Pub Date : 2024-05-01DOI: 10.1136/bmjmed-2023-000791
Christine Cunningham, Louis Fisher, Christopher Wood, Victoria Speed, Andrew D Brown, Helen Curtis, Rose Higgins, Richard Croker, Ben FC Butler-Cole, David Evans, Peter Inglesby, Iain Dillingham, Sebastian CJ Bacon, Elizabeth Beech, Kieran Hand, Simon Davy, Tom Ward, George Hickman, Lucy Bridges, Thomas O'Dwyer, Steven Maude, Rebecca M. Smith, Amir Mehrkar, Liam C Hart, Chris Bates, Jonathan Cockburn, John Parry, Frank Hester, Sam Harper, Ben Goldacre, Brian MacKenna
To investigate the effect of the covid-19 pandemic on the number of patients with group A streptococcal infections and related antibiotic prescriptions.Retrospective cohort study in England using OpenSAFELY-TPP.Primary care practices in England that used TPP SystmOne software, 1 January 2018 to 31 March 2023, with the approval of NHS England.Patients registered at a TPP practice at the start of each month of the study period. Patients with missing data for sex or age were excluded, resulting in a population of 23 816 470 in January 2018, increasing to 25 541 940 by March 2023.Monthly counts and crude rates of patients with group A streptococcal infections (sore throat or tonsillitis, scarlet fever, and invasive group A streptococcal infections), and recommended firstline, alternative, and reserved antibiotic prescriptions linked with a group A streptococcal infection before (pre-April 2020), during, and after (post-April 2021) covid-19 restrictions. Maximum and minimum count and rate for each infectious season (time from September to August), as well as the rate ratio of the 2022-23 season compared with the last comparably high season (2017-18).The number of patients with group A streptococcal infections, and antibiotic prescriptions linked to an indication of group A streptococcal infection, peaked in December 2022, higher than the peak in 2017-18. The rate ratios for monthly sore throat or tonsillitis (possible group A streptococcal throat infection), scarlet fever, and invasive group A streptococcal infection in 2022-23 relative to 2017-18 were 1.39 (95% confidence interval (CI) 1.38 to 1.40), 2.68 (2.59 to 2.77), and 4.37 (2.94 to 6.48), respectively. The rate ratio for prescriptions of first line, alternative, and reserved antibiotics to patients with group A streptococcal infections in 2022-23 relative to 2017-18 were 1.37 (95% CI 1.35 to 1.38), 2.30 (2.26 to 2.34), and 2.42 (2.24 to 2.61), respectively. For individual antibiotic prescriptions in 2022-23, azithromycin showed the greatest relative increase versus 2017-18, with a rate ratio of 7.37 (6.22 to 8.74). This finding followed a marked decrease in the recording of patients with group A streptococcal infections and associated prescriptions during the period of covid-19 restrictions where the maximum count and rates were lower than any minimum rates before the covid-19 pandemic.Recording of rates of scarlet fever, sore throat or tonsillitis, and invasive group A streptococcal infections, and associated antibiotic prescribing, peaked in December 2022. Primary care data can supplement existing infectious disease surveillance through linkages with relevant prescribing data and detailed analysis of clinical and demographic subgroups.
{"title":"Incidence and treatment of group A streptococcal infections during covid-19 pandemic and 2022 outbreak: retrospective cohort study in England using OpenSAFELY-TPP","authors":"Christine Cunningham, Louis Fisher, Christopher Wood, Victoria Speed, Andrew D Brown, Helen Curtis, Rose Higgins, Richard Croker, Ben FC Butler-Cole, David Evans, Peter Inglesby, Iain Dillingham, Sebastian CJ Bacon, Elizabeth Beech, Kieran Hand, Simon Davy, Tom Ward, George Hickman, Lucy Bridges, Thomas O'Dwyer, Steven Maude, Rebecca M. Smith, Amir Mehrkar, Liam C Hart, Chris Bates, Jonathan Cockburn, John Parry, Frank Hester, Sam Harper, Ben Goldacre, Brian MacKenna","doi":"10.1136/bmjmed-2023-000791","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000791","url":null,"abstract":"To investigate the effect of the covid-19 pandemic on the number of patients with group A streptococcal infections and related antibiotic prescriptions.Retrospective cohort study in England using OpenSAFELY-TPP.Primary care practices in England that used TPP SystmOne software, 1 January 2018 to 31 March 2023, with the approval of NHS England.Patients registered at a TPP practice at the start of each month of the study period. Patients with missing data for sex or age were excluded, resulting in a population of 23 816 470 in January 2018, increasing to 25 541 940 by March 2023.Monthly counts and crude rates of patients with group A streptococcal infections (sore throat or tonsillitis, scarlet fever, and invasive group A streptococcal infections), and recommended firstline, alternative, and reserved antibiotic prescriptions linked with a group A streptococcal infection before (pre-April 2020), during, and after (post-April 2021) covid-19 restrictions. Maximum and minimum count and rate for each infectious season (time from September to August), as well as the rate ratio of the 2022-23 season compared with the last comparably high season (2017-18).The number of patients with group A streptococcal infections, and antibiotic prescriptions linked to an indication of group