirements raised to 50% on thebasis of 21 May-20 June 1972 balance (1 July 1972)
Marginal reserve requirements lowered from 50% to 10% on the basis of 21 May-20 June1972 balance (10 December 1973); the yen conversion limit abolished for free-yen deposits,with the exception of head office/branch accounts (17 December 1973)
Marginal reserve requirement lowered from 10% to zero (12 September 1974)
0.25% reserve requirements introduced on the outstanding balances of free-yen deposits (1June 1977).
50% marginal reserve requirements introduced on the basis of 1-31 October 1977 balance(22 November 1977)
Marginal reserve requirements raised from 50% to 100%, on the basis of 1-28 Februarybalance (18 March 1978)

9. Easing, JanuaryFebruary1979

Marginal reserve requirements lowered from 100% to 50%, on the basis of 1-28 February
1978 balance (17 January 1979); marginal reserve requirements further reduced to zero
percent (10 February 1989)
Sources: Bank of Japan (1988); Fukao (1990); and Inuta (2000).

- 30 -

Table 2
An Event Study: The Impact of Tightening or Easing Controls on Free-yen Deposits, 1971-1980

Date(number of measures)

1 August-September 1971 (2)December 1971-January21972 (2)3 May-July 1972 (3)4 December 1973 (2)5 September 1974 (1)6 June 1977 (1)
7 November 1977 (1)8 March 1978 (1)9 January-February 1979 (2)Tighteningor easingDirection of
Two-monthOne-monthchange relativechangechangeto the counter(%)(%)
factual of 2.75

Direction ofchange relative
to the counterfactual of 5.29

Tightening-3.45

Successful

6.63

Not successful

Easing4.71

Successful-0.44

Not successful

Tightening
EasingEasingTightening
Tightening
Tightening
Easing

20.67
-2.14
-3.53
-1.29
7.94
-2.42
1.24

Not successful
Not successful
Not successful
Successful
Not successful
Successful
Not successful

31.36
-5.55
-4.85
1.98
9.12
-2.66
-7.59

Not successful
Not successful
Not successful
Successful
Not successful
Successful
Not successful

Note: The counterfactuals are simply calculated as the average one-month and two-month changes in the sample that
excludes the event month and subsequent three months.

- 31 -

Table 3
An Event Study: The Impact of Tightening or Easing Controls on the Total Balance of Free-yen Deposits,
1971-1980
Date (number ofmeasures)1
2
3

August-September
1971 (2)December 1971January 1972 (2)May-July 1972 (3)Counterfactual:
predictedchange(%)

Direction ofchangerelative tocounterfactual

Twomonthchange(%)

Counterfactual:
predictedchange (%)

Direction ofchangerelative tocounterfactual

Tighteningor easing

One-monthchange (%)

Tightening

-3.45

9.99

Successful

6.63

18.09

Successful

Easing

4.71

5.42

Not successful

-0.44

9.86

Not successful

Tightening

20.67

3.02

Not successful

31.36

5.57

Not successful

4

December 1973 (2)Easing-2.14-0.35 Not successful-5.55-0.25 Not successful
September 1974
5
Easing-3.53
3.31 Not successful-4.85
6.15 Not successful(1)
6 June 1977 (1)Tightening
-1.29
2.51
Successful
1.98
4.81
Successful
November 1977
7
Tightening
7.94
4.93 Not successful
9.12
9.28
Successful(1)
8 March 1978 (1)Tightening
-2.42
5.45
Successful
-2.66
10.21
Successful
January-February
9
Easing1.24
1.32 Not successful
-7.59
2.99
Not success
1979 (2)Note: The counterfactuals are calculated as the predicted one-month and two-month changes from a simple linear prediction
model.

Table 4The Average Treatment Effect on the Treated (ATT):
Months of Intensive Controls versus Non-Treated Months
Number ofmonths“Treated” months: intensive
38
controls (T=1)
All “non-treated” months (T=0) 81
Memorandum: all months
119

Average monthly rate of change
in free-yen deposits (%)
3.648
2.074
2.576

ATT (t-statistic)
1.575(1.457)

Note: The index (T) takes the value of one for: August-November 1971; May 1972-November 1973; and November
1977-January 1979.

