BACKGROUND: Most patients diagnosed with lung cancer present with symptoms. It is not known if the proportions of patients presenting with each symptom has changed over time. Identifying trends in lung cancer's presenting symptoms is important for medical education and early-diagnosis initiatives. AIM: to identify the first reported symptom of possible lung cancer (index symptom), and to test whether the percentages of patients with each index symptom changed during 2000-2017. DESIGN AND SETTING: This was a serial, cross-sectional, observational study using UK Clinical Practice Research Datalink (CPRD) data with cancer registry linkage. METHOD: the index symptom was identified for patients with an incident diagnosis of lung cancer in annual cohorts between 1 January 2000 and 31 December 2017. Searches were constrained to symptoms in National Institute for Health and Care Excellence (NICE) suspected-cancer referral guidelines, and to the year before diagnosis. Generalised linear models (with a binomial function) were used to test if the percentages of patients with each index symptom varied during 2000-2017. RESULTS: the percentage of patients with an index symptom of cough (odds ratio [OR] 1.01; 95% confidence interval [CI] = 1.00 to 1.02 per year; P

BACKGROUND: Pre-existing concurrent medical conditions (multimorbidity) complicate cancer diagnosis when they provide plausible diagnostic alternatives for cancer symptoms. AIM: to investigate associations in bladder cancer between: first, pre-existing condition count and advanced-stage diagnosis; and, second, comorbidities that share symptoms with bladder cancer and advanced-stage diagnosis. DESIGN AND SETTING: This observational UK cohort study was set in the Clinical Practice Research Datalink with Public Health England National Cancer Registration and Analysis Service linkage. METHOD: Included participants were aged ≥40 years with an incident diagnosis of bladder cancer between 1 January 2000 and 31 December 2015, and primary care records of attendance for haematuria, dysuria, or abdominal mass in the year before diagnosis. Stage at diagnosis (stage 1 or 2 versus stage 3 or 4) was the outcome variable. Putative explanatory variables using logistic regression were examined, including patient-level count of pre-existing conditions and 'alternative-explanations', indicating whether pre-existing condition(s) were plausible diagnostic alternatives for the index cancer symptom. RESULTS: in total, 1468 patients (76.4% male) were studied, of which 399 (35.6%) males and 217 (62.5%) females had alternative explanations for their index cancer symptom, the most common being urinary tract infection with haematuria. Females were more likely than males to be diagnosed with advanced-stage cancer (adjusted odds ratio [aOR] 1.62; 95% confidence interval [CI] = 1.20 to 2.18; P = 0.001). Alternative explanations were strongly associated with advanced-stage diagnosis in both sexes (aOR 1.69; 95% CI = 1.20 to 2.39; P = 0.003). CONCLUSION: Alternative explanations were associated with advanced-stage diagnosis of bladder cancer. Females were more likely than males to be diagnosed with advanced-stage disease, but the effect was not driven entirely by alternative explanations.

We examine the suitability of three methods using patient-level data to evaluate the time-varying impacts of national healthcare guidelines. Such guidelines often codify progressive change and are implemented gradually; for example, National Institute for Health and Care Excellence (NICE) suspected-cancer referral guidelines. These were revised on June 23, 2015, to include more cancer symptoms and test results ("features"), partly reflecting changing practice. We explore the time-varying impact of guideline revision on time to colorectal cancer diagnosis, which is linked to improved outcomes in decision-analytic models. We included 11,842 patients diagnosed in 01/01/2006-31/12/2017 in the Clinical Practice Research Datalink with England cancer registry data linkage. Patients were classified by whether their first pre-diagnostic cancer feature was in the original guidelines (NICE-2005) or was added during the revision (NICE-2015-only). Outcome was diagnostic interval: time from first cancer feature to diagnosis. All analyses adjusted for age and sex. Two difference-in-differences analyses used either a Pre (01/08/2012-31/12/2014, n = 2243) and Post (01/08/2015-31/12/2017, n = 1017) design, or event-study cohorts (2006-2017 vs 2015) to estimate change in diagnostic interval attributable to official implementation of the revised guidelines. A semiparametric varying-coefficient model analysed the difference in diagnostic interval between the NICE groups over time. After model estimation, primary and broader treatment effects of guideline content and implementation were measured. The event-study difference-in-differences and the semiparametric varying-coefficient methods showed that shorter diagnostic intervals were attributable to official implementation of the revised guidelines. This impact was only detectable by pre-to-post difference-in-differences when the pre/post periods were selected according to the estimation results from the varying-coefficient model. Formal tests of the parametric models, which are special cases of the semiparametric model, suggest that they are misspecified. We conclude that the semiparametric method is well suited to explore the time-varying impacts of guidelines codifying progressive change.