A streptococcal infection, peaked in December 2022, higher than the peak in 2017-18. The rate ratios for monthly sore throat or tonsillitis (possible group A streptococcal throat infection), scarlet fever, and invasive group A streptococcal infection in 2022-23 relative to 2017-18 were 1.39 (95% confidence interval (CI) 1.38 to 1.40), 2.68 (2.59 to 2.77), and 4.37 (2.94 to 6.48), respectively. The rate ratio for prescriptions of first line, alternative, and reserved antibiotics to patients with group A streptococcal infections in 2022-23 relative to 2017-18 were 1.37 (95% CI 1.35 to 1.38), 2.30 (2.26 to 2.34), and 2.42 (2.24 to 2.61), respectively. For individual antibiotic prescriptions in 2022-23, azithromycin showed the greatest relative increase versus 2017-18, with a rate ratio of 7.37 (6.22 to 8.74). This finding followed a marked decrease in the recording of patients with group A streptococcal infections and associated prescriptions during the period of covid-19 restrictions where the maximum count and rates were lower than any minimum rates before the covid-19 pandemic.Recording of rates of scarlet fever, sore throat or tonsillitis, and invasive group A streptococcal infections, and associated antibiotic prescribing, peaked in December 2022. Primary care data can supplement existing infectious disease surveillance through linkages with relevant prescribing data and detailed analysis of clinical and demographic subgroups.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":" 6","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-05-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"141130559","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-04-01DOI: 10.1136/bmjmed-2023-000544
Jane E Harding, J. Alsweiler, Taygen Edwards, C. J. McKinlay
Low blood concentrations of glucose (hypoglycaemia) soon after birth are common because of the delayed metabolic transition from maternal to endogenous neonatal sources of glucose. Because glucose is the main energy source for the brain, severe hypoglycaemia can cause neuroglycopenia (inadequate supply of glucose to the brain) and, if severe, permanent brain injury. Routine screening of infants at risk and treatment when hypoglycaemia is detected are therefore widely recommended. Robust evidence to support most aspects of management is lacking, however, including the appropriate threshold for diagnosis and optimal monitoring. Treatment is usually initially more feeding, with buccal dextrose gel, followed by intravenous dextrose. In infants at risk, developmental outcomes after mild hypoglycaemia seem to be worse than in those who do not develop hypoglycaemia, but the reasons for these observations are uncertain. Here, the current understanding of the pathophysiology of neonatal hypoglycaemia and recent evidence regarding its diagnosis, management, and outcomes are reviewed. Recommendations are made for further research priorities.
{"title":"Neonatal hypoglycaemia","authors":"Jane E Harding, J. Alsweiler, Taygen Edwards, C. J. McKinlay","doi":"10.1136/bmjmed-2023-000544","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000544","url":null,"abstract":"Low blood concentrations of glucose (hypoglycaemia) soon after birth are common because of the delayed metabolic transition from maternal to endogenous neonatal sources of glucose. Because glucose is the main energy source for the brain, severe hypoglycaemia can cause neuroglycopenia (inadequate supply of glucose to the brain) and, if severe, permanent brain injury. Routine screening of infants at risk and treatment when hypoglycaemia is detected are therefore widely recommended. Robust evidence to support most aspects of management is lacking, however, including the appropriate threshold for diagnosis and optimal monitoring. Treatment is usually initially more feeding, with buccal dextrose gel, followed by intravenous dextrose. In infants at risk, developmental outcomes after mild hypoglycaemia seem to be worse than in those who do not develop hypoglycaemia, but the reasons for these observations are uncertain. Here, the current understanding of the pathophysiology of neonatal hypoglycaemia and recent evidence regarding its diagnosis, management, and outcomes are reviewed. Recommendations are made for further research priorities.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"37 4","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-04-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140796871","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-04-01DOI: 10.1136/bmjmed-2023-000802
Maryam Mooghali, Aaron P. Mitchell, Joshua J. Skydel, Joseph S. Ross, J. Wallach, Reshma Ramachandran
