Table 2 shows the pairwise correlation coefficients and their statistical significance levels for the variables used in our models to address the two core research questions.Footnote 6 The results indicate that disability, gender, race/ethnicity, and organization type and size alone are significantly associated with accommodation requests. Being a person with a disability, a woman, and a racial/ethnic minority is positively correlated with having requested an accommodation. Working in a private and large organization, and being a man, is negatively associated with requests for accommodations. The results also indicate that disability, gender, and type of organization are associated with accommodation outcomes. Specifically, being a woman and working for a private organization is positively correlated with having a request granted. Conversely, and counter to predictions, being a person with a disability, and being a man, is negatively correlated with having accommodation requests granted.
To aid interpretation of these results, and for future research, we include associations with type of disability in Table 2. Results show that having a general health condition is negatively associated with requesting accommodations. This result may be predictable because many such general health conditions may not be perceived as, or actually covered as, disabilities under the ADA. By contrast, having more than one disability, which likely is associated with more severely compromised health, is positively associated with accommodation requests. Interestingly and perhaps predictably, having a mental disability or mental health condition is negatively associated with accommodation receipt, which may be reflective of stigma associated with mental disabilities in general and difficulties associated with documenting some forms of mental disabilities.
In terms of other associations to be explored in future research, being LGBQ is positively associated with having a disability generally, and a mental health condition or disability in particular, as well as with “other” disability, but negatively associated with having a sensory or mobility (arguably visible) disability. Being a woman is positively correlated with report of a mental disability and negatively correlated with having a sensory or more than one disability. Being a man, on the other hand, is positively associated with having a sensory disability, but negatively correlated with having a mental disability. Finally, being older is positively associated with having a disability in general, and with having a sensory or mobility disability, and multiple disabilities, but negatively associated with having a mental health disability.
Tables 3 and 7 provide descriptive statistics on accommodation requests separated by group. Table 3 shows frequency distributions for accommodation requests by individual characteristics, with row percentages (Table 7, in the Appendix, presents column percentages). There is an expected significant relationship between disability status and accommodation requests. Specifically, 42.88% of lawyers with disabilities reported having requested accommodations as compared to 23.23% of those without disabilities.
Distribution of accommodation requests varies by the obviousness or visibility of disability. As predicted, 36.73% of lawyers with non-apparent disabilities report having requested accommodations compared to 49.01% of those with disabilities that fluctuate, and 54.92% of those with apparent disabilities. Disability visibility appears to be associated with an increased likelihood of accommodation request. The proportion of those with non-apparent disabilities (e.g., mental health disabilities) is much higher among those who did not request accommodations. Certain attorneys who otherwise report a health condition, disability, or impairment may choose not to request accommodations, given the disclosure and stigma associated with mental disability.
Sexual Orientation and Gender Identity
The results show that the likelihood of requesting accommodations is associated with an individual’s sexual orientation. Thus, 28.61% of straight/heterosexual respondents report requesting accommodations, as compared to 22.45% of gay/lesbian respondents, 31.82% of respondents who identified as bisexual, 20.00% of respondents who identified as queer, and 37.80% of respondents who reported other sexual orientations. With regard to gender, the results show that women and transgender lawyers are significantly more likely to request accommodations as compared to men.
There is a strong relationship between race and accommodation requests, which suggests that lawyers who are racial/ethnic minorities are more likely to report having requested workplace accommodations. Results show that 27.55% of White non-Hispanic respondents requested accommodations as compared to 33.48% of racial/ethnic minority respondents.
Middle-aged respondents (36-55 years of age) are more likely to request accommodations than either those who are younger or older. Similarly, those with job tenure between six and twenty years are more likely to request accommodations as compared to other groups.
Estimating the Odds of Accommodation Requests
The results in Table 4 show the odds ratio (“OR”) for requesting accommodation as estimated from a series of logistic regression models. We start with a basic model, progressively adding individual characteristics such as disability status, sexual orientation, gender, race, and age. The model including all these characteristics without interactions is our baseline model. We then add control variables, such as job tenure, type and size of organization, 2 × 2 interactions, and illustrative 3 × 3 interactions to the following models. We also conduct a Likelihood Ratio Test (“LR”) to compare the nature of these models. Results from the LR test show that Model 4 is not significantly better than Model 3 in the prediction of our outcome variables. Thus, to simplify interpretation, we focus on the results from Model 3, but have made available for review the results from Model 4 as well.Footnote 7
Results from Model 2 in Table 4 show that, controlling for firm tenure, and type and size of organization, being a person with a disability, a woman, or a transgender individual increases the odds of requesting accommodations. Accordingly, people with disabilities, women, and transgender individuals have higher odds of requesting accommodations as compared to people without disabilities and men, respectively.
