Do minor account laws affect bank account ownership among minors? Does access to an account impact young adult banking and credit behavior? This section discusses the results of our estimates and provides evidence on the downstream outcomes after minors become young adults, including account ownership, alternative financial services use, credit scores, and loan delinquency.
5.1 Minor account laws increase account ownership
In Table 1, data from the SIPP show that among 16-year-olds, state policies allowing minors to own independent bank accounts increase the likelihood that those eligible have accounts. MBL increases the likelihood that 16-year-olds have a checking or savings account by 8.1 percentage points (Column 1), which is about a 29 percent as a marginal effect relative to the mean rate of 28 percent banked. Column (2) further shows that MBL increases the likelihood of having over $100 saved in a checking or savings account by about 8.8 points.[15] The $100 threshold suggests that the individual did not simply open the account with a minimum balance and not use it. Column (3) shows that the MBL policy increases the likelihood of having an independently owned “solo” account. Specifically, the MBL shows a 5.4 percentage point increase in solo-owned accounts, which is 41 percent of the mean rate of 13 percent.[16] Column (4) shows that the MBL policy may shift minors away from joint accounts, though the magnitude is small (0.2 percentage points) and not statistically different from zero. We take this as evidence that the policy generates an overall increase in account ownership among minors, especially solo minor-owned accounts.
The findings from the SIPP data provide evidence that there is a first stage effect of MBL: minor account laws increase account ownership among minors.[17] To validate this even further, we investigate the effect of the MBL on having a checking or savings account among 18 and 19-year-olds in the FDIC data, as displayed in Table 2. Restricting the sample to the ages where an individual reaches adulthood (age 18) provides additional evidence of a first stage from the prior estimates. Indeed, we find that the MBL policy increases the likelihood of having an account by 8 percentage points in Column (1) among 18- to 19-year-olds. While this seems on par with the magnitude in the SIPP, the mean account ownership is much higher among 18- and 19-year-olds (90 percent compared to 28 percent). Thus, the effect is economically and statistically significant, though smaller in size than for 16-year-olds, at only 9 percent of the mean. These results are robust to dropping states that were always treated (as shown in Appendix A, Table A.3). We further show the two-way decomposition of the has account results in Table A.7, where all of the average estimates for each treatment and comparison group are positive and economically significant (Goodman-Bacon 2021).
Table 1: Overall Effects of Minor Account Laws on Account Use for 16-Year-Olds
|
Has Account
|
Over $100 Saved
|
Has Solo Account
|
Has Joint Account
|
|
(1)
|
(2)
|
(3)
|
(4)
|
Minor Banking Law
|
0.0811*
(0.0480)
|
0.0883**
(0.0384)
|
0.0536*
(0.0317)
|
-0.0019
(0.0420)
|
|
|
|
|
|
Mean DV
|
0.2779
|
0.0852
|
0.1329
|
0.1548
|
N
|
3,417
|
3,417
|
3,417
|
3,417
|
*Statistically significant at 10% level; ** at 5% level; *** at 1% level.
|
Notes: Robust standard errors clustered at the state level in parentheses. Linear probability models estimated. All models estimate Equation (1). The data include 16-year-olds only. Data from the 2014 and 2018 waves of the SIPP.
5.2 Minor account laws and financial behavior
We have documented an increase in account ownership after MBL policies allow minors to have independent accounts. We next seek to understand if access to accounts as minors changes the trajectory of financial outcomes as young adults. We again assign treatment based on whether the policy existed in the individual’s state when they were aged 15 through 17.
One important question is: do those early accounts result in higher rates of account ownership at older ages? Using the FDIC data, we expand our sample to follow those treated as minors when they are 18 through 24 years of age. In Table 2 Column (2), the effect of the policy at 18 to 20 years of age is one-quarter the size of the effect at 18 and 19 years of age and is no longer statistically different from zero. Further, Column (3) shows that changing the ages for the sample to 21 through 24 flips the sign of the estimate, although it is not statistically different from zero. While the states being used to identify the variation in Columns (1) and (2) are the same, fewer states identify the variation in Column (3). If we restrict the sample in Column (1) to include only states that identify the effects in Column (3), our effect size becomes larger (0.147 with a standard error of 0.048).
Account ownership is extremely common even among 18- to 20-year-olds (89 percent have accounts). A supplemental analysis in Figure 4 plots coefficients and 95 percent confidence intervals for the effect of minor account laws on whether an individual is banked by each age group through age 29. These estimates confirm that the estimated effect falls to zero by age 20 and becomes a more precisely estimated null at later ages.
Though we see no change in the likelihood of having an account as individuals age into their early 20s, it could still be that there is some learning from early accounts that shifts them away from making financial decisions that are potentially costly. In Columns (4) and (5) of Table 2, we seek to understand how the MBLs affect the use of alternative financial services, which arguably are a substitute for being banked. When focusing on 18- through 20-year-olds in Column (4), we find that the MBLs decrease the use of AFS in the last twelve months by 7.7 percentage points, or 21 percent relative to the mean rate of AFS use of 36 percent. While the coefficient is large, it is not statistically different from zero at the 90 percent level; we are limited in our sample size because the 2009 survey wave asked the AFS question differently (ever versus in the last year). The magnitude of the coefficient suggests that early in life account access may potentially shift young adults away from high-cost borrowing. However, the final column shows that this effect flips signs and remains statistically indistinguishable from zero when we consider those 21- through 24-year-olds. Thus, the relationship changes relatively quickly.
