Antenatal screening for Down's syndrome relies on the use of multiple markers in combination. Markers that are highly correlated can cause statistical instability. We used the maximum variance inflation factor (VIFmax) to determine whether a screening test using multiple markers was robust to imprecision in the estimation of the marker distribution parameters.
The VIFmax for a specified screening test was calculated from the correlations between markers in Down's syndrome pregnancies for six tests: integrated and serum integrated tests without repeat measurements, both tests with repeat measurements across trimesters analysed in the standard way, and both tests with repeat measurements analysed as cross-trimester (CT) marker ratios. The screening performance of each test using published parameter values, in terms of the false-negative rates for a 3% false-positive rate (FN3), were calculated for simulated populations with medians 0.2 standard deviations (SD) higher or lower than the published values (to reflect imprecision in parameter estimation) for pregnancy-associated plasma protein A and unconjugated oestriol in affected pregnancies. For each test, the VIFmax value was compared with the coefficient of variation of the FN3 (FN3 CV). An independent set of 27 Down's syndrome pregnancies was used to determine how many had meaningless low risks (<1 in 10,000) with each test.
Tests with VIFmax values greater than 5 had FN3CV values over 50%, but those with VIFmax values less than 5 had FN3 CV values less than 21%. The numbers of Down's syndrome pregnancies with meaningless low risk estimates in the independent set were 18 (64%) in tests with VIFmax values ≥5 and none for those with values <5.
VIFmax values of 5 or more suggest instability. The tests using CT marker ratios were stable (VIFmax < 3), but the tests using repeat measurements in the standard manner were not (VIFmax > 5).
To determine the prevalence of glucose-6-phosphate dehydrogenase (G6PD) deficiency in Isfahan, the central state of Iran.
From February to March 2006, a total of 2501 samples were screened for the quantitative measurement of G6PD activity by enzymatic colorimetric assay by a commercial kit (GAMMA, Belgium). The neonates were referred from 17 delivery units to the Isfahan neonatal screening center at 3–7 days after birth. Any neonate with a value < 6.4 U/gHb was considered G6PD deficient.
Of the 2501 newborns (1307 males, 1194 females) screened, 79 neonates were found to have G6PD deficiency (67 males, 12 females). The overall incidence of G6PD deficiency was 3.2%. Frequency in male population was 5.1 % (67 out of 1307 male neonates) and in female population was 1% (12 out of 1194 female neonates).The female:male ratio was 1:5.5 (P = 0.0001). The mean enzyme activity in deficient patients was 3.22 ± 1.8 U/gHb (male deficient group; 3.17 ± 1.74 U/gHb, female deficient group; 3.49 ± 2.17 U/gHb, P = 0.58).
Routine neonatal screening in Isfahan, Iran with a relatively high prevalence of G6PD deficiency is justified and meets the World Health Organization recommendation.
To evaluate the screening performance of 6/6 and 6/12 vision cut-offs with an illiterate E-chart implemented by a public health nurse to test children for ocular abnormalities and uncorrected refractive error. The gold standard diagnosis is an eye examination performed by an ophthalmologist.
A cross-sectional population-based study was conducted among 2113 students' ages 6–7 and 13–14 years old in 70 Northern District Israeli schools.
Students were tested by nurses and ophthalmologists. A nurse examination was carried out using the illiterate E-chart for vision measurement. The medical examination included vision history, clinical eye examination, vision and retinoscopy testing. The Physician's evaluation of whether students needed a referral for diagnostic procedures, treatment and/or follow-up was recorded. Screening test's performance was determined using ophthalmologist's decision regarding referral as the gold standard. Detection rate (DR), false-positive rate (FPR), odds affected positive result (OAPR), positive predictive value (PPV) and negative predictive value (NPV) were estimated overall and by students' demographic characteristics.
For vision >6/6 cut-off in at least one eye (eyes tested separately): DR – 71.9% (95% CI 65.8–78.7%), FPR – 22.8% (95% CI 17.9–28.9%), OAPR – 0.98:1 (95% CI 0.84:1–1.15:1), PPV – 52.7% (95% CI 45.4–61.2%), NPV – 90.9% (95% CI 88.7–93.1%). For 6/12 vision cut-off, namely vision 6/12 or worse in both eyes (tested separately): DR – 58.6 (95% CI 51.8–66.4%), FPR – 15.2% (95% CI 10.9–21.1%), OAPR – 1.13:1 (95% CI 0.94:1–1.35:1), PPV – 61.1% (95% CI 52.9–70.6%), NPV – 87.6% (95% CI 84.9–90.4%).
Vision-screening test performance measures are mild. It is suggested to change vision cut-off level that denotes vision abnormality from current policy of vision not equal 6/6 in both eyes (tested separately) to vision 6/12 or worse in both eyes (tested separately). This change will result in reduction of FPR from 22% to 15%, concomitant with an increase in false-negative rate from 28% to 41%. Students may be equally screened by either a senior or a less experienced nurse.
To analyse the relationship between the false-positive/false-negative predictive value (FPPV/FNPV) of the HIV-antibody (HIV-Ab) test and prevalence in different Chinese population groups.
HIV prevalence among different population groups was obtained by a screening survey of blood donors and the national HIV/AIDS surveillance programme in China. Given the sensitivity and specificity of a test kit and the prevalence of HIV infection, the estimated values of FPPV/FNPV were calculated using Bayes' formula. The actual value of FPPV of blood donors was obtained by screening 1,195,286 blood donors.
