Summary Background Sexual behaviour and relationships are key components of wellbeing and are affected by social norms, attitudes, and health. We present data on sexual behaviours and attitudes in ...Britain (England, Scotland, and Wales) from the three National Surveys of Sexual Attitudes and Lifestyles (Natsal). Methods We used a multistage, clustered, and stratified probability sample design. Within each of the 1727 sampled postcode sectors for Natsal-3, 30 or 36 addresses were randomly selected and then assigned to interviewers. To oversample individuals aged 16–34 years, we randomly allocated addresses to either the core sample (in which individuals aged 16–74 years were eligible) or the boost sample (in which only individuals aged 16–34 years were eligible). Interviewers visited all sampled addresses between Sept 6, 2010, and Aug 31, 2012, and randomly selected one eligible individual from each household to be invited to participate. Participants completed the survey in their own homes through computer-assisted face-to-face interviews and self-interview. We analysed data from this survey, weighted to account for unequal selection probabilities and non-response to correct for differences in sex, age group, and region according to 2011 Census figures. We then compared data from participants aged 16–44 years from Natsal-1 (1990–91), Natsal-2 (1999–2001), and Natsal-3. Findings Interviews were completed with 15 162 participants (6293 men, 8869 women) from 26 274 eligible addresses (57·7%). 82·1% (95% CI 81·0–83·1%) of men and 77·7% (76·7–78·7%) of women reported at least one sexual partner of the opposite sex in the past year. The proportion generally decreased with age, as did the range of sexual practices with partners of the opposite sex, especially in women. The increased sexual activity and diversity reported in Natsal-2 in individuals aged 16–44 years when compared with Natsal-1 has generally been sustained in Natsal-3, but in men has generally not risen further. However, in women, the number of male sexual partners over the lifetime (age-adjusted odds ratio 1·18, 95% CI 1·08–1·28), proportion reporting ever having had a sexual experience with genital contact with another woman (1·69, 1·43–2·00), and proportion reporting at least one female sexual partner in the past 5 years (2·00, 1·59–2·51) increased in Natsal-3 compared with Natsal-2. While reported number of occasions of heterosexual intercourse in the past 4 weeks had reduced since Natsal-2, we recorded an expansion of heterosexual repertoires—particularly in oral and anal sex—over time. Acceptance of same-sex partnerships and intolerance of non-exclusivity in marriage increased in men and women in Natsal-3. Interpretation Sexual lifestyles in Britain have changed substantially in the past 60 years, with changes in behaviour seeming greater in women than men. The continuation of sexual activity into later life—albeit reduced in range and frequency—emphasises that attention to sexual health and wellbeing is needed throughout the life course. Funding Grants from the UK Medical Research Council and the Wellcome Trust, with support from the Economic and Social Research Council and the Department of Health.
Clinical researchers have often preferred to use a fixed effects model for the primary interpretation of a meta-analysis. Heterogeneity is usually assessed via the well known Q and I2 statistics, ...along with the random effects estimate they imply. In recent years, alternative methods for quantifying heterogeneity have been proposed, that are based on a 'generalised' Q statistic.
We review 18 IPD meta-analyses of RCTs into treatments for cancer, in order to quantify the amount of heterogeneity present and also to discuss practical methods for explaining heterogeneity.
Differing results were obtained when the standard Q and I2 statistics were used to test for the presence of heterogeneity. The two meta-analyses with the largest amount of heterogeneity were investigated further, and on inspection the straightforward application of a random effects model was not deemed appropriate. Compared to the standard Q statistic, the generalised Q statistic provided a more accurate platform for estimating the amount of heterogeneity in the 18 meta-analyses.
Explaining heterogeneity via the pre-specification of trial subgroups, graphical diagnostic tools and sensitivity analyses produced a more desirable outcome than an automatic application of the random effects model. Generalised Q statistic methods for quantifying and adjusting for heterogeneity should be incorporated as standard into statistical software. Software is provided to help achieve this aim.
