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AbstractObjectiveTo use the relation between cigarette consumption and cardiovascular disease to quantify the risk of coronary heart disease and stroke for light smoking (one to five cigarettes/day).DesignSystematic review and meta-analysis.Data sourcesMedline 1946 to May 2015, with manual searches of references.Eligibility criteria for selecting studiesProspective cohort studies with at least 50 events, reporting hazard ratios or relative risks (both hereafter referred to as relative risk) compared with never smokers or age specific incidence in relation to risk of coronary heart disease or stroke.Data extraction/synthesisMOOSE guidelines were followed. For each study, the relative risk was estimated for smoking one, five, or 20 cigarettes per day by using regression modelling between risk and cigarette consumption. Relative risks were adjusted for at least age and often additional confounders. The main measure was the excess relative risk for smoking one cigarette per day (RR1_per_day−1) expressed as a proportion of that for smoking 20 cigarettes per day (RR20_per_day−1), expected to be about 5% assuming a linear relation between risk and consumption (as seen with lung cancer). The relative risks for one, five, and 20 cigarettes per day were also pooled across all studies in a random effects meta-analysis. Separate analyses were done for each combination of sex and disorder.ResultsThe meta-analysis included 55 publications containing 141 cohort studies. Among men, the pooled relative risk for coronary heart disease was 1.48 for smoking one cigarette per day and 2.04 for 20 cigarettes per day, using all studies, but 1.74 and 2.27 among studies in which the relative risk had been adjusted for multiple confounders. Among women, the pooled relative risks were 1.57 and 2.84 for one and 20 cigarettes per day (or 2.19 and 3.95 using relative risks adjusted for multiple factors). Men who smoked one cigarette per day had 46% of the excess relative risk for smoking 20 cigarettes per day (53% using relative risks adjusted for multiple factors), and women had 31% of the excess risk (38% using relative risks adjusted for multiple factors). For stroke, the pooled relative risks for men were 1.25 and 1.64 for smoking one or 20 cigarettes per day (1.30 and 1.56 using relative risks adjusted for multiple factors). In women, the pooled relative risks were 1.31 and 2.16 for smoking one or 20 cigarettes per day (1.46 and 2.42 using relative risks adjusted for multiple factors). The excess risk for stroke associated with one cigarette per day (in relation to 20 cigarettes per day) was 41% for men and 34% for women (or 64% and 36% using relative risks adjusted for multiple factors). Relative risks were generally higher among women than men.ConclusionsSmoking only about one cigarette per day carries a risk of developing coronary heart disease and stroke much greater than expected: around half that for people who smoke 20 per day. No safe level of smoking exists for cardiovascular disease. Smokers should aim to quit instead of cutting down to significantly reduce their risk of these two common major disorders.

SummaryBackgroundCancer services worldwide had to adapt in response to the COVID-19 pandemic to minimise risk to patients and staff. We aimed to assess the national impact of COVID-19 on the prescribing of systemic anticancer treatment in England, immediately after lockdown and after the introduction of new treatments to reduce patient risk. MethodsWe did a retrospective analysis using data from a central National Health Service England web database mandated for clinicians to register intention to start all new systemic anticancer treatments approved for use in England since 2016. We analysed the monthly number of treatment registrations in April, 2020, after the implementation of societal lockdown on March 23, 2020, and after implementation of treatment options to reduce patient risk such as oral or less immunosuppressive drugs, in May and June, 2020. We compared the number of registrations in April–June, 2020, with the mean number of registrations and SD during the previous 6 months of unaffected cancer care (September, 2019, to February, 2020). We calculated the percentage change and absolute difference in SD units for the number of registrations overall, by tumour type, and by type and line of therapy. FindingsIn April, 2020, 2969 registrations were recorded, representing 1417 fewer registrations than in the control period (monthly mean 4386; 32% reduction, absolute difference 4·2 SDs, p<0·0001). In May, 2020, total registrations increased to 3950, representing a 10% reduction compared with the control period (absolute difference 1·3 SDs, p<0·0001). In June, 2020, 5022 registrations were recorded, representing a 15% increase compared with the control period (absolute difference 1·9 SDs; p<0·0001]). InterpretationAfter the onset of the COVID-19 pandemic, there was a reduction in systemic anticancer treatment initiation in England. However, following introduction of treatment options to reduce patient risk, registrations began to increase in May, 2020, and reached higher numbers than the pre-pandemic mean in June, 2020, when other clinical and societal risk mitigation factors (such as telephone consultations, facemasks and physical distancing) are likely to have contributed. However, outcomes of providing less treatment or delaying treatment initiation, particularly for advanced cancers and neoadjuvant therapies, require continued assessment. FundingNone.