Objectives To evaluate National Comprehensive Cancer Network (NCCN) guideline recommendations for oncology drug treatments that have been granted accelerated approval, and to determine whether recommendations are updated based on the results of confirmatory trials after approval and based on status updates from the US Food and Drug Administration (FDA). Design Cross sectional study. Setting US FDA and NCCN guidelines. Population Oncology therapeutic indications (ie, specific oncological conditions for which the drug is recommended) that have been granted accelerated approval in 2009-18. Main outcome measures NCCN guideline reporting of accelerated approval status and postapproval confirmatory trials, and guideline recommendation alignment with postapproval confirmatory trial results and FDA status updates. Results 39 oncology drug treatments were granted accelerated approval for 62 oncological indications. Although all indications were recommended in NCCN guidelines, accelerated approval status was reported for 10 (16%) indications. At least one postapproval confirmatory trial was identified for all 62 indications, 33 (53%) of which confirmed benefit; among these indications, NCCN guidelines maintained the previous recommendation or strengthened the category of evidence for 27 (82%). Postapproval confirmatory trials failed to confirm benefit for 12 (19%) indications; among these indications, NCCN guidelines removed the previous recommendation or weakened the category of evidence for five (42%). NCCN guidelines reflected the FDA's decision to convert 30 (83%) of 36 indications from accelerated to traditional approval, of which 20 (67%) had guideline updates before the FDA's conversion decision. NCCN guidelines reflected the FDA's decision to withdraw seven (58%) of 12 indications from the market, of which four (57%) had guidelines updates before the FDA's withdrawal decision. Conclusions NCCN guidelines always recommend drug treatments that have been granted accelerated approval for oncological indications, but do not provide information about their accelerated approval status, including surrogate endpoint use and status of postapproval confirmatory trials. NCCN guidelines consistently provide information on postapproval trial results confirming clinical benefit, but not on postapproval trials failing to confirm clinical benefit. NCCN guidelines more frequently update recommendation for indications converted to traditional approval than for those approvals that were withdrawn.
{"title":"Characterization of accelerated approval status, trial endpoints and results, and recommendations in guidelines for oncology drug treatments from the National Comprehensive Cancer Network: cross sectional study","authors":"Maryam Mooghali, Aaron P. Mitchell, Joshua J. Skydel, Joseph S. Ross, J. Wallach, Reshma Ramachandran","doi":"10.1136/bmjmed-2023-000802","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000802","url":null,"abstract":"Objectives To evaluate National Comprehensive Cancer Network (NCCN) guideline recommendations for oncology drug treatments that have been granted accelerated approval, and to determine whether recommendations are updated based on the results of confirmatory trials after approval and based on status updates from the US Food and Drug Administration (FDA). Design Cross sectional study. Setting US FDA and NCCN guidelines. Population Oncology therapeutic indications (ie, specific oncological conditions for which the drug is recommended) that have been granted accelerated approval in 2009-18. Main outcome measures NCCN guideline reporting of accelerated approval status and postapproval confirmatory trials, and guideline recommendation alignment with postapproval confirmatory trial results and FDA status updates. Results 39 oncology drug treatments were granted accelerated approval for 62 oncological indications. Although all indications were recommended in NCCN guidelines, accelerated approval status was reported for 10 (16%) indications. At least one postapproval confirmatory trial was identified for all 62 indications, 33 (53%) of which confirmed benefit; among these indications, NCCN guidelines maintained the previous recommendation or strengthened the category of evidence for 27 (82%). Postapproval confirmatory trials failed to confirm benefit for 12 (19%) indications; among these indications, NCCN guidelines removed the previous recommendation or weakened the category of evidence for five (42%). NCCN guidelines reflected the FDA's decision to convert 30 (83%) of 36 indications from accelerated to traditional approval, of which 20 (67%) had guideline updates before the FDA's conversion decision. NCCN guidelines reflected the FDA's decision to withdraw seven (58%) of 12 indications from the market, of which four (57%) had guidelines updates before the FDA's withdrawal decision. Conclusions NCCN guidelines always recommend drug treatments that have been granted accelerated approval for oncological indications, but do not provide information about their accelerated approval status, including surrogate endpoint use and status of postapproval confirmatory trials. NCCN guidelines consistently provide information on postapproval trial results confirming clinical benefit, but not on postapproval trials failing to confirm clinical benefit. NCCN guidelines more frequently update recommendation for indications converted to traditional approval than for those approvals that were withdrawn.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"17 9","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-04-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140775096","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-04-01DOI: 10.1136/bmjmed-2023-000783
A. Boulesteix, Sabine Hoffmann