- 32 -

Table 5
Probit Estimates of Intensive Capital Controls
[1]
0.045
(1.04)[2]-0.233***(-3.89)3-month moving averageinterest rate differential (t-1)(%)
3-month moving average rateof increase in industrial
production (t-1) (%; year-onyear)
Dummy for the floatingexchange rate period
Constant

3-month moving averagemonthly rate of change infree-yen deposits (t-1) (%)
3-month moving average rateof change in EMPI (t-1)

Number of observations
Log pseudo likelihood
Pseudo R2
H0: All slopes=0, Wald chi2[p-value]
AIC
BIC
Correctly classified (%)

[3]
0.070
(1.44)

0.033(0.68)[4]-0.257***(-4.29)-0.403***(-4.27)-0.224***(-3.18)-0.165**(-2.35)-0.181***(-2.61)-0.239***(-2.87)-0.163***(-2.26)0.174***(4.04)

0.172***(4.03)

0.006(0.33)

0.175***(4.09)-0.929*(-1.90)-1.160***(-2.82)0.925*(1.80)-0.901*(-1.73)-1.647***(-3.02)
119
-50.0
0.33

-1.389***(-3.29)
119
-50.6
0.32

-5.540***(-3.42)
100
-23.6
0.52

-1.649***(-3.05)
115
-49.2
0.30

32.31 [0.00]

29.06 [ 0.00]

25.64 [0.00]

25.44 [0.00]

112.0
128.7
79.0

111.2
125.1
76.5

59.1
74.8
88.0

110.4
126.8
79.1

Notes: The numbers in parentheses are z-statistics, which are calculated using heteroskedasticity-robust standard
errors; ***, **, * indicate statistical significance at the 1, 5, 10 % levels, respectively.

- 33 -

Table 6
Propensity Score Matching Methods:
The Effect of Intensive Capital Controls on Free-yen Deposits
Matching method

Nearestneighbor

Model (covariates)
Average treatment effect on treated(ATT)
Analytical standard errors
T-statistics
Treatment/non-treatment
Common support condition satisfied

[1A]

-1.66

-1.51

2.60
-0.36
38/17

2.63
-0.37
16/17

2.05
-0.81
27/25

1.66
-0.91
30/38

Yes

Yes

Yes

Yes

[2A]

Probit model [2]
[2B]
[2C]

[2D]

-2.91

0.49

-0.40

-0.38

2.83
-1.03
38/18

2.10
0.23
19/26

1.58
-0.26
28/40

1.47
-0.26
35/40

Yes

Yes

Yes

Yes

[3A]

Probit model [3]
[3B]
[3C]

[3D]

-2.95

-2.38

-3.04

-3.47

2.35
-1.26
19/8

3.90
-0.61
4/5

2.97
-1.02
11/10

2.44
-1.42
13/14

Yes

Model (covariates)
Average treatment effect on treated(ATT)
Analytical standard errors
T-statistics
Treatment/non-treatment
Common support condition satisfied

[1D]

-0.97

Model (covariates)
Average treatment effect on treated(ATT)
Analytical standard errors
T-statistics
Treatment/non-treatment
Common support condition satisfied

Radius (r=0.05)-0.93

Model (covariates)
Average treatment effect on treated(ATT)
Analytical standard errors
T-statistics
Treatment/non-treatment
Common support condition satisfied

Radius(r=0.025)
Probit model [1]
[1B]
[1C]

Radius (r=0.01)

[4A]

Yes

Yes

Yes

Probit model [4]: excluding administrative controls
[4B]
[4C]
[4D]

-2.65

-3.81

-1.91

-1.16

2.09
-1.27
34/17

3.05
-1.25
15/13

1.92
-0.99
25/32

1.68
-0.69
30/39

Yes

Yes

Yes

Yes

Note: ATT is the mean difference between the treated (with intensive capital controls) and the non-treated (without
such controls) observations in terms of monthly changes in free-yen deposits. We employ propensity score matchingbased on the estimated probit model of Table 5 to choose the non-treated observations against which the treatedobservations are compared.

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