This study indicates that the FPPV of HIV-Ab enzyme-linked immunosorbent assay (ELISA) assays varies widely in different Chinese populations: about 99.5% in the blood donor population, but only 3.2% in the injecting-drug users in high-risk areas. In 1,195,286 sera specimens from the blood donors, 2439 specimens were HIV-Ab positive by third ELISA, and 11 HIV cases were confirmed by Western blot. The HIV prevalence of the blood donor population in this survey was 0.0009% (11/1,195,286), but the HIV-Ab positive rate of third ELISA is 0.2% (2439/1,195,286) and 222 times higher than the prevalence.
Evaluation of HIV prevalence through the HIV-Ab positive rate by third ELISA will significantly overestimate the true prevalence in a low-prevalence population. Individual HIV-infection status should be taken into consideration when analysing the results of HIV-Ab tests in a population with low infection.
Previous studies have not assessed whether evidence-based information about the outcomes of colorectal cancer screening increases informed choice among people from a range of socioeconomic backgrounds nor have they assessed whether this can be administered away from a health-care provider.
Randomized controlled trial in six primary care locations. Three hundred and fourteen people aged 50–74 years received a self-administered decision aid (DA) booklet about outcomes of biennial faecal occult blood testing (FOBT) screening or government consumer guidelines (G).
Significantly more DA recipients (20.9%) were ‘informed’ compared with G recipients (5.8%) (P = 0.0001, OR 4.32; 95% CI 2.49 to 7.52); the DA did not affect values clarity (61.9% clear after DA versus 59.1% after G) nor screening decisions overall (87.3% would screen after DA versus 90.5% after G). Test uptake at one month was uniformly low (5.2% DA versus 6.6% G); mostly due to being ‘too busy’. DA recipients were more likely to make decisions ‘integrating’ knowledge with values (10.4% DA versus 1.5% G). Decisions not to screen were equally uncommon in both groups but more likely to be uninformed in G (P = 0.03). More DA recipients from all education levels were ‘informed’ (P = 0.02), particularly in lower education (50.0% DA versus 17.8% G) and university-educated groups (79.4% DA versus 32.1% G).
Detailed absolute risk and benefit information about FOBT screening can be effectively used at home by people to increase informed choice. The DA was effective in people with lower education levels.
Unique Protocol ID 211705 ClinicalTrials.gov ID NCT 00148226.
Mean sojourn time (MST) and screening test sensitivity (STS), is usually estimated by Markov models using incidence data from the first screening round and the interval between screening examinations. However, several screening programmes do not have full registration of cancers submerging after screening, and increased use of opportunistic screening over time can raise questions regarding the quality of interval cancer registration.
Based on the earlier used Markov model, formulas for expected number of cases given time since former screening activity was developed. Using questionnaire data for 336,533 women in the Norwegian Breast Cancer Screening Programme (NBCSP), mean square regression estimates of MST and STS were calculated.
In contrast to the previously used method, the new approach gave satisfactory model fit. MST was estimated to 5.6 years for women aged 50–59 years, and 6.9 years for women aged 60–69 years, and STS was estimated to 55% and 60%, respectively. Attempts to add separate parameters for breast cancer incidence without screening, or previous STS, resulted in wide confidence intervals if estimated separately, and non-identifiably if combined.
Previously published results of long MST and low screen test sensitivity were confirmed with the new approach. Questionnaire data on time since previous screening can be used to estimate MST and STS, but the approach is sensitive to relaxing the assumptions regarding the expected breast cancer incidence without screening and constant STS over time.
To examine sociodemographic predictors of self-reported screening attendance, intention to accept human papillomavirus (HPV) testing and willingness to accept vaccination for a daughter under 16.
Home-based, computer-assisted interviews with a population representative sample of British women.
Participants were selected using random probability sampling of the Postcode Address File, 994 women aged 25–64 were included in these analyses. Women reported their attendance at cervical screening and intention to accept an HPV test. A subsample of those with a daughter under 16 years (n = 296) reported their willingness to accept HPV vaccination for their daughter.
Screening attendance was associated with education level (odds ratio [OR] = 1.66, confidence interval [95% CI]: 1.07–2.56) and being married (OR = 2.04, 95% CI: 1.37–3.03). Acceptance of HPV testing was predicted by regular attendance for cervical screening (OR = 1.58, 95% CI: 1.03–2.42) and being from a white background (OR = 2.20, 95% CI: 1.18–4.13). Daughter's age was the only predictor of HPV vaccine acceptance, with mothers whose youngest daughter was 13–16 years old being the most likely to accept vaccination (OR = 2.91, 95% CI: 1.27–6.65).
In contrast to screening attendance, ethnicity plays an important role in HPV testing. Specific cultural barriers should be identified and addressed to ensure ethnic disparities in testing are limited. While marital status is associated with screening attendance, HPV testing could overcome this bias. Sociodemographic variables seem to play a limited role in HPV vaccine acceptance among mothers making vaccine decisions for their daughters, but as with other studies, age of daughter is important. The scientific reasons for vaccinating at 12–13 years should be emphasized in HPV information.
To assess whether human papillomavirus (HPV) testing is a safe enough approach to warrant extension of the screening intervals of baseline Papanicolaou (Pap)–/HPV– women in low-income settings.
