Saker

S H Saker

Eduardo Liz

Rebuttal of "Compatibility of local and global stability conditions for some discrete population models", by

Rebuttal of "Compatibility of local and global stability

conditions for some discrete population models," by S.H.

Paper [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] contains two main results (Theorems 2.1 and 2.3), and a corollary (labelled as Theorem 2.2). The author claims that these results are new, and establishes a comparison with some previous papers. The aim of this rebuttal is not only to clarify that this claim is false (that is, these results are not new), but also to show that the proofs in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] are incorrect.

Regarding the novelty of the results, the reader can check that the statement of Theorem 2.1 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] is proved in [2, Theorem 2], and the statement of Theorem 2.3 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] was previously demonstrated in [START_REF] Liz | Local stability implies global stability in some onedimensional discrete single-species models[END_REF]Example 2.3]. We notice that Theorem 2.2 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] is a direct consequence of both Theorem 2.1 and Theorem 2.3. Amazingly, paper [START_REF] Liz | Local stability implies global stability in some onedimensional discrete single-species models[END_REF] is included in the list of references of [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF], but the author does not recognize in his discussion that Theorem 2.3 was already proven there; rather than this, he choose to compare his result with other papers, that were already cited in [START_REF] Liz | Local stability implies global stability in some onedimensional discrete single-species models[END_REF].

The note [START_REF] Liz | A sharp global stability result for a discrete population model[END_REF] is not included in the bibliography of paper [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF], but I suspect that Dr. Saker knows the paper. For instance, he uses exactly the same motivation and references as in [START_REF] Liz | A sharp global stability result for a discrete population model[END_REF].

Secondly, it should be noticed that the techniques used in the proofs of Theorems 2.1 and 2.3 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] are different from the ones employed in [START_REF] Liz | A sharp global stability result for a discrete population model[END_REF][START_REF] Liz | Local stability implies global stability in some onedimensional discrete single-species models[END_REF]. However, as I show below, Saker's proofs are wrong. To justify my assertion, I focus my discussion on the first result, which establishes a global stability condition for the positive E. Liz Departamento de Matemática Aplicada II, E.T.S.E. Telecomunicación, Universidade de Vigo, Campus Marcosende, 36310 Vigo, Spain Tel.: +34-986 812127 Fax: +34-986 812116 E-mail: eliz@dma.uvigo.es equilibrium of the first-order difference equation

x n+1 = f (x n ), n = 0, 1, . . . ( 1
)
where

f (x) = αx + βx 1 + x γ .
(

Here, α, β, γ are real constants satisfying 0 ≤ α < 1 < α + β, γ > 0. Equation (1) has a unique positive equilibrium x, which is locally asymptotically stable if the following condition holds:

γ < 2β (1 -α)(α + β -1) . ( 3
)
Theorem 2.1 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF] states that condition (3) ensures the global asymptotic stability of the positive equilibrium x, that is, all solutions of (1) with positive initial condition converge to x. For it, the author claims that the map V (x) = (x-x) 2 is a (strict) Lyapunov function for [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF]. The same idea is used in the proof of Theorem 2.3.

If one looks closely to the proof of Theorem 2.1 in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF], it becomes apparent that there are some gaps; namely, it is proved that function ΔV (x) := V (f (x)) -V (x) has a local maximum at x = x, with ΔV (x) = 0, but existence of other possible critical points is not excluded. This gap does really matter. Actually, function ΔV can possess other local maxima reaching positive values, so in that case V is not a Lyapunov function for (1), in contradiction with the claim in [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF].

For example, take α = 1/2, β = 9, and γ = 4 in (2). It can be easily verified that inequality (3) holds, but V (x) = (xx) 2 is not a Lyapunov function for [START_REF] Saker | Compatibility of local and global stability conditions for some discrete population models[END_REF]. See Figure 1, where the graph of ΔV (x) is represented.