Results from Model 3 with 2 × 2 interactions added show that being disabled and a woman increases the odds of requesting accommodations, while being older decreases the odds. Thus, lawyers with disabilities have 2.86 times higher odds (95% CI 1.95–4.20) than those without disabilities of requesting accommodations, among those who identify as men, White, straight, and are 49 years old, holding the control variables constant. The odds of requesting accommodations for women are 2.35 times higher than for men (95% CI 1.76–3.12), among lawyers who are white, straight/heterosexual, without disabilities, and 49 years of age, holding other variables constant. Being LGBQ, transgender, or a racial/ethnic minority was not related to the odds of requesting accommodations.Footnote 8
Once we include 2 × 2 interaction terms, age becomes a significant predictor of accommodation requests, as does the interaction term between sexual orientation and age. This suggests that age is an important conditional contributor when interacted with the other individual characteristics. The coefficient for age implies that the odds of requesting workplace accommodations decline by 2% as age increases (95% CI 0.97–0.99) among lawyers who are men, White, straight, and without disabilities.
None of the interaction terms, except for sexual orientation and age, are associated with the odds of requesting accommodation.Footnote 9 The results indicate that for accommodation requests, while there are no contributing interaction effects between sexual orientation and gender, and sexual orientation and race/ethnicity, there is a substantial relationship between sexual orientation and age. Specifically, the odds of requesting accommodations increase by 1.03 times with every year of tenure for LGBQ lawyers (95% CI 1.00–1.06). Tenure in a law firm is associated generally with increased economic power, which may be particularly enhancing for LGBQ lawyers.
In regard to the control variables, each year of job tenure—again likely reflective of individual economic power in the firm—increases the odds of requesting accommodations by a small magnitude of 1.01 (95% CI 1.00–1.02). However, working for a private organization decreases, relatively, the odds by 0.76 (95% CI 0.62–0.94), controlling for other variables in the model.
Overall, the results suggest that disability, gender, and age are associated with increased odds of requesting accommodations. Since generally the variable interaction terms were not substantial, except for one, we cannot conclude that the effects of disability and sexual orientation vary with other individual characteristics in the models, such as gender, race, and age, in terms of being associated with increases in the probability of accommodation requests.
We next convert the OR of requesting workplace accommodations from Model 3 into predicted probabilities for nine non-differentiated identities of individuals (see Fig. 1). To calculate these probabilities for each identity characteristic, we set all other variables in the model at their overall sample mean.
The results of this analysis show that men have the lowest probability of requesting accommodations (20%), as compared to those without disabilities (23%), White individuals (27%), those who identify as straight/heterosexual (27%), those who identify as LGBQ (28%), racial/ethnic minorities (30%), women (34%), transgender individuals (42%), and individuals with disabilities (44%). Thus, individuals with disabilities have the highest relative probability of requesting accommodations.
In addition, we calculate the probabilities of requesting accommodations from Model 3 for twenty-four multiple identities, or intersectional groupings, with the results presented in Fig. 5 in the Appendix. Results show that White and racial/ethnic minority transgender individuals who also identify as LGBQ and have a disability have the highest probability of requesting accommodations (72%). The top ten intersectional identity groupings with the highest probability of requesting accommodations all include individuals with disabilities.
By comparison, non-Hispanic White men who identify as straight/heterosexual and who do not have disabilities (15%, bottom line of Fig. 5 in the Appendix) show the lowest probability of requesting accommodations, followed by non-Hispanic White men who identify as LGBQ and who do not have disabilities (17%). In Fig. 5 in the Appendix, the bottom nine groups of people, with the lowest probability of requesting accommodations, do not report having disabilities.