Of the >1000 women prospectively followed up as part of the Latin American Screening (LAMS) Study in São Paulo, Campinas, Porto Alegre) and Buenos Aires, 470 women with both baseline cytology and Hybrid Capture 2 (HC2) results available were included in this analysis. These baseline Pap-negative and HC2– or HC2+ women were controlled at six-month intervals with colposcopy, HC2 and Pap to assess the cumulative risk of incident Pap smear abnormalities and their predictive factors.
Of the 470 women, 324 (68.9%) were high-risk HPV (hrHPV) positive and 146 (31.1%) were negative. Having two or more lifetime sex partners (odds ratio [OR] = 2.63; 95% CI 1.70–3.51) and women using hormonal contraception (OR = 2.21; 95% CI 1.40–3.51) were at increased risk for baseline hrHPV infection. Baseline hrHPV+ women had a significantly increased risk of incident abnormal Pap smears during the follow-up. Survival curves deviate from each other starting at month 24 onwards, when hrHPV+ women start rapidly accumulating incident Pap smear abnormalities, including atypical squamous cells (ASC) or worse (log-rank; P < 0.001), low-grade squamous intraepithelial lesions (LSIL) or worse (P < 0.001) and high-grade squamous intraepithelial lesions (HSIL) (P = 0.03). Among the baseline hrHPV– women, the acquisition of incident hrHPV during the follow-up period significantly increased the risk of incident cytological abnormalities (hazard ratio = 3.5; 95% CI 1.1–11.7).
These data implicate that HPV testing for hrHPV types might be a safe enough approach to warrant extension of the screening interval of hrHPV–/Pap–women even in low-resource settings. Although some women will inevitably contract hrHPV, the process to develop HSIL will be long enough to enable their detection at the next screening round (e.g. after three years).
The main objective of the national screening programme is to reduce the risk of sight loss among people with diabetes due to diabetic retinopathy (DR).
Offering two-field mydriatic digital photographic screening to all people with diabetes in England over the age of 12 years.
The programme is in its infancy, receiving the first year's annual reports from approximately 96 screening programmes, each of which have developed to offer screening to a minimum number of 12,000 people with diabetes, which would cover a population of 350,000 people with 3.4% diabetes prevalence. The national programme has commenced the External quality assurance (QA) programme in order to achieve and sustain the highest possible standards.
England has a population of two million people with diabetes over the age of 12 and it is believed that there is a prevalence of blindness of 4200 and an annual incidence of blindness of 1280 people with diabetes. This programme has the potential to reduce the prevalence of blindness in England from 4200 people to 1000 people and a conservative estimate of reducing the annual incidence of DR blindness by one-third would save 427 people per annum from blindness. These figures are based on the UK certification of blindness but if World Health Organization (WHO) definitions are used the prevalence, incidence and potential reductions in blindness are much greater.
To see whether revised screening standards and health-professional training are associated with changes in the performance of a neonatal screening programme for congenital hypothyroidism (CHT).
Screening data from the regional screening service in Durham and Newcastle, which covers north-east England and North Cumbria.
We assessed the timing of the different stages of the screening process leading up to the introduction of the revised guidelines between April 2004 and March 2005 (year 1) and afterwards between April 2005 and March 2006 (year 2) in all babies notified as having CHT. We also assessed the interval between sampling and specimen arrival in the laboratory at the beginning and end of year 2 in all babies screened.
Twenty-three babies tested positive or borderline in year 1 and 18 babies in year 2. There was reduced variability in the overall time from birth to notification in year 2 versus year 1 (P = 0.001). This reduction was a consequence of a reduced interval between sample collection and arrival in the laboratory (P = 0.047) and for the laboratory to notify the positive test result (P = 0.003). There was a reduction in the mean time from sampling to receipt by the laboratory in the 2997 babies screened in the final month compared with the 2498 babies screened in the first month of year 2 (P = 0.01).
There was an improvement in neonatal screening programme performance around the time that revised neonatal screening guidelines were introduced. This highlights the importance of ongoing education and training for those involved in screening programmes.
This paper reports early screening results from the newborn sickle cell disease screening programme recently implemented in England.
England. Screening is offered at 5–8 days of age as part of the existing bloodspot test and offered to all babies irrespective of ethnicity.
The laboratory methods recommended are high performance liquid chromatography (HPLC) and iso-electric focusing (IEF).15 Two methods of analysis must be applied to all screen positive results. The conditions screened for are:- Sickle cell anaemia (Hb SS), Hb SC disease, Hb S/β-thalassaemia, Hb S/DPunjab, Hb S/OArab, Hb S/HPFH. Carriers identified for the common haemoglobin variants are reported to parents and follow-up counselling is offered. A bespoke laboratory quality assurance programme has been established which has defined standards of satifactory performance.
Provisional figures from the first seven months of screening (up to March 2004) 108,255 infants were screened gave a screen positive rate of 1:900 for these high prevalence areas and a carrier rate of 2.7%. Figures for 2004–2005 show about 250 significant screen positive results for sickle cell disorders and about 6,500 carriers were identified. The birth prevalence for screen positive results from 2004–05 is 1:1500. We estimate that when there is countrywide data, the national birth prevalence will be about 1:2000–1:2,500.