Summary Background Despite its importance to sexual health and wellbeing, sexual function is given little attention in sexual health policy. Population-based studies are needed to understand sexual ...function across the life course. Methods We undertook a probability sample survey (the third National Survey of Sexual Attitudes and Lifestyles Natsal-3) of 15 162 individuals aged 16–74 years who lived in Britain (England, Scotland, and Wales). Interviews were done between Sept 6, 2010, and Aug 31, 2012. We assessed the distribution of sexual function by use of a novel validated measure (the Natsal-SF), which assessed problems with individual sexual response, sexual function in a relationship context, and self-appraisal of sex life (17 items; 16 items per gender). We assess factors associated with low sexual function (defined as the lowest quintile of distribution of Natsal-SF scores) and the distribution of components of the measure. Participants reporting one or more sexual partner in the past year were given a score on the Natsal-SF (11 690 participants). 4122 of these participants were not in a relationship for all of the past year and we employed the full information maximum likelihood method to handle missing data on four relationship items. Findings We obtained data for 4913 men and 6777 women for the Natsal-SF. For men and women, low sexual function was associated with increased age, and, after age-adjustment, with depression (adjusted odds ratio 3·70 95% CI 2·90–4·72 for men and 4·11 3·36–5·04 for women) and self-reported poor health status (2·63 1·73–3·98 and 2·41 1·72–3·39). Low sexual function was also associated with experiencing the end of a relationship (1·52 1·18–1·95 and 1·77 1·44–2·17), inability to talk easily about sex with a partner (2·36 1·94–2·88 and 2·82 2·28–3·48), and not being happy in the relationship (2·89 2·32–3·61 and 4·10 3·39–4·97). Associations were also noted with engaging in fewer than four sex acts in the past 4 weeks (3·13 2·58–3·79 and 3·38 2·80–4·09), having had same sex partners (2·28 1·56–3·35 and 1·60 1·16–2·20), paying for sex (in men only; 2·62 1·46–4·71), and higher numbers of lifetime sexual partners (in women only; 2·12 1·68–2·67 for ten or more partners). Low sexual function was also associated with negative sexual health outcomes such as experience of non-volitional sex (1·98 1·14–3·43 and 2·18 1·79–2·66) and STI diagnosis (1·50 1·06–2·11 and 1·83 1·35–2·47). Among individuals reporting sex in the past year, problems with sexual response were common (41·6% of men and 51·2% of women reported one or more problem) but self-reported distress about sex lives was much less common (9·9% and 10·9%). For individuals in a sexual relationship for the past year, 23·4% of men and 27·4% of women reported an imbalance in level of interest in sex between partners, and 18·0% of men and 17·1% of women said that their partner had had sexual difficulties. Most participants who did not have sex in the past year were not dissatisfied, distressed, or avoiding sex because of sexual difficulties. Interpretation Wide variability exists in the distribution of sexual function scores. Low sexual function is associated with negative sexual health outcomes, supporting calls for a greater emphasis on sexual function in sexual health policy and interventions. Funding Grants from the UK Medical Research Council and the Wellcome Trust, with support from the Economic and Social Research Council and the Department of Health.