Broadly, the results indicate that individuals with disabilities are more likely to request accommodations. There also appears to be a separate effect for gender, with transgender individuals being more likely to request accommodations, as compared to men being less likely to request accommodations. Nonetheless, these intersectional identity interpretations must be viewed as exploratory because the interaction terms between disability and gender are affected by small cell sample sizes.
Finally, we consider the probabilities of requesting accommodations from Model 3 by age for eight distinct intersectional identity groups (illustrated in Fig. 2). The top panel of Fig. 2 illustrates that the probability of requesting accommodations decreases somewhat with age for White non-Hispanic women without disabilities, and decreases for White non-Hispanic women with disabilities. In contrast, women who identify as racial and ethnic minorities, regardless of disability status, show an increase in the probability of requesting accommodations with age (Fig. 2, top panel).
In addition to the race/ethnicity differences, these trends suggest that women with disabilities generally have a higher probability of requesting accommodations, as compared to women without disabilities, at all life points (Fig. 2, top panel). The results also show that LGBQ men and women with and without disabilities show increases in the probability of requesting accommodations with age (Fig. 2, bottom panel). LGBQ men and women with disabilities, at all ages, have an overall higher probability of requesting accommodations as compared to those without disabilities (Fig. 2, bottom panel).
Estimating the Odds of Accommodation Receipt
Table 5 shows the distribution of accommodation requests that were fully, partially, or not granted among groups of lawyers. To ease interpretation, we present row percentages, but column percentages can be found in Table 8 in the Appendix. The results show that, except for race/ethnicity, we are not able to reject the null hypothesis that the variables in this table are independent from the accommodation outcomes. Nonetheless, the following discussion provides some insights on the basic distribution and magnitude of accommodation outcomes.
First, and counter-intuitively, people who do not report a disability have their accommodation requests fully or partially granted at slightly higher rates as compared to those who report having a disability. Specifically, among people with disabilities, 71.96% had their request fully approved as compared to 76.37% of people without disabilities. Similarly, and also unpredicted, people with disabilities are more likely not to have their accommodation requests granted (12.84%), as compared to people without disabilities (8.75%). Those who report their disability as apparent (i.e., more visible or obvious) are relatively more likely to have their accommodation request approved (77.42%) as compared to those who reported their disability as non-apparent (71.11%), or who have a disability that fluctuates (70.83%). Nonetheless, these trends are not statistically significant.
In regard to sexual orientation, straight/heterosexual respondents had a slightly higher likelihood of having their accommodations approved (75.28%) as compared to LGBQ respondents (72.80%). Similarly, while gender does not seem to explain differences in accommodation approval rates, women have the relatively highest likelihood of having their accommodation request fully or partially granted; they also have the lowest likelihood of having their accommodation request not granted when compared to men and transgender lawyers. In considering race, White non-Hispanic respondents are more likely to have their requests fully provided as compared to other racial and ethnic groups. Middle-aged respondents and those with long tenure (more than 20 years) are more likely to have their requests fully granted. Except for race/ethnicity, these other differences are not statistically significant.
The results in Table 6 present the odds ratio (“OR”) of having accommodation requests granted as estimated from a series of logistic regression models. Similar to the previous models, we start with a basic model, progressively adding individual characteristics such as disability status, sexual orientation, gender, race, and age. The model including all these characteristics is again considered the baseline model. We then add control variables (tenure, type of organization, and size of organization), 2 × 2 interactions, and illustrative 3 × 3 interactions in the following models, as we have done prior. We interpret results from this Model 3, but also present results from Model 4 for informational purposes.
Results from Model 2 in Table 6 show that none of the individual characteristic variables are statistically significant.Footnote 10 However, once we include 2 × 2 interactions in Model 3, age becomes statistically significant. Similar to Model 2, the results in Model 3 do not support that any individual characteristics are associated with the likelihood of having accommodations granted other than age. Thus, having a disability, being LGBQ, a woman, transgender, or a racial/ethnic minority is not associated with the odds of a positive accommodation outcome.Footnote 11 The coefficient of age implies that the odds of having a request granted increase with age (OR 1.07, 95% CI 1.01–1.14).