{"title":"To adjust or not to adjust: it is not the tests performed that count, but how they are reported and interpreted","authors":"A. Boulesteix, Sabine Hoffmann","doi":"10.1136/bmjmed-2023-000783","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000783","url":null,"abstract":"","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"1177 1","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-04-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140774078","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-04-01DOI: 10.1136/bmjmed-2023-000807
S. Santhakumaran, Louis Fisher, Bang Zheng, V. Mahalingasivam, Lucy Plumb, E. Parker, R. Steenkamp, Caroline Morton, A. Mehrkar, S. Bacon, Sue Lyon, Rob Konstant-Hambling, B. Goldacre, B. Mackenna, Laurie A. Tomlinson, D. Nitsch
Objective To validate primary and secondary care codes in electronic health records to identify people receiving chronic kidney replacement therapy based on gold standard registry data. Design Validation study using data from OpenSAFELY and the UK Renal Registry, with the approval of NHS England. Setting Primary and secondary care electronic health records from people registered at 45% of general practices in England on 1 January 2020, linked to data from the UK Renal Registry (UKRR) within the OpenSAFELY-TPP platform, part of the NHS England OpenSAFELY covid-19 service. Participants 38 745 prevalent patients (recorded as receiving kidney replacement therapy on 1 January 2020 in UKRR data, or primary or secondary care data) and 10 730 incident patients (starting kidney replacement therapy during 2020), from a population of 19 million people alive and registered with a general practice in England on 1 January 2020. Main outcome measures Sensitivity and positive predictive values of primary and secondary care code lists for identifying prevalent and incident kidney replacement therapy cohorts compared with the gold standard UKRR data on chronic kidney replacement therapy. Agreement across the data sources overall, and by treatment modality (transplantation or dialysis) and personal characteristics. Results Primary and secondary care code lists were sensitive for identifying the UKRR prevalent cohort (91.2% (95% confidence interval (CI) 90.8% to 91.6%) and 92.0% (91.6% to 92.4%), respectively), but not the incident cohort (52.3% (50.3% to 54.3%) and 67.9% (66.1% to 69.7%)). Positive predictive values were low (77.7% (77.2% to 78.2%) for primary care data and 64.7% (64.1% to 65.3%) for secondary care data), particularly for chronic dialysis (53.7% (52.9% to 54.5%) for primary care data and 49.1% (48.0% to 50.2%) for secondary care data). Sensitivity decreased with age and index of multiple deprivation in primary care data, but the opposite was true in secondary care data. Agreement was lower in children, with 30% (295/980) featuring in all three datasets. Half (1165/2315) of the incident patients receiving dialysis in UKRR data had a kidney replacement therapy code in the primary care data within three months of the start date of the kidney replacement therapy. No codes existed whose exclusion would substantially improve the positive predictive value without a decrease in sensitivity. Conclusions Codes used in primary and secondary care data failed to identify a small proportion of prevalent patients receiving kidney replacement therapy. Codes also identified many patients who were not recipients of chronic kidney replacement therapy in UKRR data, particularly dialysis codes. Linkage with UKRR kidney replacement therapy data facilitated more accurate identification of incident and prevalent kidney replacement therapy cohorts for research into this vulnerable population. Poor coding has implications for any patient care (including eligibility for vacc
{"title":"Identification of patients undergoing chronic kidney replacement therapy in primary and secondary care data: validation study based on OpenSAFELY and UK Renal Registry","authors":"S. Santhakumaran, Louis Fisher, Bang Zheng, V. Mahalingasivam, Lucy Plumb, E. Parker, R. Steenkamp, Caroline Morton, A. Mehrkar, S. Bacon, Sue Lyon, Rob Konstant-Hambling, B. Goldacre, B. Mackenna, Laurie A. Tomlinson, D. Nitsch","doi":"10.1136/bmjmed-2023-000807","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000807","url":null,"abstract":"Objective To validate primary and secondary care codes in electronic health records to identify people receiving chronic kidney replacement therapy based on gold standard registry data. Design Validation study using data from OpenSAFELY and the UK Renal Registry, with the approval of NHS England. Setting Primary and secondary care electronic health records from people registered at 45% of general practices in England on 1 January 2020, linked to data from the UK Renal Registry (UKRR) within the OpenSAFELY-TPP platform, part of the NHS England OpenSAFELY covid-19 service. Participants 38 745 prevalent patients (recorded as receiving kidney replacement therapy on 1 January 2020 in UKRR data, or primary or secondary care data) and 10 730 incident patients (starting kidney replacement therapy during 2020), from a population of 19 million people alive and registered with a general practice in England on 1 January 2020. Main outcome measures Sensitivity and positive predictive values of primary and secondary care code lists for identifying prevalent and incident kidney replacement therapy cohorts compared with the gold standard UKRR data on chronic kidney