The results from the national newborn sickle cell screening programme in England — show that the sickle cell disorders are as common as cystic fibrosis (CF) in England, although the distribution of cases is concentrated in London and other urban areas. The findings and approach to implementation adopted in England may be of interest to other Western European countries with increasing rates of sickle cell disease who are considering such programmes and also to other developed countries.
To analyse the effect of an organized, nationwide breast cancer screening programme on non-organized mammography activities in Hungary.
The nationwide dataset of the Hungarian National Health Insurance Fund Administration covering the years 2000–2005.
Data derived from the nationwide database of the Hungarian National Health Insurance Fund Administration. The study includes all women undergoing mammography before (2000–2001) and after (2002–2003/2004–2005) the introduction of organized screening.
The number of women having non-organized (opportunistic/diagnostic) mammograms was around 250,000 in 2000–2001, but increased to 350,000 in 2005. In the age group 45–64 years in 2000–2001, only 27.4% of all women undergoing mammography were examined within locally-organized programmes. After the introduction of the nationwide programme, this percentage increased to 61.0% in 2002–2003, and 56.3% in 2004–2005. After the introduction of the nationwide organized programme (2002–2003), the proportion of organized screening mammographies remained among the highest in county Hajdú-Bihar (78.4%) and Zala (88.3%) and increased significantly in county Vas (87.7%).
The introduction of an organized nationwide screening programme in Hungary resulted in increases in the number of screening mammographies, and also of non-organized mammographies. Although the ratio of organized screening versus non-organized mammography changed in favour of screening mammographies, there are large within-country differences between counties.
When evaluating screening for the early detection of cancer, it is important to estimate both harms and benefits. One common harm is a false-positive (FP), which is a positive screening result, perhaps followed by an invasive test, with no cancer detected on the diagnostic work-up or within a specified time period. An important goal is to estimate the risk of at least one FP, which we call the cumulative risk of an FP, if persons took a regimen of various screening tests, as is commonly recommended. The estimation is complicated because the data come from a study in which subjects are offered various screening tests in rounds with some missing tests in most subjects. Previous methods for estimating cumulative risk of FPs with a single type of test are not directly applicable, so a new approach was developed.
The tests were ordered by appearance, where the last test was either the first FP (analogous to a failure time) or the last test taken with no FPs having occurred on that test or previously (analogous to a censoring time). We applied a Kaplan-Meier approach for survival analysis with the innovation that the hazard for a first FP for a given test depends on the type of test and number of previous tests of that type which were taken.
The method is illustrated with data from the screening arm of the randomized Prostate, Lung, Colorectal and Ovarian (PLCO) Cancer Screening Trial. With an FP defined as a diagnostic work-up in the absence of cancer (or advanced adenoma) within three years, the probability of at least one FP among 14 tests in men was 60.5% with 95% confidence interval of (59.3%, 61.6%).
A simple estimate is proposed for the probability of at least one FP if persons took a regimen of multiple screening tests of different types. The methodology is useful for summarizing the burden of multiphasic screening programmes.
Evaluation and comparison of the performance of organized and opportunistic screening mammography.
Women attending screening mammography in Denmark in 2000. The study included 37,072 women attending organized screening. Among these, 320 women were diagnosed with breast cancer during follow-up. Opportunistic screening was attended by 2855 women with 26 women being diagnosed with breast cancer. Data on women attending screening were linked with information on cancer status. Each woman was followed with respect to diagnosis of breast cancer (invasive as well as in situ) for a period of two years. Screening outcome and cancer status during follow-up were combined to assess whether the result of the examination was true-positive, true-negative, false-positive or false-negative. Based on this classification, age-adjusted sensitivity and specificity of organized and opportunistic screening were calculated.
Defining BI-RADSTM 4–5 as a positive screening outcome, the overall sensitivity of opportunistic screening was 33.6% and the specificity was 99.1%. Using BI-RADSTM 3–5 as positive, the sensitivity was 37.4% and the specificity was 97.9%. Organized screening (which was not categorized according to BI-RADSTM) had an overall sensitivity of 67.2% and a specificity of 98.4%.
Our study showed a considerably higher sensitivity in organized screening than in opportunistic screening, while the specificity was fairly similar in the two settings. The findings support implementation of population-based breast screening programmes, as recommended in the ‘European guidelines for quality assurance in breast cancer screening and diagnosis’.
The objective of this study is to analyse cervical screening coverage data over time.
Routinely collected cervical screening statistics, in England, on the proportion of women who have undergone cervical screening with cytology during the preceding five years. The participants included all women residents eligible for cervical screening.
Overall coverage remained at about 82% or over between 1995 and 2000. Since 2000, however, coverage has drifted slowly down to just over 80% in 2005. Coverage has long been observed to be related to age. In 2005, the coverage rate was 71% in women aged 25–29, 83% in those aged 35–54 and 75% in those aged 55–64. Comparing coverage by age in the three years, 1995 – 2000 and 2005 – shows broad conformity with this pattern in each of the three years, but overlaid is the fact that at ages below 50, the rate has been falling while at ages above 55, the rate has been rising. The fall in screening coverage appears to be largely a cohort effect, with women born in the 1960s and later being increasingly less likely to participate.
No specific reason for this effect is evident. Action could be targeted at women aged 25–34 to address falling coverage.
