Home smoking bans significantly reduce secondhand smoke exposure among children, but parents may offer discordant reports on whether there is a home smoking ban. The purpose of this study was to examine national trends in (a) parental discordance/concordance in the reporting of home smoking bans and (b) correlates of discordant/concordant reports among two-parent households with underage children from 1995 to 2007. Data from the 1995/1996, 1998/1999, 2001/2002, 2003, and 2006/2007 Tobacco Use Supplement of the U.S. Current Population Survey were used to estimate prevalence rates and multinomial logistic regression models of discordant/concordant parental smoking ban reports by survey period. Overall, the percentage of households in which the 2 parents gave discordant reports on a complete home smoking ban decreased significantly from 12.7% to 2.8% from 1995 to 2007 (p < .001). Compared with households where both parents reported a complete smoking ban, discordant reports were more likely to be obtained from households with current smokers (p < .01) across survey periods. Compared with households where both parents reported the lack of a complete home smoking ban, discordant reports were more likely among households with college graduates, no current smokers, and parents with Hispanic ethnicity (p < .05). Parental concordance on the existence of a home smoking ban increased from 1995 to 2007. This suggests estimates of home smoking bans based on just one parent may be more reliable now than they were in the past. Interventions to improve the adoption and enforcement of home smoking bans should target households with current smoker parents.
Background Home smoking bans significantly reduce the likelihood of secondhand smoke exposure among children and non-smoking adults. The purpose of this study was to examine national trends in (1) the adoption of home smoking bans, (2) discrepancies in parental smoking ban reports and (3) household and parental correlates of home smoking bans among households with underage children from 1995 to 2007. Methods The authors used data from the 1995–1996, 1998–1999, 2001–2002, 2003 and 2006–2007 Tobacco Use Supplement of the US Current Population Survey to estimate prevalence rates and logistic regression models of parental smoking ban reports by survey period. Results Overall, the prevalence of a complete home smoking bans increased from 58.1% to 83.8% (p<0.01), while discrepancies in parental reports decreased from 12.5% to 4.6% (p<0.01) from 1995 to 2007. Households with single parent, low income, one or two current smokers, parents with less than a college education or without infants were consistently less likely to report a home smoking ban over this period (p<0.05). Conclusions Despite general improvements in the adoption of home smoking bans and a reduction on parental discrepancies, disparities in the level of protection from secondhand smoke have persisted over time. Children living in households with single parents, low income, current smoker parents, less educated parents or without infants are less likely to be protected by a home smoking ban. These groups are in need of interventions promoting the adoption of home smoking bans to reduce disparities in tobacco-related diseases.
Background: A smoke-free home rule has been associated with reduced cigarette consumption; however, it is unknown whether a home rule is associated with the use of alternative tobacco products (ATP) such as smokeless tobacco products, regular and water pipes, and cigars. This study examined the association between the smoke-free home rules and ever and current use of ATP. Method: Data from the 2010–2011 US Tobacco Use Supplements to the Current Population Survey were analysed using multivariable logistic regressions, including variables related to smoke-free home rules. Results: Overall, 83.9% respondents reported a smoke-free home rule inside their homes; 20.6% of respondents had tried at least one type of ATP, and 3.9% were current users in 2010–2011. Having a smoke-free home rule was associated with lower likelihood of current versus never use of any ATP (adjusted odds ratio (AOR) = 0.80, 95% confidence interval (CI): 0.77–0.83). Among ever users of any ATP, the existence of a smoke-free home rule was associated with lower odds of being a current user (AOR = 0.49, 95% CI: 0.43–0.56). Similar associations were observed for each type of ATP examined (p < 0.05). Conclusion: Smoke-free home rules are associated with lower current ATP use among the US population. Future research should examine whether promoting smoke-free home rules could help to reduce ATP use and related diseases.
We use a combination of administrative and survey data to estimate the fraction of individuals newly enrolled in public health coverage (Wisconsin's combined Medicaid and CHIP program) that had access to private, employer-sponsored health insurance at the time of their enrollment and the fraction that dropped this coverage. We estimate that after expansion of eligibility for public coverage, approximately 20% of new enrollees had access to private health insurance at the time of enrollment and that only 8% dropped this coverage (with the remaining 12% having both private and public coverage). We also identify an upper bound estimate, which suggests that the percentage of new enrollees with private insurance coverage at the time of enrollment is, at most, 27%. These estimates of crowd-out are relatively low compared with estimates from the literature based on Medicaid and CHIP expansions, although based both on different data and on a different method.