Fig. 1

 1 Fig. 1 Graph of the function ΔV (x). Clearly, it does not reach a global maximum at x = x.


Timothy B Hallett
email: timothy.hallett@imperial.ac.uk

Ramzi A Alsallaq

Jared M Baeten

Helen Weiss

Connie L Celum

Ron Gray

Laith Abu-Raddad

Will Circumcision Provide Even More Protection From HIV to Women and Men? New Estimates of the Population Impact of Circumcision Interventions

Keywords: HIV, MATHEMATICAL MODEL, CIRCUMCISION, EPIDEMIOLOGY, AFRICA

come

Background

Since three randomised controlled trials demonstrated that adult male circumcision reduces men's chance of HIV acquisition by about 60% (1-3), there has been substantial interest in developing and expanding services providing male circumcision as an additional HIV prevention strategy [START_REF] Who | New Data on Male Circumcision and HIV Prevention: Policy and Programme Implications[END_REF]. An important part of the planning and strategic decision making process for expanding circumcision services has been quantifying the expected impact such interventions might have on the rate of new HIV infections in populations, and several mathematical models have been developed to generate estimates [START_REF] Williams | The potential impact of male circumcision on HIV in Sub-Saharan Africa[END_REF][START_REF] Hallett | Understanding the impact of male circumcision interventions on the spread of HIV in southern Africa[END_REF][START_REF] White | Male circumcision for HIV prevention in sub-Saharan Africa: who, what and when?[END_REF][START_REF] Kahn | Cost-effectiveness of male circumcision for HIV prevention in a South African setting[END_REF][START_REF] Unaids/Who/ | SACEMA Expert Group on Modelling the Impact Cost of Male Circumcision for HIV Prevention. Male Circumcision for HIV Prevention in High HIV Prevalence Settings: What Can Mathematical Modelling Contribute to Informed Decision Making?[END_REF]. To guide HIV prevention policy decisions, these estimates are compared to others evaluating different prevention strategies, such as scaling-up testing and counselling services [START_REF] Who | Guidance On Provider-Initiated HIV Testing And Counselling In Health facilities[END_REF][START_REF] Hallett | The Role of Testing and Counselling For HIV Prevention and Care in the Era of Scaling-Up Antiretroviral Therapy[END_REF], campaigns with messages about abstinence, condom use and reducing multiple partners [START_REF] Halperin | The time has come for common ground on preventing sexual transmission of HIV[END_REF][START_REF] Shelton | Why multiple sexual partners?[END_REF], and universal HIV testing with immediate initiation of antiretroviral therapy (i.e. 'Universal Test and Treat') [START_REF] Granich | Universal voluntary HIV testing with immediate antiretroviral therapy as a strategy for elimination of HIV transmission: a mathematical model[END_REF][START_REF] Dodd | Examining the promise of HIV elimination by 'test and treat' in hyperendemic settings[END_REF]. Previous modelling of the impact of male circumcision on population-level HIV had two important limitations: lack of data about the rate of HIV transmission (both female-to-male and male-to-female) before the wound has healed, and the long-term effect of male circumcision on risk of male-to-female transmission. We hypothesised that these limitations resulted in substantial imprecision in the estimate of the impact of circumcision interventions on HIV spread, both overall and especially among women.