Summary Background Unplanned pregnancy is a key public health indicator. We describe the prevalence of unplanned pregnancy, and associated factors, in a general population sample in Britain (England, ...Scotland, and Wales). Method We did a probability sample survey, the third National Survey of Sexual Attitudes and Lifestyles (Natsal-3), of 15 162 men and women aged 16–74 years in Britain, including 5686 women of child-bearing age (16–44 years) who were included in the pregnancy analysis, between Sept 6, 2010, and Aug 31, 2012. We describe the planning status of pregnancies with known outcomes in the past year, and report the annual population prevalence of unplanned pregnancy, using a validated, multicriteria, multi-outcome measure (the London Measure of Unplanned Pregnancy). We set the findings in the context of secular trends in reproductive health-related events, and patterns across the life course. Findings 9·7% of women aged 16–44 years had pregnancies with known outcome in the year before interview, of which 16·2% (95% CI 13·1–19·9) scored as unplanned, 29·0% (25·2–33·2) as ambivalent, and 54·8% (50·3–59·2) as planned, giving an annual prevalence estimate for unplanned pregnancy of 1·5% (1·2–1·9). Pregnancies in women aged 16–19 years were most commonly unplanned (45·2% 30·8–60·5). However, most unplanned pregnancies were in women aged 20–34 years (62·4% 50·2–73·2). Factors strongly associated with unplanned pregnancy were first sexual intercourse before 16 years of age (age-adjusted odds ratio 2·85 95% CI 1·77–4·57, current smoking (2·47 1·46–4·18), recent use of drugs other than cannabis (3·41 1·64–7·11), and lower educational attainment. Unplanned pregnancy was also associated with lack of sexual competence at first sexual intercourse (1·90 1·14–3·08), reporting higher frequency of sex (2·11 1·25–3·57 for five or more times in the past 4 weeks), receiving sex education mainly from a non-school-based source (1·84 1·12–3·00), and current depression (1·96 1·10–3·47). Interpretation The increasing intervals between first sexual intercourse, cohabitation, and childbearing means that, on average, women in Britain spend about 30 years of their life needing to avert an unplanned pregnancy. Our data offer scope for primary prevention aimed at reducing the rate of unplanned conceptions, and secondary prevention aimed at modification of health behaviours and health disorders in unplanned pregnancy that might be harmful for mother and child. Funding Grants from the UK Medical Research Council and the Wellcome Trust, with support from the Economic and Social Research Council and the Department of Health.
Cluster-randomized trials (CRTs) involve randomizing groups of individuals (e.g. hospitals, schools or villages) to different interventions. Various approaches exist for analysing CRTs but there has ...been little discussion around the treatment effects (estimands) targeted by each.
We describe the different estimands that can be addressed through CRTs and demonstrate how choices between different analytic approaches can impact the interpretation of results by fundamentally changing the question being asked, or, equivalently, the target estimand.
CRTs can address either the participant-average treatment effect (the average treatment effect across participants) or the cluster-average treatment effect (the average treatment effect across clusters). These two estimands can differ when participant outcomes or the treatment effect depends on the cluster size (referred to as 'informative cluster size'), which can occur for reasons such as differences in staffing levels or types of participants between small and large clusters. Furthermore, common estimators, such as mixed-effects models or generalized estimating equations with an exchangeable working correlation structure, can produce biased estimates for both the participant-average and cluster-average treatment effects when cluster size is informative. We describe alternative estimators (independence estimating equations and cluster-level analyses) that are unbiased for CRTs even when informative cluster size is present.
We conclude that careful specification of the estimand at the outset can ensure that the study question being addressed is clear and relevant, and, in turn, that the selected estimator provides an unbiased estimate of the desired quantity.
The effectiveness of SARS-CoV-2 vaccines in older adults living in long-term care facilities is uncertain. We investigated the protective effect of the first dose of the Oxford-AstraZeneca ...non-replicating viral-vectored vaccine (ChAdOx1 nCoV-19; AZD1222) and the Pfizer-BioNTech mRNA-based vaccine (BNT162b2) in residents of long-term care facilities in terms of PCR-confirmed SARS-CoV-2 infection over time since vaccination.
The VIVALDI study is a prospective cohort study that commenced recruitment on June 11, 2020, to investigate SARS-CoV-2 transmission, infection outcomes, and immunity in residents and staff in long-term care facilities in England that provide residential or nursing care for adults aged 65 years and older. In this cohort study, we included long-term care facility residents undergoing routine asymptomatic SARS-CoV-2 testing between Dec 8, 2020 (the date the vaccine was first deployed in a long-term care facility), and March 15, 2021, using national testing data linked within the COVID-19 Datastore. Using Cox proportional hazards regression, we estimated the relative hazard of PCR-positive infection at 0–6 days, 7–13 days, 14–20 days, 21–27 days, 28–34 days, 35–48 days, and 49 days and beyond after vaccination, comparing unvaccinated and vaccinated person-time from the same cohort of residents, adjusting for age, sex, previous infection, local SARS-CoV-2 incidence, long-term care facility bed capacity, and clustering by long-term care facility. We also compared mean PCR cycle threshold (Ct) values for positive swabs obtained before and after vaccination. The study is registered with ISRCTN, number 14447421.