The EAR (Excision, Alcoholization, Replantation) method consists of a proper tumor resection, removal of tumor tissue extracorporally, soaking the residual bone shell in 95% alcohol for half an hour, and replantation in situ, the cavity being filled with bone graft or bone cement. Eighty-three cases were treated in this manner, of which 95% were followed for 2 years or more, with complete success in cases with IA lesions and no recurrence in two-thirds of those with IB-IIB lesions. Experimentally, it is proved that alcohol can kill tumor cells completely without interfering with osteogenesis. The joint cartilage, although degenerated, is replaced by newly formed fibrocartilage, thus preserving joint function.
OBJECTIVE: To evaluate the association between prenatal prescription opioid analgesic exposure (duration, timing) and neonatal opioid withdrawal syndrome (NOWS). METHODS: We conducted a retrospective cohort study of Wisconsin Medicaid–covered singleton live births from 2011 to 2019. The primary outcome was a NOWS diagnosis in the first 30 days of life. Opioid exposure was identified with any claim for prescription opioid analgesic fills during pregnancy. We measured exposure duration cumulatively in days (1–6, 7–29, 30–89, and 90 or more) and identified timing as early (first two trimesters only) or late (third trimester, regardless of earlier pregnancy use). We used logistic regression modeling to assess NOWS incidence by exposure duration and timing, with and without propensity score matching. RESULTS: Overall, 31,456 (14.3%) of 220,570 neonates were exposed to prescription opioid analgesics prenatally. Among exposed neonates, 19,880 (63.2%) had 1–6 days of exposure, 7,694 (24.5%) had 7–29 days, 2,188 (7.0%) had 30–89 days, and 1,694 (5.4%) had 90 or more days of exposure; 15,032 (47.8%) had late exposure. Absolute NOWS incidence among neonates with 1–6 days of exposure was 7.29 per 1,000 neonates (95% CI 6.11–8.48), and incidence increased with longer exposure: 7–29 days (19.63, 95% CI 16.53–22.73); 30–89 days (58.96, 95% CI 49.08–68.84); and 90 or more days (177.10, 95% CI 158.90–195.29). Absolute NOWS incidence for early and late exposures were 11.26 per 1,000 neonates (95% CI 9.65–12.88) and 35.92 per 1,000 neonates (95% CI 32.95–38.90), respectively. When adjusting for confounders including timing of exposure, neonates exposed for 1–6 days had no increased odds of NOWS compared with unexposed neonates, whereas those exposed for 30 or more days had increased odds of NOWS (30–89 days: adjusted odds ratio [aOR] 2.15, 95% CI 1.22–3.79; 90 or more days: 2.80, 95% CI 1.36–5.76). Late exposure was associated with elevated odds of NOWS (aOR 1.57, 95% CI 1.25–1.96) when compared with unexposed after adjustment for exposure duration. CONCLUSION: More than 30 days of prenatal prescription opioid exposure was associated with NOWS regardless of exposure timing. Third-trimester opioid exposure, irrespective of exposure duration, was associated with NOWS.
We use a combination of administrative and survey data to estimate the fraction of individuals newly enrolled in public health coverage (Wisconsin's combined Medicaid and CHIP program) that had access to private, employer-sponsored health insurance at the time of their enrollment and the fraction that dropped this coverage.We estimate that after expansion of eligibility for public coverage, approximately 20% of new enrollees had access to private health insurance at the time of enrollment and that only 8% dropped this coverage (with the remaining 12% having both private and public coverage).We also identify an "upper bound" estimate, which suggests that the percentage of new enrollees with private insurance coverage at the time of enrollment is, at most, 27%.These estimates of crowd-out are relatively low compared with estimates from the literature based on Medicaid and CHIP expansions, although based both on different data and on a different method.
We use a combination of administrative and survey data to estimate the fraction of individuals newly enrolled in public health coverage (Wisconsin's combined Medicaid and CHIP program) that had access to private, employer-sponsored health insurance at the time of their enrollment and the fraction that dropped this coverage. We estimate that after expansion of eligibility for public coverage, approximately 20% of new enrollees had access to private health insurance at the time of enrollment and that only 8% dropped this coverage (with the remaining 12% having both private and public coverage). We also identify an upper bound estimate, which suggests that the percentage of new enrollees with private insurance coverage at the time of enrollment is, at most, 27%. These estimates of crowd-out are relatively low compared with estimates from the literature based on Medicaid and CHIP expansions, although based both on different data and on a different method.