Recently, new data have become available to address these limitations. A pooled analysis from the three trials evaluating the effect of male circumcision on female-to-male HIV transmission found that sex before the wound has healed was associated with a transient increased risk of HIV acquisition in men [START_REF] Mehta | Does sex in the early period after circumcision increase HIV-seroconversion risk? Pooled analysis of adult male circumcision clinical trials[END_REF]. In addition, two recent studies have reported on the relationship between male circumcision and male-to-female HIV transmission. The first, a randomized trial of circumcision from Uganda, found no short-term benefit of circumcision (i.e. to 24 months after the operation) and some suggestion of increased HIV transmission risk to women from circumcised HIV-infected men who resumed sex before the wound was healed [START_REF] Wawer | Circumcision in HIV-infected men and its effect on HIV transmission to female partners in Rakai, Uganda: a randomised controlled trial[END_REF]. The second, an observational analysis from a cohort of 3408 HIV discordant couples from 14 sites in eastern and southern Africa followed for up to 2 years, found that male-to-female HIV transmission risk was 41% lower from circumcised men compared to uncircumcised men (relative transmission rate of 0.59, 95% confidence interval 0.31-1.13; the corresponding estimate adjusted for viral load is: 0.60 (95% confidence interval 0.31-1.16)) [START_REF] Baeten | Male circumcision and risk of male-to-female HIV-1 transmission: a multinational prospective study in African HIV-1-serodiscordant couples[END_REF]. These data mirror those from an earlier observational study in Rakai, Uganda that found a relative male-to-female HIV transmission rate of 0.41 (95% confidence interval 0.10-1.14) in couples with circumcised versus uncircumcised HIV-infected men [START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF]. Together, these data suggest that association between circumcision and male-to-female HIV transmission risk is influenced by the timing of circumcision, with recent circumcision offering no benefit but circumcision at an earlier age (and thus a longer interval between surgery to sex) significantly reducing HIV transmission risk to female partners.

To explore the implications of these new observations for the estimates of the potential impact of circumcision interventions, we adapted two previous published mathematical models -by Hallett et al. [START_REF] Hallett | Understanding the impact of male circumcision interventions on the spread of HIV in southern Africa[END_REF] for Zimbabwe and Alsallaq et al. for Kisumu, Kenya [START_REF] Alsallaq | Quantitative assessment of the role of male circumcision in HIV epidemiology at the population level[END_REF].

Methods

Full descriptions of the two mathematical models have already been published [START_REF] Hallett | Understanding the impact of male circumcision interventions on the spread of HIV in southern Africa[END_REF][START_REF] Alsallaq | Quantitative assessment of the role of male circumcision in HIV epidemiology at the population level[END_REF], and brief descriptions are provided here.

Model 1: Zimbabwe (Hallett et al.) This model is a deterministic compartmental representation of the heterosexual spread of HIV in a Zimbabwe, with HIV prevalence of approximately 20% and initially low rates of circumcision [START_REF] Hallett | Assessing evidence for behaviour change affecting the course of HIV epidemics: A new mathematical modelling approach and application to data from Zimbabwe Epidemics[END_REF]. To capture the heterogeneity in the number of sexual partners, men and women in the model were stratified into risk groups. Men and women form partnerships so that it is more likely that high risk individuals form partnerships with one another. Published data from eastern Zimbabwe [START_REF] Gregson | Sexual mixing patterns and sex-differentials in teenage exposure to HIV infection in rural Zimbabwe[END_REF] were used to inform these parameters specifying sexual behaviour (although the broad behavioural patterns are similar to reports in other settings). Based on observational data from other longitudinal studies [START_REF] Wawer | Rates of HIV-1 transmission per coital act, by stage of HIV-1 infection, in Rakai, Uganda[END_REF], the course of infection is represented by individuals progressing through several stages: acute infection (short duration, high infectiousness), latent infection (long duration, low infectiousness) and pre-AIDS (short duration, high infectiousness). The Zimbabwe model was fit to both the antenatal clinic prevalence data and the DHS prevalence estimate.