10 412 care home residents aged 65 years and older from 310 LTCFs were included in this analysis. The median participant age was 86 years (IQR 80–91), 7247 (69·6%) of 10 412 residents were female, and 1155 residents (11·1%) had evidence of previous SARS-CoV-2 infection. 9160 (88·0%) residents received at least one vaccine dose, of whom 6138 (67·0%) received ChAdOx1 and 3022 (33·0%) received BNT162b2. Between Dec 8, 2020, and March 15, 2021, there were 36 352 PCR results in 670 628 person-days, and 1335 PCR-positive infections (713 in unvaccinated residents and 612 in vaccinated residents) were included. Adjusted hazard ratios (HRs) for PCR-positive infection relative to unvaccinated residents declined from 28 days after the first vaccine dose to 0·44 (95% CI 0·24–0·81) at 28–34 days and 0·38 (0·19–0·77) at 35–48 days. Similar effect sizes were seen for ChAdOx1 (adjusted HR 0·32, 95% CI 0·15–0·66) and BNT162b2 (0·35, 0·17–0·71) vaccines at 35–48 days. Mean PCR Ct values were higher for infections that occurred at least 28 days after vaccination than for those occurring before vaccination (31·3 SD 8·7 in 107 PCR-positive tests vs 26·6 6·6 in 552 PCR-positive tests; p<0·0001).
Single-dose vaccination with BNT162b2 and ChAdOx1 vaccines provides substantial protection against infection in older adults from 4–7 weeks after vaccination and might reduce SARS-CoV-2 transmission. However, the risk of infection is not eliminated, highlighting the ongoing need for non-pharmaceutical interventions to prevent transmission in long-term care facilities.
UK Government Department of Health and Social Care.
There are sometimes cost, scientific, or logistical reasons to allocate individuals unequally in an individually randomized trial. In cluster randomized trials we can allocate clusters unequally ...and/or allow different cluster size between trial arms. We consider parallel group designs with a continuous outcome, and optimal designs that require the smallest number of individuals to be measured given the number of clusters. Previous authors have derived the optimal allocation ratio for clusters under different variance and/or intracluster correlations (ICCs) between arms, allowing different but prespecified cluster sizes by arm. We derive closed‐form expressions to identify the optimal proportions of clusters and of individuals measured for each arm, thereby defining optimal cluster sizes, when cluster size can be chosen freely. When ICCs differ between arms but the variance is equal, the optimal design allocates more than half the clusters to the arm with the higher ICC, but (typically only slightly) less than half the individuals and hence a smaller cluster size. We also describe optimal design under constraints on the number of clusters or cluster size in one or both arms. This methodology allows trialists to consider a range for the number of clusters in the trial and for each to identify the optimal design. Except if there is clear prior evidence for the ICC and variance by arm, a range of values will need to be considered. Researchers should choose a design with adequate power across the range, while also keeping enough clusters in each arm to permit the intended analysis method.