replacement therapy. Agreement across the data sources overall, and by treatment modality (transplantation or dialysis) and personal characteristics. Results Primary and secondary care code lists were sensitive for identifying the UKRR prevalent cohort (91.2% (95% confidence interval (CI) 90.8% to 91.6%) and 92.0% (91.6% to 92.4%), respectively), but not the incident cohort (52.3% (50.3% to 54.3%) and 67.9% (66.1% to 69.7%)). Positive predictive values were low (77.7% (77.2% to 78.2%) for primary care data and 64.7% (64.1% to 65.3%) for secondary care data), particularly for chronic dialysis (53.7% (52.9% to 54.5%) for primary care data and 49.1% (48.0% to 50.2%) for secondary care data). Sensitivity decreased with age and index of multiple deprivation in primary care data, but the opposite was true in secondary care data. Agreement was lower in children, with 30% (295/980) featuring in all three datasets. Half (1165/2315) of the incident patients receiving dialysis in UKRR data had a kidney replacement therapy code in the primary care data within three months of the start date of the kidney replacement therapy. No codes existed whose exclusion would substantially improve the positive predictive value without a decrease in sensitivity. Conclusions Codes used in primary and secondary care data failed to identify a small proportion of prevalent patients receiving kidney replacement therapy. Codes also identified many patients who were not recipients of chronic kidney replacement therapy in UKRR data, particularly dialysis codes. Linkage with UKRR kidney replacement therapy data facilitated more accurate identification of incident and prevalent kidney replacement therapy cohorts for research into this vulnerable population. Poor coding has implications for any patient care (including eligibility for vacc","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"76 17","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-04-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140794531","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-04-01DOI: 10.1136/bmjmed-2024-000896
Saketh Guntupalli
A compelling analysis supports consideration of ADNEX in triage of women with adnexal masses
一项令人信服的分析支持在对患有附件肿块的妇女进行分流时考虑使用 ADNEX
{"title":"Practical implications of the ADNEX risk prediction model for diagnosis of ovarian cancer","authors":"Saketh Guntupalli","doi":"10.1136/bmjmed-2024-000896","DOIUrl":"https://doi.org/10.1136/bmjmed-2024-000896","url":null,"abstract":"A compelling analysis supports consideration of ADNEX in triage of women with adnexal masses","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"1419 ","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-04-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140773965","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-03-06DOI: 10.1136/bmjmed-2023-000771
Shi Zhao, Wangnan Cao, Gengze Liao, Zihao Guo, Lufei Xu, Chen Shen, Chun Chang, Xiao Liang, Kai Wang, Daihai He, Shengzhi Sun, Rui Chen, Francesca Dominici
Objectives To estimate the association between the transition to daylight saving time and the risks of all cause and cause specific mortality in the US. Design Nationwide time series observational study based on weekly data. Setting US state level mortality data from the National Center for Health Statistics, with death counts from 50 US states and the District of Columbia, from the start of 2015 to the end of 2019. Population 13 912 837 reported deaths in the US. Main outcome measures Weekly counts of mortality for any cause, and for Alzheimer's disease, dementia, circulatory diseases, malignant neoplasms, and respiratory diseases. Results During the study period, 13 912 837 deaths were reported. The analysis found no evidence of an association between the transition to spring daylight saving time (when clocks are set forward by one hour on the second Sunday of March) and the risk of all cause mortality during the first eight weeks after the transition (rate ratio 1.003, 95% confidence interval 0.987 to 1.020). Autumn daylight saving time is defined in this study as the time when the clocks are set back by one hour (ie, return to standard time) on the first Sunday of November. Evidence indicating a substantial decrease in the risk of all cause mortality during the first eight weeks after the transition to autumn daylight saving time (0.974, 0.958 to 0.990). Overall, when considering the transition to both spring and autumn daylight saving time, no evidence of any effect of daylight saving time on all cause mortality was found (0.988, 0.972 to 1.005). These patterns of changes in mortality rates associated with transition to daylight saving time were consistent for Alzheimer's disease, dementia, circulatory diseases, malignant neoplasms, and respiratory diseases. The protective effect of the transition to autumn daylight saving time on the risk of mortality was more pronounced in elderly people aged ≥75 years, in the non-Hispanic white population, and in those residing in the eastern time zone. Conclusions In this study, transition to daylight saving time was found to affect mortality patterns in the US, but an association with additional deaths overall was not found. These findings might inform the ongoing debate on the policy of shifting daylight saving time.