Table 2: Downstream Effects of Minor Account Laws on Account Ownership and AFS Use
|
Has Account
|
Use AFS
|
|
(1)
|
(2)
|
(3)
|
(4)
|
(5)
|
Minor Banking Law
|
0.0811**
|
0.0233
|
-0.0214
|
-0.0773
|
0.0706
|
|
(0.0325)
|
(0.0236)
|
(0.0144)
|
(0.0814)
|
(0.0644)
|
Mean DV
|
0.896
|
0.892
|
0.903
|
0.359
|
0.319
|
|
|
|
|
|
|
N
|
1,202
|
2,463
|
8,896
|
1,226
|
5,138
|
Ages
|
18-19
|
18-20
|
21-24
|
18-20
|
21-24
|
Notes: Robust standard errors clustered at the state level in parentheses. Linear probability models estimated. All models estimate Equation (1). Data from the FDIC Un(der) Banked Survey. “Has Account” equals one if the individual has a checking or savings account and zero otherwise. “Use AFS” equals one if the individual has used alternative financial services in the last 12 months and zero otherwise.
We further probe these AFS results by type of AFS used in Figure 5 to determine if the noisy coefficient comes from one type of product. In Panel A, we plot the coefficients and 95% confidence intervals for 18- to 20-year-olds and in Panel B we do the same from 21- to 24-year-olds. We see that for the younger sample, the noisy decrease in AFS use comes from a reduction in money orders, though it is still not statistically different from zero. However, there is a clear increase in the use of check-cashing services for the older sample.
Supplementary estimates show that these results are robust to dropping states that were always treated (Appendix Table A.5) and adding state-specific linear trends (Appendix Table A.6). The decomposition for the banked outcome (Appendix Table A.7) indicates that the bulk of the average estimates are positive for 18- to 20-year-olds and negative for 21- to 24-year-olds. The substantive comparison comes from treated states versus always treated states and treated states versus never treated states (Goodman-Bacon 2021).
To provide further evidence, we examine the effects of the MBLs on credit and debt behaviors using the CCP data. In the first column of Table 3, we see if MBLs change the age at which one has their first credit report. For this specification, we create a cross-section of over 3.7 million observations that were aged 18 to 37 in 2017, indicating at what age each had their first credit report. Since having a credit report suggests that the individual had sufficient credit activity to generate a credit file, this measure reflects early experience with credit and debt. This finding suggests that the MBL decreases the age at which one had a credit file, though the magnitude is small (0.19 years) relative to the mean of 20.6 years for the first incidence of credit.
If experience with a bank account is a valuable financial experience for minors, we may expect financial decisions to be reflected in higher credit risk scores (where higher scores predict lower risk). In Table 3, we find the effects are very small in magnitude in the case of 18- to 20-year-olds on both credit scores and delinquency behaviors (Columns (2) and (4)), and not statistically different from zero. Among 21- to 24-year-olds, the estimates on credit scores are statistically significant and negative: access to minor accounts reduces credit scores by 2.8 points relative to a mean of 633. MBLs appear to drive financial decisions that slightly reduce credit scores. This is further supported by the fact that we also see increases in delinquency rates due to the MBL for those 21 through 24 years of age in the final column of Table 3. Access to minor accounts increases the likelihood of being behind on account by 0.85 percentage points or about 5 percent as a marginal effect from the mean.
Table 3: Table of Downstream Effects of Minor Account Laws on Credit Outcomes
|
Early Credit
|
Equifax Risk Score
|
Past Due
|
|
|
(1)
|
(2)
|
(3)
|
(4)
|
(5)
|
|
Minor Banking Law
|
-0.188*
|
-0.470
|
-2.801***
|
-0.00011
|
0.0085**
|
|
|
(0.112)
|
(0.917)
|
(0.953)
|
(0.0013)
|
(0.0035)
|
|
Mean DV
|
20.64
|
642.29
|
633.30
|
0.068
|
0.169
|
|
|
|
|
|
|
|
|
N
|
3,756,628
|
2,017,077
|
5,568,291
|
2,284,659
|
6,146,305
|
|
Ages
|
18-37
|
18-20
|
21-24
|
18-20
|
21-24
|
|
Years
|
2017
|
2009-2017
|
2009-2017
|
2009-2017
|
2009-2017
|
|
*Statistically significant at 10% level; ** at 5% level; *** at 1% level.
Notes: Robust standard errors clustered at the state level in parentheses. All models estimate Equation (1) with data from the Federal Reserve Bank of New York/Equifax Consumer Credit Panel (CCP). Early Credit is the first year the individual shows up in the credit file. For this specification, we use a large age range in one point of time (2017). Risk Score is the CCP’s measure for credit scores and ranges from 280 to 850. Past due equals one if the individual is behind on any account and zero otherwise. We estimate linear probability models for that variable.
|
In comparing our effects on AFS use from the FDIC and on credit scores from the CCP data, we note a few caveats. First, the CCP data do not include information on AFS loans or repayments. Second, the FDIC data do not include young adults who still live with their parents, since we only observe cases where the primary respondent in the household is 18- to 24-years-old. Third, the FDIC account ownership question uses a 12-month look-back period. People who had minor-owned accounts but have not opened a new adult-owned account could report being banked at the same time they are using AFS for their financial services. Finally, the FDIC data are self-reported, and as discussed earlier, 28 percent of the FDIC sample report that they are unsure of whether they have used AFS in the last year. While these caveats do not undermine our main estimates, they do highlight the differences across datasets.
[15] Mean savings among those with a checking or savings account is $418.
[16] The reason that average is not zero before the start of the policy is because nearly all states allow state-chartered credit unions to offer independent accounts to minors.
[17] It could be that MBLs affect youth employment by making it easier to have one’s own money. We check this in Table A.4, and though the effects are positive, they are not statistically different from zero.