Guidelines for anal cancer recommend assessment of response at 6–12 weeks after starting treatment. Using data from the ACT II trial, we determined the optimum timepoint to assess clinical tumour response after chemoradiotherapy. The previously reported ACT II trial was a phase 3 randomised trial of patients of any age with newly diagnosed, histologically confirmed, squamous cell carcinoma of the anus without metastatic disease from 59 centres in the UK. We randomly assigned patients (by minimisation) to receive either intravenous mitomycin (one dose of 12 mg/m2 on day 1) or intravenous cisplatin (one dose of 60 mg/m2 on days 1 and 29), with intravenous fluorouracil (one dose of 1000 mg/m2 per day on days 1–4 and 29–32) and radiotherapy (50·4 Gy in 28 daily fractions); and also did a second randomisation after initial therapy to maintenance chemotherapy (fluorouracil and cisplatin) or no maintenance chemotherapy. The primary outcome was complete clinical response (the absence of primary and nodal tumour by clinical examination), in addition to overall survival and progression-free survival from time of randomisation. In this post-hoc analysis, we analysed complete clinical response at three timepoints: 11 weeks from the start of chemoradiotherapy (assessment 1), 18 weeks from the start of chemoradiotherapy (assessment 2), and 26 weeks from the start of chemoradiotherapy (assessment 3) as well as the overall and progression-free survival estimates of patients with complete clinical response or without complete clinical response at each assessment. We analysed both the overall trial population and a subgroup of patients who had attended each of the three assessments by modified intention-to-treat. This study is registered at controlled-trials.com, ISRCTN 26715889. We enrolled 940 patients from June 4, 2001, until Dec 16, 2008. Complete clinical response was achieved in 492 (52%) of 940 patients at assessment 1 (11 weeks), 665 (71%) of patients at assessment 2 (18 weeks), and 730 (78%) of patients at assessment 3 (26 weeks). 691 patients attended all three assessments and in this subgroup, complete clinical response was reported in 441 (64%) patients at assessment 1, 556 (80%) at assessment 2, and 590 (85%) at assessments 3. 151 (72%) of the 209 patients who had not had a complete clinical response at assessment 1 had a complete clinical response by assessment 3. In the overall trial population of 940 patients, 5 year overall survival in patients who had a clinical response at assessments 1, 2, 3 was 83% (95% CI 79–86), 84% (81–87), and 87% (84–89), respectively and was 72% (66–78), 59% (49–67), and 46% (37–55) for patients who did not have a complete clinical response at assessments 1, 2, 3, respectively. In the subgroup of 691 patients, 5 year overall survival in patients who had a clinical response at assessment 1, 2, 3 was 85% (81–88), 86% (82–88), and 87% (84–90), respectively, and was 75% (68–80), 61% (50–70), and 48% (36–58) for patients who did not have a complete clinical response at assessment 1, 2, 3, respectively. Similarly, progression-free survival in both the overall trial population and the subgroup was longer in patients who had a complete clinical response, compared with patients who did not have a complete clinical response, at all three assessments. Many patients who do not have a complete clinical response when assessed at 11 weeks after commencing chemoradiotherapy do in fact respond by 26 weeks, and the earlier assessment could lead to some patients having unnecessary surgery. Our data suggests that the optimum time for assessment of complete clinical response after chemoradiotherapy for patients with squamous cell carcinoma of the anus is 26 weeks from starting chemoradiotherapy. We suggest that guidelines should be revised to indicate that later assessment is acceptable. Cancer Research UK.

A Concise Guide to Observational Studies in Healthcare provides busy healthcare professionals with an easy-to-read introduction and overview to conducting, analysing and assessing observational studies. It is a suitable introduction for anyone without prior knowledge of study design, analysis or conduct as the important concepts are presented throughout the text. It provides an overview to the features of design, analyses and conduct of observational studies, without using mathematical formulae, or complex statistics or terminology and is a useful guide for researchers conducting their own studies, those who participate in studies co-ordinated by others, or who read or review a published report of an observational study. Examples are based on clinical features of people, biomarkers, lifestyle habits and environmental exposures, and evaluating quality of care.