To estimate the total proportional incidence of interval breast cancers in a two-yearly mammography screening programme, and to perform subgroup analyses by woman's age, screening centre-specific recall rate and screening round.
Using unconverted electronic data-sets from the 13 screening centres in the Emilia-Romagna Region of northern Italy (540,450 women aged 50–69 years), a database of 919,538 mammography records was created. Of these, 655,175 eligible single-mammography records (1997–2002) from 379,318 women were record-linked with the regional Breast Cancer Registry. In the two-year inter-screening interval, a total of 1,022,694.3 woman-years at risk were accumulated, with 695 interval cancers observed and 2428.3 expected. The observed number of interval cancers was divided by the expected number to obtain the proportional incidence.
The total proportional incidence of first- and second-year interval cancers was 0.18 (95% CI 0.15–0.20) and 0.43 (0.39–0.47), respectively. Woman's age was inversely associated with proportional incidence in both interval years, with a cut-off point at age 60. A screening centre-specific recall rate greater than the regional average of 5% was associated with a proportional incidence of 0.14 (0.11–0.17) versus 0.20 (0.17–0.24) in the first interval year, and of 0.36 (0.31–0.41) versus 0.50 (0.44–0.56) in the second. The proportional incidence remained unchanged between the first and subsequent screening rounds.
The results were in line with the previous Italian data and with the recommended European standards. The inverse effect of woman's age and of recall rate was expected.
This study aims to determine the association of two versus three year screening intervals with the likelihood of detection of a high-grade cervical abnormality and cervical cancer.
Data were obtained from the New South Wales (NSW) Papanicolaou (Pap) Test Register (PTR) and NSW Central Cancer Registry (CCR). Subjects were human papillomavirus (HPV) unvaccinated women aged 20–69 years who had a minimum of two Pap tests with a negative result at their first recorded Pap test (n = 1,213,295). Logistic regression was used to determine the association between screening interval and the likelihood of: (1) a cytological prediction of high-grade abnormality, defined as cervical intraepithelial neoplasia (CIN) 2 or greater; (2) a histologically confirmed high-grade abnormality; (3) a cytological prediction of cervical cancer and (4) a confirmed diagnosis of cervical cancer, controlling for potential confounders of age and socioeconomic status (SES) of area of residence.
For each year increase in the screening interval, the odds of a histologically confirmed high-grade abnormality increased significantly in women aged 20–29 years (odds ratio [OR] 1.24, 95% confidence interval [CI] 1.20–1.28) and in women aged 30–49 years (OR 1.11, 95% CI 1.06–1.16), but not in women aged 50–69 years (OR 1.08, 95% CI 0.89–1.32). Similar results were observed for cytologically detected high-grade abnormalities. The screening interval was significantly and positively associated with a cytological prediction of cervical cancer (OR 1.40, 95% CI 1.28–1.54) and a confirmed cervical cancer diagnosis (OR 1.66, 95% CI 1.33–2.07) in women aged 20–69 years. We estimate that if the screening interval were increased from two to three years, and the number of women participating in triennial screening participation was the same as for biennial participation in NSW, then 267 (95% CI 186–347) extra cases of high-grade abnormalities would be detected annually by cytology and 225 extra cases (95% CI 160–291) confirmed by histology, mostly confined to women aged 20–49 years. Equivalently, 2.3 (95% CI 1.8–2.8) and 1.9 (95% CI 1.5–2.4) extra cases of high-grade cytology and histology, respectively, would be expected per 1000 women with initially negative cytology if the screening interval were extended from two to three years.
Increasing the cervical screening interval from two to three years would be expected to significantly increase the odds of detection of a high-grade abnormality for NSW women aged 20–49 years and cervical cancer for NSW women aged 20–69 years. Accordingly, our study provides evidence in support of retaining the recommended cervical screening interval at two years for HPV unvaccinated, well women.
Settings: Community survey of asymptomatic volunteers.
Methods: Using the NESTROFT results, haematological data and haemoglobin A2 concentration from 1435 young, asymptomatic Sindhi subjects recruited in a population-based survey, and statistical analysis by classification tree approach, we examined whether haematological parameters have discriminatory utility additional to that of NESTROFT in screening for β-thalassaemia trait.
Results: We observed that in the derivation subset from which the classification tree was generated, there was only a marginal – albeit statistically significant – improvement in the screening performance of NESTROFT, whereas there was no such improvement attributable to the use of haematological parameters in a separate validation subset.
Conclusion: Our results further substantiate the claim that the use of NESTROFT is highly indicated for screening for β-thalassaemia trait in regions where the prevalence is high and the resources are constrained.
]]>Conceptual entities: Sensitivity is the indicator on the ability of screening to find cancer in the detectable preclinical phase (DPCP). The ability is usually specified as to the screening test. We call this entity the test sensitivity. Test positivity with histological confirmation refers to the full diagnostic process and we call the corresponding entity as episode sensitivity. Ultimately, a screening programme identifies a proportion of cancers in the DPCP in the total target population, that we call programme sensitivity. We derive the formulae for these three sensitivities consistent with the incidence method.
Example: Our example on estimation of the three sensitivities is from a randomized screening trial for prostate cancer in Finland. The estimates by incidence method were substantially different, 85% for test sensitivity, 48% for episode sensitivity and 36% for programme sensitivity.
Conclusion: More than one concept of sensitivity with standard method of estimation is needed to describe the ability of screening to identify the disease in the DPCP.