{"title":"All cause and cause specific mortality associated with transition to daylight saving time in US: nationwide, time series, observational study","authors":"Shi Zhao, Wangnan Cao, Gengze Liao, Zihao Guo, Lufei Xu, Chen Shen, Chun Chang, Xiao Liang, Kai Wang, Daihai He, Shengzhi Sun, Rui Chen, Francesca Dominici","doi":"10.1136/bmjmed-2023-000771","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000771","url":null,"abstract":"Objectives To estimate the association between the transition to daylight saving time and the risks of all cause and cause specific mortality in the US. Design Nationwide time series observational study based on weekly data. Setting US state level mortality data from the National Center for Health Statistics, with death counts from 50 US states and the District of Columbia, from the start of 2015 to the end of 2019. Population 13 912 837 reported deaths in the US. Main outcome measures Weekly counts of mortality for any cause, and for Alzheimer's disease, dementia, circulatory diseases, malignant neoplasms, and respiratory diseases. Results During the study period, 13 912 837 deaths were reported. The analysis found no evidence of an association between the transition to spring daylight saving time (when clocks are set forward by one hour on the second Sunday of March) and the risk of all cause mortality during the first eight weeks after the transition (rate ratio 1.003, 95% confidence interval 0.987 to 1.020). Autumn daylight saving time is defined in this study as the time when the clocks are set back by one hour (ie, return to standard time) on the first Sunday of November. Evidence indicating a substantial decrease in the risk of all cause mortality during the first eight weeks after the transition to autumn daylight saving time (0.974, 0.958 to 0.990). Overall, when considering the transition to both spring and autumn daylight saving time, no evidence of any effect of daylight saving time on all cause mortality was found (0.988, 0.972 to 1.005). These patterns of changes in mortality rates associated with transition to daylight saving time were consistent for Alzheimer's disease, dementia, circulatory diseases, malignant neoplasms, and respiratory diseases. The protective effect of the transition to autumn daylight saving time on the risk of mortality was more pronounced in elderly people aged ≥75 years, in the non-Hispanic white population, and in those residing in the eastern time zone. Conclusions In this study, transition to daylight saving time was found to affect mortality patterns in the US, but an association with additional deaths overall was not found. These findings might inform the ongoing debate on the policy of shifting daylight saving time.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"15 4","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-03-06","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140077742","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-03-01DOI: 10.1136/bmjmed-2023-000624
C. Razieh, S. Shabnam, H. Dambha‐Miller, Eva J A Morris, T. Yates, Y. Chudasama, F. Zaccardi, C. Gillies, Amitava Banerjee, Manish Pareek, Ben Lacey, Martin White, K. Khunti, N. Islam
To explore the risk of a positive test result for the delta or omicron variant of the SARS-CoV-2 virus in different occupations and deprivation groups in the UK.Analysis of the longitudinal COVID-19 Infection Survey.COVID-19 Infection Survey, conducted by the Office for National Statistics and the University of Oxford, UK, a nationwide longitudinal survey to monitor SARS-CoV-2 infection in the community, 26 April 2020 to 31 January 2022.Survey participants recruited from randomly selected households to reflect the UK population (England, Scotland, Wales, and Northern Ireland) were divided into the delta cohort (2 July 2020 to 19 December 2021) and the omicron variant (on or after 20 December 2021), the dominant variants during our study period.Incidence rate and incidence rate ratio for the presence of the delta and omicron variants by area level deprivation and occupation sector. Multivariable Poisson regression models were fitted to estimate the adjusted incidence rate ratio after adjusting for age, sex, ethnic group, comorbid conditions, urban or rural residence, household size, patient or client facing job, and time (as quarters of the year).329 356 participants were included in the delta cohort and 246 061 in the omicron cohort. The crude incidence rate for the presence of the delta and omicron variants of the SARS-CoV-2 virus were higher in the most deprived group (based on the index of multiple deprivation divided by deciles; delta cohort 4.33 per 1000 person months, 95% confidence interval 4.09 to 4.58; omicron cohort 76.67 per 1000 person months, 71.60 to 82.11) than in the least deprived group (3.18, 3.05 to 3.31 and 54.52, 51.93 to 57.24, respectively); the corresponding adjusted incidence rate ratios were 1.37 (95% confidence interval 1.29 to 1.47) and 1.34 (1.24 to 1.46) during the delta and omicron variant dominant periods, respectively. The adjusted incidence rate ratios for a positive test result in the most deprived group compared with the least deprived group in the delta cohort were 1.59 (95% confidence interval 1.25 to 2.02) and 1.50 (1.19 to 1.87) in the healthcare and manufacturing or construction sectors, respectively. Corresponding values in the omicron cohort were 1.50 (1.15 to 1.95) and 1.43 (1.09 to 1.86) in the healthcare and teaching and education sectors, respectively. Associations between SARS-CoV-2 infection and other employment sectors were not significant or were not tested because of small numbers.In this study, the risk of a positive test result for the SARS-CoV-2 virus in the delta and omicron cohorts was higher in the most deprived than in the least deprived group in the healthcare, manufacturing or construction, and teaching and education sectors.