Model 2: Kisumu, Kenya (Alsallaq et al.) This model is deterministic compartmental representation of the heterosexual spread of HIV in Kisumu, Kenya calibrated by three sets of prevalence data: the Four City study population survey data [START_REF] Buvé | The multicentre study on factors determining the differential spread of HIV in four African cities: summary and conclusions[END_REF], the Kenya AIDS indicator survey (KAIS) data for Kisumu [START_REF]Kenya AIDS Indicator Survey[END_REF], and the ANC surveillance data [START_REF] Unaids | AIDS epidemic update[END_REF][START_REF] Who/Afro | HIV/AIDS Epidemiological Surveillance Update for the WHO African Region 2002 Country Profiles[END_REF]. Before the intervention, HIV prevalence is approximately 15% and 25% of adult men are circumcised [START_REF] Auvert | Male circumcision and HIV infection in four cities in sub-Saharan Africa[END_REF]. A hierarchy of effective rates of partnership formation were used as a surrogate of sexual-risk to stratify the population into four sexual-risk groups. The effective rates of partnership formation for females were assumed to balance those for males which were adjusted by fitting the model to empirical behaviour data of Kisumu [START_REF] Buvé | The multicentre study on factors determining the differential spread of HIV in four African cities: summary and conclusions[END_REF][START_REF]Kenya AIDS Indicator Survey[END_REF][START_REF] Unaids | AIDS epidemic update[END_REF][START_REF] Who/Afro | HIV/AIDS Epidemiological Surveillance Update for the WHO African Region 2002 Country Profiles[END_REF]. Other parameters were informed by empirical data including transmission probabilities per coital act per HIV stage [START_REF] Wawer | Rates of HIV-1 transmission per coital act, by stage of HIV-1 infection, in Rakai, Uganda[END_REF][START_REF] Abu-Raddad | Dual infection with HIV and malaria fuels the spread of both diseases in sub-Saharan Africa[END_REF], frequency of coital acts per HIV stage [START_REF] Wawer | Rates of HIV-1 transmission per coital act, by stage of HIV-1 infection, in Rakai, Uganda[END_REF], duration of sexual partnerships [START_REF] Morison | Commercial sex and the spread of HIV in four cities in sub-Saharan Africa[END_REF], duration of each HIV stage [START_REF] Wawer | Rates of HIV-1 transmission per coital act, by stage of HIV-1 infection, in Rakai, Uganda[END_REF][START_REF] Morgan | The natural history of HIV-1 infection in Africa[END_REF][START_REF]UNAIDS Reference Group on Estimates, Modelling and Projections[END_REF], and fraction of the population initially in each risk group [START_REF] Morison | Commercial sex and the spread of HIV in four cities in sub-Saharan Africa[END_REF][START_REF] Ferry | Comparison of key parameters of sexual behaviour in four African urban populations with different levels of HIV infection[END_REF]34).

An important limitation is that neither model explicitly accounts for possible interactions between circumcision and other STIs.

Parameterisation of Circumcision and Intervention

In both models, seven independent parameters specify the effect of circumcision on transmission (see Table 1; ranges for sensitivity analysis presented in Table S1). The reduction in male-to-female transmission of HIV was assumed to begin two years after the operation to be consistent with the observational data showing a large effect [START_REF] Baeten | Male circumcision and risk of male-to-female HIV-1 transmission: a multinational prospective study in African HIV-1-serodiscordant couples[END_REF][START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF] and trial data, which did not find evidence for such an effect within 2 years of the operation (recognizing that the trial was not sufficiently powered to do so [START_REF] Wawer | Circumcision in HIV-infected men and its effect on HIV transmission to female partners in Rakai, Uganda: a randomised controlled trial[END_REF]). In the modelled interventions, unless otherwise specified, a constant rate of operations was assumed such that 50% of non-circumcised men in a given population were circumcised in the first 10 years after the intervention starts (HIV negative and positive men had the same chance of being circumcised).

Results

First, we explored the potential impact of the risk of HIV acquisition and transmission being higher during the wound healing period. In a hypothetical intervention which reduced the fraction of the population that are uncircumcised to 50% over 10 years and no men resume sex during wound healing, the HIV incidence rate (new infections per person-years at risk) in the whole population after 10 years is reduced by 20.5% in Zimbabwe (0% circumcised before intervention), and by 7.4% in Kisumu (25% circumcised before intervention). Alternatively, making the most pessimistic assumptions that all men remain sexually active throughout the wound healing period and the mean duration of wound healing is as long as six weeks, incidence rate is still projected to reduce by 19% in Zimbabwe and 6.2% in Kisumu over the same period. Differences between the simulations reduce over time, so the long-term effect of elevated chance of HIV acquisition and transmission during the circumcision wound healing period is likely to be even less.