]]>Setting: Data were from a subset of people who received iFOBT at Taipei county of Taiwan in 2005.
Methods: All subjects with positive iFOBT were referred to hospital for further diagnostic examinations. In total, 226 such subjects who did not accept referral within 60 days were identified as the non-compliant group from the record of Public Health Bureau. Frequency-matched 219 subjects were sampled from the 599 people who accepted referral within 60 days as the compliant group. Telephone interviews were performed according to questionnaire designed basically under the Health Belief Model. Multiple logistic regression was used to assess effects of possible associated factors for referral compliance.
Results: A total of 145 persons in the compliant group and 115 persons in the non-compliant group completed the interview. Factors including 'perceived susceptibility' and 'cue to action: information' were positively associated with, while 'casual personality' was negatively associated with referral compliance.
Conclusions: Three factors in Health Belief Model were associated with referral compliance after positive FOBT.
]]>Methods: A total of 22,672 subjects aged 50 years or older were invited to have an uptake of iFOBT. We collected data from screen-detected cases for the cut-off above 100 ng/mL and obtained interval cancers from a nationwide cancer registry for a cut-off below 100 ng/mL. We found a total of 65 colorectal cancer (CRC) cases, including 43 detected by screen and 22 diagnosed between screens (interval cases). The optimal cut-off was first determined by receiver operating characteristics (ROC) curve analysis. Formal economic evaluation was further applied to identifying the optimal cut-off by assessing the minimum incremental cost-effectiveness ratio (ICER), an indicator for cost per life year gained (effectiveness), given a series of cut-offs of iFOBT, ranging from 30 to 200 ng/mL compared with no screening.
Results: ROC curve analysis found the optimal cut-off of iFOBT to be 100 ng/mL at which the sensitivity, false-positive and odds of being affected given a positive result were 81.5% (70.2%–89.2%), 5.7% (5.4%–6.0%) and 1.24 (1.19–1.32), respectively. The area under ROC curve was 0.87 (0.81–0.93). In economic appraisal, the screening programme irrespective of any cut-off dominated (less cost and more effectiveness) over the control group. The optimal cut-off (the lowest ICER) was 110 ng/mL at which an average of 0.054 life year was gained and that of 950 ($US) was saved.
Conclusions: We used cost-effectiveness to identify 110 ng/mL as the optimal cut-off of iFOBT in a Taiwanese population-based screening for CRC. Our model provides a useful approach for health policy-makers in designing population-based screening for CRC to determine the optimal cut-off of iFOBT when cost and effectiveness need to be taken into account.
]]>Methods: Data are collected annually from all UK screening units on standard KC62 return forms.
Results: The prevalence of screen-positive cancer (cancer detection rate) has increased at both rounds during the six-year period. At the incident round, cancer detection rates increased by 24%, from 5.4 per 1000 in 2000 to 6.7 per 1000 in 2005 and the detection of small cancers (≤ 10 mm) has increased by 40%. Generally, quality measures in the programme continue to improve. However, while rates of recall at the incident screen decreased from 3.8% in 2000 to 3.6% in 2005, at the prevalent round, in 2005, 22% of units continued to recall more than 10% of women to assessment.
Conclusions: The results suggest that the performance of the programme continues to improve. In the future, analysis of data on interval cancers will assist the interpretation of cancer detection rates.
]]>Setting: Female breast cancer data in the Korea Central Cancer Registry 2002.
Methods: The stochastic model was based on the threshold method, in which the schedule is determined by a pre-specified threshold value. The threshold value was defined as the probability of being in a preclinical state of breast cancer at age 40 years. The sensitivity of the mammography was specified as 0.7. Two models for mean sojourn time (MST) in the preclinical state were considered; MSTs for Model I were 2 (ages < 50 years), 3 (ages 50–59 years), and 4 years (ages ≥ 60 years), and MSTs for Model II were 3, 4, and 5 years for the corresponding age groups.
Results: The threshold method for Model I generated 19 examinations within the screening ages of 40–69 years. Each screening time was determined at ages 40.0, 41.6, 43.2, 44.8, 46.0, 47.2, 48.4, 49.6, 50.7, 51.7, 52.7, 53.7, 54.7, 56.2, 57.8, 59.4, 61.3, 63.1, and 64.9 years. The schedule sensitivity of Model I was 64.2%, which was higher than that (57.5%) of the biennial periodic schedule. Model II included 11 screenings between the ages of 40 and 69 years and also showed a higher schedule sensitivity, especially for women aged 40 years as compared with the biennial screening.
Conclusions: This finding suggests that the threshold screening schedule for breast cancer increase the schedule sensitivity by reflecting the age-specific incidence rate of a population.
]]>Methods: The study group was derived by finding women common to a consecutive series of 1944 invasive breast cancers diagnosed in Nottingham between 1986 and 1998 with a known basal phenotype status and a prospectively collected database of all screen-detected breast cancers. In total, 356 women constituted the study group. Pathological and radiological features were recorded. Basal cell markers used were CK5/6 (cloneD5/16134) and CK14 (clone LL002). Tumours were classified as of basal phenotype if > or = 10% staining was seen with either marker.