Background In rare cancers or subtypes of common cancers, a comparison of multiple promising treatments may be required. The selected treatment can then be assessed against the standard of care (if it exists) or used as a backbone for combinations with new, possibly targeted, agents. There could be different experimental therapies or different doses of the same therapy, and either can be done in combination with standard treatments. A ’pick-the-winner’ design is often used, which focuses on efficacy to select the most promising treatment. However, a treatment with a slightly lower efficacy compared to another treatment may actually be preferred if it has a better toxicity or quality of life profile, is easier to administer, or cheaper. Methods By pre-defining a margin of practical equivalence in order to calculate the sample size, a more flexible assessment can be made of whether the treatments have very different effects or are sufficiently close so that other factors can be used to choose between them. Using exact binomial probabilities, we calculated the sample size for two- and three-arm randomised selection trials including a margin of practical equivalence with a variety of input parameters. Results We explain conceptually the margin of practical equivalence in this paper, and provide a free user-friendly web application to calculate the required sample size for a variety of input parameters. Conclusion The web application should help promote the randomised selection design with a margin of practical equivalence, which provides greater flexibility than the ’pick-the-winner’ approach in assessing the results of selection trials.

Uptake in the national colorectal cancer screening programme in England varies by socioeconomic status. We assessed four interventions aimed at reducing this gradient, with the intention of improving the health benefits of screening. All people eligible for screening (men and women aged 60–74 years) across England were included in four cluster-randomised trials. Randomisation was based on day of invitation. Each trial compared the standard information with the standard information plus the following supplementary interventions: trial 1 (November, 2012), a supplementary leaflet summarising the gist of the key information; trial 2 (March, 2012), a supplementary narrative leaflet describing people's stories; trial 3 (June, 2013), general practice endorsement of the programme on the invitation letter; and trial 4 (July–August, 2013) an enhanced reminder letter with a banner that reiterated the screening offer. Socioeconomic status was defined by the Index of Multiple Deprivation score for each home address. The primary outcome was the socioeconomic status gradient in uptake across deprivation quintiles. This study is registered, number ISRCTN74121020. As all four trials were embedded in the screening programme, loss to follow-up was minimal (less than 0·5%). Trials 1 (n=163 525) and 2 (n=150 417) showed no effects on the socioeconomic gradient of uptake or overall uptake. Trial 3 (n=265 434) showed no effect on the socioeconomic gradient but was associated with increased overall uptake (adjusted odds ratio [OR] 1·07, 95% CI 1·04–1·10, p<0·0001). In trial 4 (n=168 480) a significant interaction was seen with socioeconomic status gradient (p=0·005), with a stronger effect in the most deprived quintile (adjusted OR 1·11, 95% CI 1·04–1·20, p=0·003) than in the least deprived (1·00, 0·94–1·06, p=0·98). Overall uptake was also increased (1·07, 1·03–1·11, p=0·001). Of four evidence-based interventions, the enhanced reminder letter reduced the socioeconomic gradient in screening uptake, but further reducing inequalities in screening uptake through written materials alone will be challenging. National Institute for Health Research.

Many patients with advanced non-small-cell lung cancer (NSCLC) receive only active supportive care because of poor performance status or presence of several comorbidities. We investigated whether erlotinib improves clinical outcome in these patients. TOPICAL was a double-blind, randomised, placebo-controlled, phase 3 trial, done at 78 centres in the UK. Eligibility criteria were newly diagnosed, pathologically confirmed NSCLC; stage IIIb or IV; chemotherapy naive; no symptomatic brain metastases; deemed unsuitable for chemotherapy because of poor (≥2) Eastern Cooperative Oncology Group performance status or presence of several comorbidities, or both; and estimated life expectancy of at least 8 weeks. Patients were randomly assigned (by phone call, in a 1:1 ratio, stratified by disease stage, performance status, smoking history, and centre, block size 10) to receive oral placebo or erlotinib (150 mg per day) until disease progression or unacceptable toxicity. Investigators, clinicians, and patients were masked to assignment. The primary endpoint was overall survival. Analyses were by intention to treat, and prespecified subgroup analyses included development of a rash due to erlotinib within 28 days of starting treatment. This study is registered, number ISRCTN 77383050. Between April 14, 2005, and April 1, 2009, we randomly assigned 350 patients to receive erlotinib and 320 to receive placebo. We followed up patients until March 31, 2011. 657 patients died; median overall survival did not differ between groups (erlotinib, 3·7 months, 95% CI 3·2–4·2, vs placebo, 3·6 months, 3·2–3·9; unadjusted hazard ratio [HR] 0·94, 95% CI 0·81–1·10, p=0·46). 59% (178 of 302) of patients assigned erlotinib and who were assessable at 1 month developed first-cycle rash, which was the only independent factor associated with overall survival. Patients with first-cycle rash had better overall survival (HR 0·76, 95% CI 0·63–0·92, p=0·0058), compared with placebo. Compared with placebo, overall survival seemed to be worse in the group that did not develop first-cycle rash (1·30, 1·05–1·61, p=0·017). Grade 3 or 4 diarrhoea was more common with erlotinib than placebo (8% [28 of 334] vs 1% [four of 313], p=0·0001), as was high-grade rash (23% [79 of 334] vs 2% [five of 313], p<0·0001); other adverse events were much the same between groups. Patients with NSCLC who are deemed unsuitable for chemotherapy could be given erlotinib. Patients who develop a first-cycle rash should continue to receive erlotinib, whereas those who do not have a rash after 28 days should discontinue erlotinib, because of the possibility of decreased survival. Cancer Research UK, Roche.