Next we conducted a meta-analysis of data from the two independent observational cohorts [START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF][START_REF] Baeten | Male Circumcision and Risk of Male-to-Female HIV-1 Transmission: A Multinational Prospectiv Study[END_REF] on the long-term effect of male circumcision on male-to-female HIV transmission. The overall fixed effect point estimate hazard rate was 0.54 (95% confidence interval: 0.31-0.96, p=0.04), indicating 46% reduction in transmission rate from 2 years after the operation. There was no evidence for heterogeneity in effect size (p=0.61).

On the basis of the meta-analysis results, we explored the impact of the reduction in transmission rate from circumcised HIV infected males to their uninfected female partners, after the wound healing period and the subsequent 2 years (as defined in the Methods). Assuming first no effect of circumcision on male-to-female transmission, and using best estimates for the duration of wound healing (Table S1), our simulations suggest that incidence rate would reduce by 19.8% in Zimbabwe or 9.4% in Kisumu after 20 years. Then, assuming an estimated 46% reduction in male-tofemale transmission rate associated with circumcision, commencing 2 years after the procedure, corresponding estimates for the reduction in incidence are 28.0% (Zimbabwe) and 16.8% (Kisumu) (Figure 1(a-b)). This is an increase in epidemiological impact of at least 40% overall.

As expected, the reduction in incidence with the male-to-female transmission effect is most enhanced among women (Figure 1(c-d)). Without an effect of circumcision on male-to-female HIV transmission, incidence is reduced among women by 11.7% in Zimbabwe or 3.9% in Kisumu after 20 years; but, with the direct effect on male-to-female transmission, incidence is instead reduced by 23.7% (Zimbabwe) or 13.9% (Kisumu). Thus the epidemiological impact among women at least almost doubles.

The number of operations required for each HIV infection averted is also substantially reduced if there is a long-term reduction in male-to-female transmission of HIV as a result of male circumcision: in Zimbabwe, from 11 operations per infection averted after 20 years if there is no effect on transmission, to 8 operations (down 28%); and, in Kisumu, from 14 operations per infection to 8 operations over the same period (down 41%).

The multivariate uncertainty was used to determine how the uncertainty in the estimated effects of circumcision propagates to the projections in epidemiological impact (Zimbabwe model only, but qualitatively similar results would be expected for the Kisumu model). With the original assumptions (no effect of wound healing and no effect on male-to-female HIV transmission), the uncertainty intervals for the reduction in incidence after 20 years are: 15-32% overall and 9-21% among women. The new empirical data (Table S1) support corresponding updated estimates of 22-43% reduction in incidence overall, and a 15-39% reduction among women (Figure 2).

If there is no effect of circumcision on male-to-female transmission and a moderate degree of increased risk behaviour following the operation, then it is possible that HIV incidence among women and the population overall could increase following the intervention (Figure S1) -a serious adverse outcome. However, assuming a 46% reduction in the transmission rate with circumcision (2 years after the operation), then our modelling suggests that even very high degrees of increased risk among circumcised men would not lead to increases in incidence overall under the intervention.

Discussion

Our synthesis of new data about the effect of circumcision on female-to-male and male-to-female HIV transmission provides an even greater imperative to increase scale-up of circumcision interventions to prevent HIV [START_REF] Kim | Evaluation of impact of adult male circumcision programs on HIV incidence and prevalence: Current research, gaps in knowledge and recommendations for additional research[END_REF][START_REF] Hankins | Male circumcision for HIV prevention: Operational research priorities[END_REF]. We found that a transient increased risk for HIV transmission (in both directions) during the circumcision wound healing period resulted in only a negligible effect on HIV incidence rate in populations, that potential long-term reduced risk of HIV transmission from HIV-infected circumcised men amplified the benefits of circumcision implementation, and that the benefits of circumcision in these scenarios were particularly great for women.