{"title":"Socioeconomic inequalities in risk of infection with SARS-CoV-2 delta and omicron variants in the UK, 2020-22: analysis of the longitudinal COVID-19 Infection Survey","authors":"C. Razieh, S. Shabnam, H. Dambha‐Miller, Eva J A Morris, T. Yates, Y. Chudasama, F. Zaccardi, C. Gillies, Amitava Banerjee, Manish Pareek, Ben Lacey, Martin White, K. Khunti, N. Islam","doi":"10.1136/bmjmed-2023-000624","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000624","url":null,"abstract":"To explore the risk of a positive test result for the delta or omicron variant of the SARS-CoV-2 virus in different occupations and deprivation groups in the UK.Analysis of the longitudinal COVID-19 Infection Survey.COVID-19 Infection Survey, conducted by the Office for National Statistics and the University of Oxford, UK, a nationwide longitudinal survey to monitor SARS-CoV-2 infection in the community, 26 April 2020 to 31 January 2022.Survey participants recruited from randomly selected households to reflect the UK population (England, Scotland, Wales, and Northern Ireland) were divided into the delta cohort (2 July 2020 to 19 December 2021) and the omicron variant (on or after 20 December 2021), the dominant variants during our study period.Incidence rate and incidence rate ratio for the presence of the delta and omicron variants by area level deprivation and occupation sector. Multivariable Poisson regression models were fitted to estimate the adjusted incidence rate ratio after adjusting for age, sex, ethnic group, comorbid conditions, urban or rural residence, household size, patient or client facing job, and time (as quarters of the year).329 356 participants were included in the delta cohort and 246 061 in the omicron cohort. The crude incidence rate for the presence of the delta and omicron variants of the SARS-CoV-2 virus were higher in the most deprived group (based on the index of multiple deprivation divided by deciles; delta cohort 4.33 per 1000 person months, 95% confidence interval 4.09 to 4.58; omicron cohort 76.67 per 1000 person months, 71.60 to 82.11) than in the least deprived group (3.18, 3.05 to 3.31 and 54.52, 51.93 to 57.24, respectively); the corresponding adjusted incidence rate ratios were 1.37 (95% confidence interval 1.29 to 1.47) and 1.34 (1.24 to 1.46) during the delta and omicron variant dominant periods, respectively. The adjusted incidence rate ratios for a positive test result in the most deprived group compared with the least deprived group in the delta cohort were 1.59 (95% confidence interval 1.25 to 2.02) and 1.50 (1.19 to 1.87) in the healthcare and manufacturing or construction sectors, respectively. Corresponding values in the omicron cohort were 1.50 (1.15 to 1.95) and 1.43 (1.09 to 1.86) in the healthcare and teaching and education sectors, respectively. Associations between SARS-CoV-2 infection and other employment sectors were not significant or were not tested because of small numbers.In this study, the risk of a positive test result for the SARS-CoV-2 virus in the delta and omicron cohorts was higher in the most deprived than in the least deprived group in the healthcare, manufacturing or construction, and teaching and education sectors.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"74 3","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-03-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140280069","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-03-01DOI: 10.1136/bmjmed-2023-000729
L. Goodfellow, Angharad Care, Ciara Curran, Devender Roberts, Mark A Turner, Marian Knight, Alfirevic Zarko
To describe perinatal and maternal outcomes of preterm prelabour rupture of membranes (PPROM) before 23 weeks' gestation in a national cohort.Prospective observational study.National population based cohort study with the UK Obstetric Surveillance System (UKOSS), a research infrastructure of all 194 obstetric units in the UK, 1 September 2019 to 28 February 2021.326 women with singleton and 38 with multiple pregnancies with PPROM between 16+0 and 22+6 weeks+days' gestation.Perinatal outcomes of live birth, survival to discharge from hospital, and severe morbidity, defined as intraventricular haemorrhage grade 3 or 4, or requiring supplemental oxygen at 36 weeks' postmenstrual age, or both. Maternal outcomes were surgery for removal of the placenta, sepsis, admission to an intensive treatment unit, and death. Clinical data included rates of termination of pregnancy for medical reasons.Perinatal outcomes were calculated with all terminations of pregnancy for medical reasons excluded, and a worst-best range was calculated assuming that all terminations for medical reasons and those with missing data would have died (minimum value) or all would be liveborn (maximum value). For singleton pregnancies, the live birth rate was 44% (98/223), range 30-62% (98/326-201/326), perinatal survival to discharge from hospital was 26% (54/207), range 17-53% (54/326-173/326), and 18% (38/207), range 12-48% (38/326-157/326) of babies survived without severe morbidity. The rate of maternal sepsis was 12% (39/326) in singleton and 29% (11/38) in multiple pregnancies (P=0.004). Surgery for removal of the placenta was needed in 20% (65/326) and 16% (6/38) of singleton and twin pregnancies, respectively. Five women became severely unwell with sepsis; two died and another three required care in the intensive treatment unit.In this study, 26% of women who had very early PPROM with expectant management had babies that survived to discharge from hospital. Morbidity and mortality rates were high for both mothers and neonates. Maternal sepsis is a considerable risk that needs more research. These data should be used in counselling families with PPROM before 23 weeks' gestation, and currently available guidelines should be updated accordingly.