Between 1990 and 2000, we examined the effect of timing of non-platinum chemotherapy when combined with radiotherapy. We aimed to determine whether giving chemotherapy concurrently with radiotherapy or as maintenance therapy, or both, affected clinical outcome. Here we report survival and recurrence after 10 years of follow-up. Between Jan 15, 1990, and June 20, 2000, 966 patients were recruited from 34 centres in the UK and two centres from Malta and Turkey. Patients with locally advanced head and neck cancer, and who had not previously undergone surgery, were randomly assigned to one of four groups in a 3:2:2:2 ratio, stratified by centre and chemotherapy regimen: radical radiotherapy alone (n=233); radiotherapy with two courses of chemotherapy given simultaneously on days 1 and 14 of radiotherapy (SIM alone; n=166); or 14 and 28 days after completing radiotherapy (SUB alone, n=160); or both (SIM+SUB; n=154). Chemotherapy was either methotrexate alone, or vincristine, bleomycin, methotrexate, and fluorouracil. Patients who had previously undergone radical surgery to remove their tumour were only randomised to radiotherapy alone (n=135) or SIM alone (n=118), in a 3:2 ratio. The primary endpoints were overall survival (from randomisation), and event-free survival (EFS; recurrence, new tumour, or death; whichever occurred first) among patients who were disease-free 6 months after randomisation. Analyses were by intention to treat. This trial is registered at www.Clinicaltrials.gov, number NCT00002476. All 966 patients were included in the analyses. Among patients who did not undergo surgery, the median overall survival was 2·6 years (99% CI 1·9–4·2) in the radiotherapy alone group, 4·7 (2·6–7·8) years in the SIM alone group, 2·3 (1·6–3·5) years in the SUB alone group, and 2·7 (1·6–4·7) years in the SIM+SUB group (p=0·10). The corresponding median EFS were 1·0 (0·7–1·4), 2·2 (1·1–6·0), 1·0 (0·6–1·5), and 1·0 (0·6–2·0) years (p=0·005), respectively. For every 100 patients given SIM alone, there are 11 fewer EFS events (99% CI 1–21), compared with 100 given radiotherapy, 10 years after treatment. Among the patients who had previously undergone surgery, median overall survival was 5·0 (99% CI 1·8–8·0) and 4·6 (2·2–7·6) years in the radiotherapy alone and SIM alone groups (p=0·70), respectively, with corresponding median EFS of 3·7 (99% CI 1·1–5·9) and 3·0 (1·2–5·6) years (p=0·85), respectively. The percentage of patients who had a significant toxicity during treatment were: 11% (radiotherapy alone, n=25), 28% (SIM alone, n=47), 12% (SUB alone, n=19), and 36% (SIM+SUB, n=55) among patients without previous surgery; and 9% (radiotherapy alone, n=12) and 20% (SIM alone, n=24) among those who had undergone previous surgery. The most common toxicity during treatment was mucositis. The percentage of patients who had a significant toxicity at least 6 months after randomisation were: 6% (radiotherapy alone, n=13), 6% (SIM alone, n=10), 4% (SUB alone, n=7), and 6% (SIM+SUB, n=9) among patients who had no previous surgery; and 7% (radiotherapy alone, n=10) and 11% (SIM alone, n=13) among those who had undergone previous surgery. The most common toxicity 6 months after treatment was xerostomia, but this occurred in 3% or less of patients in each group. Concurrent non-platinum chemoradiotherapy reduces recurrences, new tumours, and deaths in patients who have not undergone previous surgery, even 10 years after starting treatment. Chemotherapy given after radiotherapy (with or without concurrent chemotherapy) is ineffective. Patients who have undergone previous surgery for head and neck cancer do not benefit from non-platinum chemotherapy. Cancer Research UK, with support from University College London and University College London Hospital Comprehensive Biomedical Research Centre.