While short-term increased risk for both female-to-male and male-to-female HIV transmission may occur whilst the wound heals, this effect is very unlikely to materially affect the impact of circumcision interventions at the population level. Nonetheless, careful communication strategies should still be used to promote abstinence for at least two months after the operation to minimise the chance of HIV transmission to female partners.

Our meta-analysis of the best available cohort data suggest that male circumcision reduces the chance of male-to-female HIV transmission with a latent interval of at least two years [START_REF] Baeten | Male circumcision and risk of male-to-female HIV-1 transmission: a multinational prospective study in African HIV-1-serodiscordant couples[END_REF][START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF], and that this substantially amplifies the potential impact of circumcision interventions in populations. Our new projections for the reduction in HIV incidence following circumcision interventions are 40% greater in Zimbabwe and 79% greater for Kisumu, and the number of operations required per infection averted is reduced by more than 25%, when the overall benefit in reduced HIV male-to-female transmission was included. This increases the already attractive costsaving potential of circumcision interventions [START_REF] Kahn | Cost-effectiveness of male circumcision for HIV prevention in a South African setting[END_REF][START_REF] Gray | The impact of male circumcision on HIV incidence and cost per infection prevented: a stochastic simulation model from Rakai, Uganda[END_REF]. (In Text S1, we have also provided a simple formula for making a calculation of operations per infected averted that is generalised to other settings). Indeed, the combination of the two separate effects of circumcision (reducing the chance that men acquire HIV infection and reducing the chance that infected men pass on HIV infection) leads to a more powerful effect than would be expected on the basis of considering the two effects separately (Figure S2).

Importantly, our simulations show very significantly that the benefit of circumcision interventions to women is greatly increased. Although it was previously expected that women would benefit modestly from male circumcision implementation (since fewer of their sexual partners would be infected ( 6)), these new projections show that women could also receive a direct benefit from circumcision. Earlier work had predicted that the population impact of circumcision intervention would be sensitive to this factor [START_REF] Gray | The impact of male circumcision on HIV incidence and cost per infection prevented: a stochastic simulation model from Rakai, Uganda[END_REF][START_REF] Manicaland | STD Prevention Project[END_REF]. If our estimate about the long-term effect of circumcision on male-to-female transmission is correct, then women in stable partnerships with infected men would receive a similar degree of protection from circumcision as men would in partnerships with infected women. We note, however, that the standard of evidence for an effect on male-to-female transmission (meta-analysis of 2 cohort studies) is much less than that for the effect of circumcision of men's chance of acquiring HIV infection (3 randomized trials and more than 36 observational studies ( 40)). Our results do not indicate that HIV-infected men should be targeted for circumcision, since most of the impact of the reduction in male-to-female transmission comes through reducing onward transmission of 'break-through' infections among men circumcised when they were not infected (if only uninfected men are circumcised in the intervention, the eventual reduction in incidence is increased by 30% overall and 66% for women, compared with 44% and 95% respectively if HIV-infected men are circumcised too). Thus, our findings are in support of the current UNAIDS recommendations that men seeking circumcision that are already HIV-infected be discouraged but not be refused the operation.

In addition, earlier concerns about women being exposed to increased risk due to risk compensation among circumcised men are substantially allayed. Our calculations show that it is unlikely that risk compensation following circumcision will undermine the benefits for either men or women.