© British Journal of General Practice Background Decision-support tools quantify the risk of undiagnosed cancer in symptomatic patients, and may help GPs when making referrals. Aim to quantify the availability and use of cancer decision-support tools (QCancer® and risk assessment tools) and to explore the association between tool availability and 2-week-wait (2WW) referrals for suspected cancer. Design and setting a cross-sectional postal survey in UK primary care. Methods Out of 975 UK randomly selected general practices, 4600 GPs and registrars were invited to participate. Outcome measures included the proportions of UK general practices where cancer decision-support tools are available and at least one GP uses the tool. Weighted least-squares linear regression with robust errors tested the association between tool availability and number of 2WW referrals, adjusting for practice size, sex, age, and Index of Multiple Deprivation. Results in total, 476 GPs in 227 practices responded (response rates: practitioner, 10.3%; practice, 23.3%). At the practice level, 83/227 (36.6%, 95% confidence interval [CI] = 30.3 to 43.1) practices had at least one GP or registrar with access to cancer decision-support tools. Tools were available and likely to be used in 38/227 (16.7%, 95% CI = 12.1 to 22.2) practices. In subgroup analyses of 172 English practices, there was no difference in mean 2WW referral rate between practices with tools and those without (mean adjusted difference in referrals per 100 000: 3.1, 95% CI = -5.5 to 11.7). Conclusion This is the first survey of cancer decision-support tool availability and use. It suggests that the tools are an underused resource in the UK. Given the cost of cancer investigation, a randomised controlled trial of such clinical decision-support aids would be appropriate.

BACKGROUND: Pre-existing non-cancer conditions may complicate and delay colorectal cancer diagnosis. METHOD: Incident cases (aged ⩾40 years, 2007-2009) with colorectal cancer were identified in the Clinical Practice Research Datalink, UK. Diagnostic interval was defined as time from first symptomatic presentation of colorectal cancer to diagnosis. Comorbid conditions were classified as 'competing demands' (unrelated to colorectal cancer) or 'alternative explanations' (sharing symptoms with colorectal cancer). The association between diagnostic interval (log-transformed) and age, gender, consultation rate and number of comorbid conditions was investigated using linear regressions, reported using geometric means. RESULTS: Out of the 4512 patients included, 72.9% had ⩾1 competing demand and 31.3% had ⩾1 alternative explanation. In the regression model, the numbers of both types of comorbid conditions were independently associated with longer diagnostic interval: a single competing demand delayed diagnosis by 10 days, and four or more by 32 days; and a single alternative explanation by 9 days. For individual conditions, the longest delay was observed for inflammatory bowel disease (26 days; 95% CI 14-39). CONCLUSIONS: the burden and nature of comorbidity is associated with delayed diagnosis in colorectal cancer, particularly in patients aged ⩾80 years. Effective clinical strategies are needed for shortening diagnostic interval in patients with comorbidity.

OBJECTIVE: Analysis of routinely collected electronic health record (EHR) data from primary care is reliant on the creation of codelists to define clinical features of interest. To improve scientific rigour, transparency and replicability, we describe and demonstrate a standardised reproducible methodology for clinical codelist development. DESIGN: We describe a three-stage process for developing clinical codelists. First, the clear definition a priori of the clinical feature of interest using reliable clinical resources. Second, development of a list of potential codes using statistical software to comprehensively search all available codes. Third, a modified Delphi process to reach consensus between primary care practitioners on the most relevant codes, including the generation of an 'uncertainty' variable to allow sensitivity analysis. SETTING: These methods are illustrated by developing a codelist for shortness of breath in a primary care EHR sample, including modifiable syntax for commonly used statistical software. PARTICIPANTS: the codelist was used to estimate the frequency of shortness of breath in a cohort of 28 216 patients aged over 18 years who received an incident diagnosis of lung cancer between 1 January 2000 and 30 November 2016 in the Clinical Practice Research Datalink (CPRD). RESULTS: of 78 candidate codes, 29 were excluded as inappropriate. Complete agreement was reached for 44 (90%) of the remaining codes, with partial disagreement over 5 (10%). 13 091 episodes of shortness of breath were identified in the cohort of 28 216 patients. Sensitivity analysis demonstrates that codes with the greatest uncertainty tend to be rarely used in clinical practice. CONCLUSIONS: Although initially time consuming, using a rigorous and reproducible method for codelist generation 'future-proofs' findings and an auditable, modifiable syntax for codelist generation enables sharing and replication of EHR studies. Published codelists should be badged by quality and report the methods of codelist generation including: definitions and justifications associated with each codelist; the syntax or search method; the number of candidate codes identified; and the categorisation of codes after Delphi review.