The results further show that only the interaction terms between sexual orientation and race/ethnicity, gender (women) and age, and race/ethnicity and age, are statistically significant. Here, the odds of a positive accommodation outcome decline for LGBQ racial/ethnic minority lawyers, older women lawyers, and older racial/ethnic minority lawyers, relative to comparable others. The effect of sexual orientation on accommodation outcome for racial and ethnic minorities is 0.11 times (95% CI 0.01–1.00) that of White lawyers. The odds ratio for women and racial/ethnic minorities decreases by 0.95 (95% CI 0.89–1.00) and 0.93 (95% CI 0.87–0.99) respectively for a one-year increase in age.
The results additionally suggest that the effect of disability on accommodation outcomes does not vary by sexual orientation, gender, race/ethnicity, and age. Similarly, the effect of sexual orientation does not differ by gender or age. There is no evidence to conclude that the effect of age on accommodation outcome is different for transgender lawyers as compared to men, or that the effect of race/ethnicity on accommodation outcomes is different for men and women.Footnote 12
Working for a large organization, however, substantially increases the odds of having accommodation requests granted: 3.59 times higher (95% CI 0.95–13.56), controlling for other variables in the models. Firm tenure and working at a private organization alone are not associated with accommodation outcomes.Footnote 13
Taken together, the exploratory intersectional analyses suggest that the effect of sexual orientation on accommodation outcomes varies by race/ethnicity and effects of gender (older women only), and by race/ethnicity and age. Thus, the odds of a positive accommodation outcome are lower for LGBQ lawyers who are racial/ethnic minorities as compared to White lawyers, for older women as compared to younger women, and for older racial/ethnic minority lawyers as compared to younger lawyers.
We convert the OR of requesting accommodations from Model 3 in Table 6 into predicted probabilities for nine intersectional identity groupings, as before (see Fig. 3). To calculate these probabilities for each identity characteristic, we again set the other variables in the model at their sample mean. The results show that even though LGBQ lawyers and those with no disabilities have among the lowest probabilities of requesting workplace accommodations (see Fig. 1), they show the highest relative probabilities among the nine identity groups of having their requests granted (96% and 95%). Transgender lawyers and those with disabilities, despite having the highest probability of requesting workplace accommodations, have the lowest relative probabilities of having their requests granted (90%).
As in the prior model, we calculate the probabilities of having accommodation requests granted from Model 3 for twenty-four distinct intersectional identity groupings (see results presented in Fig. 6 in the Appendix). The figure illustrates, for example, that although transgender individuals who identify as racial and ethnic minorities and LGBQ and who also have a disability, have among the highest probability of requesting accommodations (see Fig. 5 earlier in the Appendix), they evidence the lowest probability of having their accommodation request granted (14%). This group is followed by men and women who identify as racial and ethnic minorities and as LGBQ, and also have a disability (19%, 43% respectively, at bottom of Fig. 6 in the Appendix).
As referenced above, the top five groups with the lowest probability of having their accommodation request approved are lawyers with disabilities and who are racial and ethnic minorities. Non-Hispanic White women, men, and transgender individuals who identify as LGBQ and who do not have disabilities have the highest probability of having their workplace accommodations granted as compared to the other groups (98%, 98%, 97%, respectively). In Fig. 6 in the Appendix, the top five groups with the highest probability of having their accommodation request approved are all individuals without disabilities. Taken together, the results indicate that lawyers with disabilities are more likely to request accommodations, but, counterintuitively, those are the lawyers who are less likely to have their accommodation request granted. These outcomes are largely moderated by intersectional multiple identities, such as gender and race/ethnicity.
As a final analysis, and as before, we model the probabilities of having accommodation requests granted from Model 3 by age for eight distinct intersectional multiple identity groupings (see Fig. 4). The top panel of Fig. 4 shows that the probability of having accommodation requests granted decreases with age for minority women with and without disabilities, whereas for White non-Hispanic women with and without disabilities the probability remains relatively constant over time.
Considered in light of the prior models testing the probability of requesting accommodations, these results in the top panel of Fig. 4 suggest that while the probability of requesting workplace accommodations increases with age for minority women, the probability of having such requests granted decreases with age. The converse is the outcome for non-Hispanic White women. In addition, the probability of having workplace accommodation requests granted increases with age for LGBQ men with and without disabilities, whereas it decreases for LGBQ women with and without disabilities. These trends are in contrast with the prior models showing that all of these four groups show an increase in the probability of requesting workplace accommodations with age.