We adapted two previously developed mathematical models, both of which make a number of simplifying assumptions about the pattern of sex partner contact and HIV natural history, but which have nevertheless proved useful in developing recommendations and strategic decision making in male circumcision interventions [START_REF] Alsallaq | Quantitative assessment of the role of male circumcision in HIV epidemiology at the population level[END_REF][START_REF] Kim | Evaluation of impact of adult male circumcision programs on HIV incidence and prevalence: Current research, gaps in knowledge and recommendations for additional research[END_REF][START_REF] Manicaland | STD Prevention Project[END_REF][START_REF]Modelling the potential impact on the HIV epidemic of scaling up male circumcision programmes: is there much consensus?[END_REF]. Applying the same analysis to two independently constructed models, parameterised for different settings, also indicates that our findings are not specific to one type of modelling approach or one population. There are similarities in the model structure (both are deterministic compartmental transmission models which include stratification by risk group and an account of mixing in sexual contact between risk groups), but substantial difference remain, besides parameterisation, such as the way that the force of infection is calculated and the assumption made about entry and exit to different risk categories. The impact of circumcision interventions is somewhat less in Kisumu (model 2) than in Zimbabwe (model 1), because the basic reproductive ratio is estimated to be higher in Kisumu (this reduces the impact of many types of intervention [START_REF] Garnett | Strategies for limiting the spread of HIV in developing countries: conclusions based on studies of the transmission dynamics of the virus[END_REF]) and 25% of men in Kisumu are already circumcised [START_REF] Alsallaq | Quantitative assessment of the role of male circumcision in HIV epidemiology at the population level[END_REF]. These results should not be interpreted as literal forecasts for either setting since the rate of scale-up of circumcision that will be achieved is far from certain, and the imprecision in the specification of the epidemiological conditions was not fully quantified [START_REF] Garnett | Mathematical models in the evaluation of health programmes[END_REF].

An important limitation is that neither model explicitly accounts for possible interactions between circumcision and other STIs [START_REF] Desai | The role of sexually transmitted infections in male circumcision effectiveness against HIV--insights from clinical trial simulation[END_REF]. Therefore the benefits of circumcision interventions in reducing the incidence of other STIs cannot be quantified here. Any indirect effects on HIV incidence that accrue through reduced STI prevalence will also not be captured in the models, so our results may in fact under-estimate the impact of the interventions. However, earlier work indicates the strength of those indirect effects is expected to small relative to the direct effect of circumcision on HIV acquisition (and transmission) [START_REF] Desai | The role of sexually transmitted infections in male circumcision effectiveness against HIV--insights from clinical trial simulation[END_REF][START_REF] Boily | Incremental role of male circumcision on a generalised HIV epidemic through its protective effect against other sexually transmitted infections: from efficacy to effectiveness to population-level impact[END_REF].

Indirect observational data have produced conflicting evidence on the effect of circumcision on male-to-female HIV transmission, but most studies were severely limited by not knowing the circumcision or infection status of men [START_REF] Kim | Evaluation of impact of adult male circumcision programs on HIV incidence and prevalence: Current research, gaps in knowledge and recommendations for additional research[END_REF]. A recent meta-analysis included data from the Rakai RCT among HIV infected men [START_REF] Wawer | Circumcision in HIV-infected men and its effect on HIV transmission to female partners in Rakai, Uganda: a randomised controlled trial[END_REF], and six longitudinal studies, with 95% confidence interval 0.53 to 1.36 (point estimate 0.80) [START_REF] Weiss | Health risks and benefits of male circumcision for women and HIV infected men: Summary of evidence to date[END_REF]. In this paper, we have restricted analyses to the only two longitudinal studies that directly measured transmission rates from HIV-infected men of known circumcision status at the time of study entry to their regular partners effect [START_REF] Baeten | Male circumcision and risk of male-to-female HIV-1 transmission: a multinational prospective study in African HIV-1-serodiscordant couples[END_REF][START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF]. Both found evidence for a reduction in transmission, although small numbers limited statistical significance in each study when analyzed separately. Our meta-analysis of these highly comparable datasets, however, provides statistically significant evidence of a true effect. The uncertainty in the estimated effect size was, nevertheless, fully represented in this analysis (Figure 2). The Rakai trial [START_REF] Wawer | Circumcision in HIV-infected men and its effect on HIV transmission to female partners in Rakai, Uganda: a randomised controlled trial[END_REF] showed no efficacy in the two years of follow-up of female partners after their male partners were circumcised, suggesting that it may take time for keratinisation of the scar and of glans or other biologic changes to develop, and here we assumed the effect developed 2 years after the operation. Encouragingly, more data will become available from on-going HIV vaccine trials [START_REF] Gray | Safety and Efficacy of a Three-Dose Regimen of an Adenoviral HIV Vaccine (MRKAd5 HIV-1 Gag/Pol/Nef) in HIV Uninfected South African Adults[END_REF][START_REF] Pitisuttithum | Randomized, double-blind, placebo-controlled efficacy trial of a bivalent recombinant glycoprotein 120 HIV-1 vaccine among injection drug users in Bangkok, Thailand[END_REF][START_REF] Rerks-Ngarm | HIV/AIDS preventive vaccine 'primeboost' phase III trial: foundations and initial lessons learned from Thailand[END_REF] and other studies which will contribute to sharpening knowledge about the effect of circumcision and to refining model projections.