{"title":"Preterm prelabour rupture of membranes before 23 weeks’ gestation: prospective observational study","authors":"L. Goodfellow, Angharad Care, Ciara Curran, Devender Roberts, Mark A Turner, Marian Knight, Alfirevic Zarko","doi":"10.1136/bmjmed-2023-000729","DOIUrl":"https://doi.org/10.1136/bmjmed-2023-000729","url":null,"abstract":"To describe perinatal and maternal outcomes of preterm prelabour rupture of membranes (PPROM) before 23 weeks' gestation in a national cohort.Prospective observational study.National population based cohort study with the UK Obstetric Surveillance System (UKOSS), a research infrastructure of all 194 obstetric units in the UK, 1 September 2019 to 28 February 2021.326 women with singleton and 38 with multiple pregnancies with PPROM between 16+0 and 22+6 weeks+days' gestation.Perinatal outcomes of live birth, survival to discharge from hospital, and severe morbidity, defined as intraventricular haemorrhage grade 3 or 4, or requiring supplemental oxygen at 36 weeks' postmenstrual age, or both. Maternal outcomes were surgery for removal of the placenta, sepsis, admission to an intensive treatment unit, and death. Clinical data included rates of termination of pregnancy for medical reasons.Perinatal outcomes were calculated with all terminations of pregnancy for medical reasons excluded, and a worst-best range was calculated assuming that all terminations for medical reasons and those with missing data would have died (minimum value) or all would be liveborn (maximum value). For singleton pregnancies, the live birth rate was 44% (98/223), range 30-62% (98/326-201/326), perinatal survival to discharge from hospital was 26% (54/207), range 17-53% (54/326-173/326), and 18% (38/207), range 12-48% (38/326-157/326) of babies survived without severe morbidity. The rate of maternal sepsis was 12% (39/326) in singleton and 29% (11/38) in multiple pregnancies (P=0.004). Surgery for removal of the placenta was needed in 20% (65/326) and 16% (6/38) of singleton and twin pregnancies, respectively. Five women became severely unwell with sepsis; two died and another three required care in the intensive treatment unit.In this study, 26% of women who had very early PPROM with expectant management had babies that survived to discharge from hospital. Morbidity and mortality rates were high for both mothers and neonates. Maternal sepsis is a considerable risk that needs more research. These data should be used in counselling families with PPROM before 23 weeks' gestation, and currently available guidelines should be updated accordingly.","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"61 ","pages":""},"PeriodicalIF":0.0,"publicationDate":"2024-03-01","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140274799","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
Pub Date : 2024-02-28eCollection Date: 2024-01-01DOI: 10.1136/bmjmed-2024-000855
Anna Head, Martin O'Flaherty, Chris Kypridemos
{"title":"Multimorbidity research: where one size does not fit all.","authors":"Anna Head, Martin O'Flaherty, Chris Kypridemos","doi":"10.1136/bmjmed-2024-000855","DOIUrl":"10.1136/bmjmed-2024-000855","url":null,"abstract":"","PeriodicalId":72433,"journal":{"name":"BMJ medicine","volume":"3 1","pages":"e000855"},"PeriodicalIF":0.0,"publicationDate":"2024-02-28","publicationTypes":"Journal Article","fieldsOfStudy":null,"isOpenAccess":false,"openAccessPdf":"https://www.ncbi.nlm.nih.gov/pmc/articles/PMC10910389/pdf/","citationCount":null,"resultStr":null,"platform":"Semanticscholar","paperid":"140029671","PeriodicalName":null,"FirstCategoryId":null,"ListUrlMain":null,"RegionNum":0,"RegionCategory":"","ArticlePicture":[],"TitleCN":null,"AbstractTextCN":null,"PMCID":"OA","EPubDate":null,"PubModel":null,"JCR":null,"JCRName":null,"Score":null,"Total":0}
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