Results: Of all screen-detected lesions, 43 (12%) had a basal immunophenotype and 313 (88%) were non-basal. There were 15 (35%) and 40 (13%) breast cancer deaths in the basal group and nonbasal groups, respectively ( P = 0.0006). On univariate analysis, nodal stage, histological grade, lympho-vascular invasion (LVI) status, invasive size and basal phenotype had prognostic significance. On multivariate analysis, basal phenotype, LVI and nodal stage maintain prognostic significance. Of the 189 women with < 15 mm lesions, eight of 20 (40%) of the basal group and eight of 169 (5%) of the non-basal group died of breast cancer ( P < 0.0001). On multivariate analysis, basal phenotype was the only factor to maintain independent prognostic significance.
Conclusions: Basal phenotype is a powerful prognostic factor for women with small screen-detected invasive breast cancer.
]]>Study design: We retrospectively reviewed case-notes of 34 consecutive singleton fetuses with cystic hygroma referred at 15–18 weeks of gestation. All cases had maternal blood sampled for triple screening at the time of the ultrasound scan.
Results In total, 62% of fetuses with cystic hygroma had abnormal chromosome complements and 80% had a poor outcome. Six fetuses presenting normal values of human chorionic gonadotropin (0.5–2.5 MoM [multiples of the median]), serum alpha-fetoprotein (0.5–2.5 MoM) and unconjugated estriol (>0.5 MoM), normal karyotype and absence of associated structural anomalies had an uneventful outcome.
Conclusions Our data demonstrated that cystic hygroma at 15–18 weeks has a strong association with chromosomal abnormalities. In euploid fetuses, maternal serum screening results may have a role in the diagnostic work-up of the pregnancy.
]]>Methods: Between July 2004 and July 2006, 1825 newborn dried blood samples were collected onto filter papers in four maternity units from Burundi, Rwanda and the East of the Democratic Republic of Congo. We tested for the presence of haemoglobin C and S in the eluted blood by an enzyme-linked immunosorbent assay (ELISA) test using a monoclonal antibody. All ELISA-positive samples (multiple of the median (MoM)≥ 1.5) were confirmed by a simple molecular test. The statistica software version 7.1 was used to create graphics and to fix the MoM cut-off, and the 2 of Pearson was used to compare the genotype incidences between countries.
Results: Of the 1825 samples screened, 97 (5.32%) were positive. Of these, 60 (3.28%) samples were heterozygous for Hb S, and four (0.22%) for Hb C; two (0.11%) newborns were Hb SS homozygotes.
Conclusions: The lower cost and the high specificity of ELISA test are appropriate for developing countries, and such systematic screening for sickle cell anaemia is therefore feasible.
]]>Methods: A retrospective analysis of infants screened during the period 8 November 1995 to 22 July 2006 was performed. Patient data for analyses was restricted to patients with homozygous (Hb SS) sickle cell disease. Published data from the Jamaican Sickle Cell Cohort Study was used to make comparisons with the study sample.
Results: The study sample consisted of 435 patients with Hb SS disease. Acute chest syndrome was the most common clinical (non-death) event accounting for ~50% of all events. Acute splenic sequestration, no longer a significant cause of mortality, was responsible for ~32% of clinical events. Seven deaths (1.8%) occurred during the study period compared with 17.6% to the same age in the Jamaican Sickle Cell Cohort Study. There was a lower proportion of hospital admissions and episodes of serious illness in the study group compared with controls.
Conclusions: Survival estimates for the study sample showed improvement compared with the Jamaican Sickle Cell Cohort Study. This study continues to demonstrate the benefits of, and as such shows support for, newborn screening and early interventions in sickle cell disease. In addition, it highlights some of the areas for continued focus and research development. Although the current system is providing an essential and beneficial service, the study emphasizes the need for newborn screening programmes to be comprehensive care systems to be fully effective.
]]>Setting: Participants were recruited from three clinical programmes in two cities in the province of Ontario, Canada. The study sample was born during two consecutive periods of newborn hearing screening. The first period, prior to 2002, was targeted on high-risk infants only, and the second, from 2002, included both high- and standard-risk infants (universal newborn hearing screening – UNHS). All children were enrolled in rehabilitation programmes focused on oral language development.
Methods: In this multicentre observational study, 65 children under the age of five years with onset of hearing loss before six months of age, 26 identified through systematic newborn screening (14 through targeted screening and 12 through UNHS) and 39 without screening, were assessed with an extensive battery of child- and parent-administered speech and language measures. The degree of hearing loss ranged from mild to profound with 22 children in the mild, moderate and moderately severe categories and 43 in the severe and profound categories. Data are reported for the three-year study period.
Results: The screened group of children was identified at a median age of 6.6 (interquartile range, 3.0–8.2) months and children referred from sources other than newborn screening were diagnosed at a median age of 16.5 (interquartile range, 10.2–29.0) months. Assessment of oral communication development showed no significant difference between the screened and unscreened groups. The communication outcomes for children identified before 12 months of age did not differ from those of later identified children.
Conclusions: Systematic screening of newborn hearing results in earlier identification and intervention for children with permanent hearing loss. Superior language outcome following newborn screening was not demonstrable in the setting of this study.
]]>Methods: Fifteen electronic databases, the internet, key journals and reference lists of included studies were searched. We included diagnostic accuracy studies that compared guaiac or immunochemical FOBTs with any reference standard, for the detection of colorectal cancer in an average-risk adult population, with sufficient data to construct a 2 x 2 table.