Background Blood‐based tests present a promising strategy to enhance cancer screening through two distinct approaches. In the traditional paradigm of “one test for one cancer”, single‐cancer early detection (SCED) tests a feature high true positive rate (TPR) for individual cancers, but high false‐positive rate (FPR). Whereas multi‐cancer early detection (MCED) tests simultaneously target multiple cancers with one low FPR, offering a new “one test for multiple cancers” approach. However, comparing these two approaches is inherently non‐intuitive. We developed a framework for evaluating and comparing the efficiency and downstream costs of these two blood‐based screening approaches at the general population level. Methods We developed two hypothetical screening systems to evaluate the performance efficiency of each blood‐based screening approach. The “SCED‐10” system featured 10 hypothetical SCED tests, each targeting one cancer type; the “MCED‐10” system included a single hypothetical MCED test targeting the same 10 cancer types. We estimated the number of cancers detected, cumulative false positives, and associated costs of obligated testing for positive results for each system over 1 year when added to existing USPSTF‐recommended cancer screening for 100,000 US adults aged 50–79. Results Compared with MCED‐10, SCED‐10 detected 1.4× more cancers (412 vs. 298), but had 188× more diagnostic investigations in cancer‐free people (93,289 vs. 497), lower efficiency (positive predictive value: 0.44% vs. 38%; number needed to screen: 2062 vs. 334), 3.4× the cost ( $329 M vs. $ 98 M), and 150× higher cumulative burden of false positives per annual round of screening (18 vs. 0.12). Conclusions A screening system for average‐risk individuals using multiple SCED tests has a higher rate of false positives and associated costs compared with a single MCED test. A set of SCED tests with the same sensitivity as standard‐of‐care screening detects only modestly more cancers than an MCED test limited to the same set of cancers.

Background On-site source data verification is a common and expensive activity, with little evidence that it is worthwhile. Central statistical monitoring (CSM) is a cheaper alternative, where data checks are performed by the coordinating centre, avoiding the need to visit all sites. Several publications have suggested methods for CSM; however, few have described their use in real trials. Methods R-programs were created to check data at either the subject level (7 tests within 3 programs) or site level (9 tests within 8 programs) using previously described methods or new ones we developed. These aimed to find possible data errors such as outliers, incorrect dates, or anomalous data patterns; digit preference, values too close or too far from the means, unusual correlation structures, extreme variances which may indicate fraud or procedural errors and under-reporting of adverse events. The methods were applied to three trials, one of which had closed and has been published, one in follow-up, and a third to which fabricated data were added. We examined how well the methods work, discussing their strengths and limitations. Results The R-programs produced simple tables or easy-to-read figures. Few data errors were found in the first two trials, and those added to the third were easily detected. The programs were able to identify patients with outliers based on single or multiple variables. They also detected (1) fabricated patients, generated to have values too close to the multivariate mean, or with too low variances in repeated measurements, and (2) sites which had unusual correlation structures or too few adverse events. Some methods were unreliable if applied to centres with few patients or if data were fabricated in a way which did not fit the assumptions used to create the programs. Outputs from the R-programs are interpreted using examples. Limitations Detecting data errors is relatively straightforward; however, there are several limitations in the detection of fraud: some programs cannot be applied to small trials or to centres with few patients (<10) and data falsified in a manner which does not fit the program’s assumptions may not be detected. In addition, many tests require a visual assessment of the output (showing flagged participants or sites), before data queries are made or on-site visits performed. Conclusions CSM is a worthwhile alternative to on-site data checking and may be used to limit the number of site visits by targeting only sites which are picked up by the programs. We summarise the methods, show how they are implemented and that they can be easy to interpret. The methods can identify incorrect or unusual data for a trial subject, or centres where the data considered together are too different to other centres and therefore should be reviewed, possibly through an on-site visit.