Summary Background Population-based estimates of prevalence, risk distribution, and intervention uptake inform delivery of control programmes for sexually transmitted infections (STIs). We undertook ...the third National Survey of Sexual Attitudes and Lifestyles (Natsal-3) after implementation of national sexual health strategies, and describe the epidemiology of four STIs in Britain (England, Scotland, and Wales) and the uptake of interventions. Methods Between Sept 6, 2010 and Aug 31, 2012, we did a probability sample survey of 15 162 women and men aged 16–74 years in Britain. Participants were interviewed with computer-assisted face-to-face and self-completion questionnaires. Urine from a sample of participants aged 16–44 years who reported at least one sexual partner over the lifetime was tested for the presence of Chlamydia trachomatis , type-specific human papillomavirus (HPV), Neisseria gonorrhoeae , and HIV antibody. We describe age-specific and sex-specific prevalences of infection and intervention uptake, in relation to demographic and behavioural factors, and explore changes since Natsal-1 (1990–91) and Natsal-2 (1999–2001). Findings Of 8047 eligible participants invited to provide a urine sample, 4828 (60%) agreed. We excluded 278 samples, leaving 4550 (94%) participants with STI test results. Chlamydia prevalence was 1·5% (95% CI 1·1–2·0) in women and 1·1% (0·7–1·6) in men. Prevalences in individuals aged 16–24 years were 3·1% (2·2–4·3) in women and 2·3% (1·5–3·4) in men. Area-level deprivation and higher numbers of partners, especially without use of condoms, were risk factors. However, 60·4% (45·5–73·7) of chlamydia in women and 43·3% (25·9–62·5) in men was in individuals who had had one partner in the past year. Among sexually active 16–24-year-olds, 54·2% (51·4–56·9) of women and 34·6% (31·8–37·4) of men reported testing for chlamydia in the past year, with testing higher in those with more partners. High-risk HPV was detected in 15·9% (14·4–17·5) of women, similar to in Natsal-2. Coverage of HPV catch-up vaccination was 61·5% (58·2–64·7). Prevalence of HPV types 16 and 18 in women aged 18–20 years was lower in Natsal-3 than Natsal-2 (5·8% 3·9–8·6 vs 11·3% 6·8–18·2; age-adjusted odds ratio 0·44 0·21–0·94). Gonorrhoea (<0·1% prevalence in women and men) and HIV (0·1% prevalence in women and 0·2% in men) were uncommon and restricted to participants with recognised high-risk factors. Since Natsal-2, substantial increases were noted in attendance at sexual health clinics (from 6·7% to 21·4% in women and from 7·7% to 19·6% in men) and HIV testing (from 8·7% to 27·6% in women and from 9·2% to 16·9% in men) in the past 5 years. Interpretation STIs were distributed heterogeneously, requiring general and infection-specific interventions. Increases in testing and attendance at sexual health clinics, especially in people at highest risk, are encouraging. However, STIs persist both in individuals accessing and those not accessing services. Our findings provide empirical evidence to inform future sexual health interventions and services. Funding Grants from the UK Medical Research Council and the Wellcome Trust, with support from the Economic and Social Research Council and the Department of Health.
Summary Background Maternal and neonatal mortality rates remain high in many low-income and middle-income countries. Different approaches for the improvement of birth outcomes have been used in ...community-based interventions, with heterogeneous effects on survival. We assessed the effects of women's groups practising participatory learning and action, compared with usual care, on birth outcomes in low-resource settings. Methods We did a systematic review and meta-analysis of randomised controlled trials undertaken in Bangladesh, India, Malawi, and Nepal in which the effects of women's groups practising participatory learning and action were assessed to identify population-level predictors of effect on maternal mortality, neonatal mortality, and stillbirths. We also reviewed the cost-effectiveness of the women's group intervention and estimated its potential effect at scale in Countdown countries. Findings Seven trials (119 428 births) met the inclusion criteria. Meta-analyses of all trials showed that exposure to women's groups was associated with a 23% non-significant reduction in maternal mortality (odds ratio 0·77, 95% CI 0·48–1·23), a 20% reduction in neonatal mortality (0·80, 0·67–0·96), and a 7% non-significant reduction in stillbirths (0·93, 0·82–1·05), with high heterogeneity for maternal ( I2 =64·0%, p=0·011) and neonatal results ( I2 =73·2%, p=0·001). In the meta-regression analyses, the proportion of pregnant women in groups was linearly associated with reduction in both maternal and neonatal mortality (p=0·019 and p=0·009, respectively). A subgroup analysis of the four studies in which at least 30% of pregnant women participated in groups showed a 49% reduction in maternal mortality (0·51, 0·29–0·89) and a 33% reduction in neonatal mortality (0·67, 0·60–0·75). The intervention was cost effective by WHO standards and could save an estimated 283 000 newborn infants and 36 600 mothers per year if implemented in rural areas of 74 Countdown countries. Interpretation With the participation of at least a third of pregnant women and adequate population coverage, women's groups practising participatory learning and action are a cost-effective strategy to improve maternal and neonatal survival in low-resource settings. Funding Wellcome Trust, Ammalife, and National Institute for Health Research Collaboration for Leadership in Applied Health Research and Care for Birmingham and the Black Country programme.
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