Results: Fifty-nine studies were included. Thirty-three evaluated guaiac FOBTs, 35 immunochemical FOBTs and one evaluated sequential FOBTs. Sensitivities for the detection of all neoplasms ranged from 6.2% (specificity 98.0%) to 83.3% (specificity 98.4%) for guaiac FOBTs, and 5.4% (specificity 98.5%) to 62.6% (specificity 94.3%) for immunochemical FOBTs. Specificity ranged from 65.0% (sensitivity 44.1%) to 99.0% (sensitivity 19.3%) for guaiac FOBTs, and 89.4% (sensitivity 30.3%) to 98.5% (sensitivity 5.4%) for immunochemical FOBTs. Diagnostic case–control studies generally reported higher sensitivities. Sensitivities were higher for the detection of CRC, and lower for adenomas, in both the diagnostic cohort and diagnostic case–control studies for both guaiac and immunochemical FOBTs.
Conclusions Immudia HemSp appeared to be the most accurate immunochemical FOBT, however, there was no clear evidence to suggest whether guaiac or immunochemical FOBTs performed better, either from direct or indirect comparisons. Poor reporting of data limited the scope of this review, and the use the Standards for Reporting of Diagnostic Accuracy guidelines is recommended for reporting future diagnostic accuracy studies.
]]>Material and methods: Stage distribution and survival of interval cancers in MMSSP were compared with screen-detected and non-attender cancer cases in MMSSP, with interval cancers in MMST and with breast cancer cases in a non-screened population five years before the start of MMSSP (pre-screening cancer cases).
Results: In MMSSP 1990–99, the interval cancers did not differ in stage distribution or survival compared with cancer cases in non-attenders, while screen-detected cancer cases had more favourable stage distribution and rate of survival than had the interval cancer cases. The MMST interval cancer cases, 1976–1986, had more favourable stage distribution but higher overall case fatality rate, relative risks (RR) 1.78 (1.00–3.20), and breast cancer case fatality rate, RR 2.05 (1.05–4.00), compared with the more recent MMSSP interval cancer cases. No significant difference in five-year survival was seen in the MMSSP interval cancer cases compared with pre-screening cancer cases not exposed to screening.
Conclusion: In this urban population invited to mammographic screening, the survival rate for women with interval cancer has improved over a period of 20 years. Further studies are needed to assess what factors might explain changes in the course of interval breast cancer.
]]>Setting: Ten cross-sectional studies in Burkina Faso, Congo, Guinea, India, Mali and Niger, between 1999 and 2003.
Methods: Using a common protocol, health workers screened 56,147 women aged 25–65 years with VIA and VILI. All women underwent a colposcopy examination and biopsies were taken when necessary. The disease reference standard was histology or negative colposcopy. A positive result on the combined test was defined if either VIA or VILI were positive. The accuracy of the combined test compared with VIA alone or VILI alone was evaluated using likelihood ratios.
Results: The estimated sensitivity and specificity were 81.3% and 87.3%, respectively, for VIA; 91.5% and 86.9% for VILI; and 92.9% and 83.5% for the combined test. The ratio of the positive likelihood ratios of the combined test and VIA alone for CIN 2–3 lesions and cancer was 0.88 (95% confidense interval [CI]: 0.86–0.90), favouring use of VIA alone. The ratio of the negative likelihood ratios was 0.40 (95% CI: 0.37–0.47), favouring use of the combined test. Similar results were obtained when the combined test was compared with VILI alone. Assuming equivalent performance of VIA alone and the combined test with a disease prevalence of 2%, there will be about 16.0 (95% CI: 13.6–18.8) additional FPs for each additional true positive (TP) detected if the combined test is used. This number will be 121.1 (95% CI: 75.4–194.6) if VILI is considered as the single test.
Conclusions: At the trade-off point between the combined test and VIA alone or VILI alone, given the numbers of additional FP results involved for each additional TP case of disease that were found, it would be more likely that settings already using VIA would advocate combined testing, and for settings using VILI to opt for the single test. The additional costs (per 1000 women) incurred with the combined test would be International $4117.68 versus either of the tests above.
]]>Methods: Monte Carlo simulation was used to generate values of two hypothetical markers, A and B, in 1000 affected and 1000 unaffected pregnancies. The means, standard deviations and correlations of A and B were varied in five different examples.
Results: If markers A and B are, on average, higher in affected than unaffected pregnancies and each marker, individually, has the same detection rate for a given false-positive rate (i.e. the same screening performance), then the screening performance of A and B together tends to decrease as A and B become more positively correlated with each other (within affected or unaffected categories) and tends to increase as A and B become more negatively correlated. If A is, on average, higher in affected pregnancies and B is, on average, lower in affected pregnancies (but again each marker has the same screening performance), the opposite pattern is observed; screening performance increases as A and B become more positively correlated and screening performance decreases as they become more negatively correlated. If A and B have unequal screening performances, modest correlations between A and B have little effect on the screening performance of A and B together, but when the correlations are strong whether positive or negative (with r values greater than about 0.45 or less than –0.45) screening performance progressively increases.
Conclusion: Correlations between screening markers considered separately in affected and unaffected pregnancies can either decrease or increase screening performance. In practice, these effects are usually modest, because most screening markers are not highly correlated with each other and the effects become important only with strong correlations, whether positive or negative.
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