OBJECTIVES: to estimate data loss and bias in studies of Clinical Practice Research Datalink (CPRD) data that restrict analyses to Read codes, omitting anything recorded as text. DESIGN: Matched case-control study. SETTING: Patients contributing data to the CPRD. PARTICIPANTS: 4915 bladder and 3635 pancreatic, cancer cases diagnosed between 1 January 2000 and 31 December 2009, matched on age, sex and general practitioner practice to up to 5 controls (bladder: n=21â€
718; pancreas: n=16â€
459). The analysis period was the year before cancer diagnosis. PRIMARY AND SECONDARY OUTCOME MEASURES: Frequency of haematuria, jaundice and abdominal pain, grouped by recording style: Read code or text-only (ie, hidden text). The association between recording style and case-control status (χ(2) test). For each feature, the odds ratio (OR; conditional logistic regression) and positive predictive value (PPV; Bayes' theorem) for cancer, before and after addition of hidden text records. RESULTS: of the 20 958 total records of the features, 7951 (38%) were recorded in hidden text. Hidden text recording was more strongly associated with controls than with cases for haematuria (140/336=42% vs 556/3147=18%) in bladder cancer (χ(2) test, p

BACKGROUND: Diagnosis of bladder cancer relies on investigation of symptoms presented to primary care, notably visible haematuria. The importance of non-visible haematuria has never been estimated. AIM: to estimate the risk of bladder cancer with non-visible haematuria. DESIGN AND SETTING: a case-control study using UK electronic primary care medical records, including uncoded data to supplement coded records. METHOD: a total of 4915 patients (aged ≥40 years) diagnosed with bladder cancer between January 2000 and December 2009 were selected from the Clinical Practice Research Datalink and matched to 21 718 controls for age, sex, and practice. Variables for visible and non-visible haematuria were derived from coded and uncoded data. Analyses used multivariable conditional logistic regression, followed by estimation of positive predictive values (PPVs) for bladder cancer using Bayes' theorem. RESULTS: Non-visible haematuria (coded/uncoded data) was independently associated with bladder cancer: odds ratio (OR) 20 (95% confidence interval [CI] =12 to 33). The PPV of non-visible haematuria was 1.6% (95% CI = 1.2 to 2.1) in those aged ≥60 years and 0.8% (95% CI = 0.1 to 5.6) in 40-59-year-olds. The PPV of visible haematuria was 2.8% (95% CI = 2.5 to 3.1) and 1.2% (95% CI = 0.6 to 2.3) for the same age groups respectively, lower than those calculated using coded data alone. The proportion of records of visible haematuria in coded, rather than uncoded, format was higher in cases than in controls (P

1. Time courses for the uptake of L-lactate, D-lactate and pyruvate into isolated cardiac ventricular myocytes from guinea pig were determined at 11 degrees C or 0 degrees C (for pyruvate) in a citrate-based buffer by using a silicone-oil-filtration technique. These conditions enabled initial rates of transport to be measured without interference from metabolism of the substrates. 2. At a concentration of 0.5 mM, transport of all these substrates was inhibited by approx. 90% by 5 mM-alpha-cyano-4-hydroxycinnamate; at 10 mM-L-lactate a considerable portion of transport could not be inhibited. 3. Initial rates of L-lactate and pyruvate uptake in the presence of 5 mM-alpha-cyano-4-hydroxycinnamate were linearly related to the concentration of the monocarboxylate and probably represented diffusion of the free acid. The inhibitor-sensitive component of uptake obeyed Michaelis-Menten kinetics, with Km values for L-lactate and pyruvate of 2.3 and 0.066 mM respectively. 4. Pyruvate and D-lactate inhibited the transport of L-lactate, with Ki values (competitive) of 0.077 and 6.6 mM respectively; the Ki for pyruvate was very similar to its Km for transport. The Ki for alpha-cyano-4-hydroxycinnamate as a non-competitive inhibitor was 0.042 mM. 5. These results indicate that L-lactate, D-lactate and pyruvate share a common carrier in guinea-pig cardiac myocytes; the low stereoselectivity for L-lactate over D-lactate and the high affinity for pyruvate distinguish it from the carrier in erythrocytes and hepatocytes. The metabolic roles for this novel carrier in heart are discussed.