In conclusion, projections for the impact of circumcision interventions on population-level HIV may need to be dramatically revised: the impact of male circumcision implementation could be realised much sooner and with greater cost-efficiency than had previously been thought. It is also crucial to recognise and communicate that although male circumcision is an intervention applied to men, it brings substantial benefits for women as well. Premature resumption of sexual activity before the wound is healed or 'compensatory' increases in risk following circumcision are both unlikely to substantively undermine the benefits of male circumcision on HIV incidence among women or men.

Parameter Description

Value Source

Mean duration of wound healing period. 14 days (95% healed within 6 weeks).

Change in the rate of female-to-male HIV transmission before wound is healed versus before circumcision.

increases 3-fold [START_REF] Mehta | Does sex in the early period after circumcision increase HIV-seroconversion risk? Pooled analysis of adult male circumcision clinical trials[END_REF] Change in rate of HIV male-to-female HIV transmission before wound is healed versus before circumcision. increases 3.5-fold

Fraction of men that remain sexually active during the wound healing period. 13%

Change in rate of female-to-male HIV transmission after the wound has healed versus before circumcision.

reduces by 65%

Change in rate of male-to-female HIV transmission prior to two years after the operation versus before circumcision.

No effect [START_REF] Wawer | Circumcision in HIV-infected men and its effect on HIV transmission to female partners in Rakai, Uganda: a randomised controlled trial[END_REF] Change in rate of male-to-female HIV transmission from two years after the operation versus before circumcision.

We estimated this after performing a metaanalysis of data from the two studies among HIV sero-discordant couples that measured this effect. [START_REF] Baeten | Male circumcision and risk of male-to-female HIV-1 transmission: a multinational prospective study in African HIV-1-serodiscordant couples[END_REF][START_REF] Gray | Male circumcision and HIV acquisition and transmission: cohort studies in Rakai, Uganda. Rakai Project Team[END_REF] Table 1: Parameters specifying the effect of male circumcision on the rate of HIV transmission. Ranges used in the uncertainty analysis are presented in Table S1 in the technical appendix. shows the estimate making the original assumptions (i.e. only the effect of circumcision on the chance of acquisition after the wound heals); the blue line show the updated estimates using the new information (i.e. includes effect of circumcision on acquisition/transmission during wound healing and effect on male-to-female transmission after wound healing). Distributions for the parameters are shown in Table S1.

Figure 1 acFigure 1 :

 11 Figure 1 a

Figure 2 :

 2 Figure 2: Multivariate uncertainty analysis of the reduction in HIV incidence 20 years after a circumcision intervention starts, among the whole population (a) and women (b). The black line

Acknowledgements

We thank The Wellcome Trust (TBH) and the US National Institutes of Health (grant 1R01AI083034) for funding support. The funders had no role in study design, data collection and analysis, decision to publish, or preparation of the manuscript

Competing Interests

None declared

Key Messages 1. Circumcision may confer a 46% reduction in the rate of HIV transmission from circumcised men to their female partners.

2. Assuming a reduction in male-to-female transmission, the projected impact of circumcision on HIV spread is substantially enhanced, especially for women. 11

3. An increase in the risk of acquisition and transmission of HIV during circumcision wound healing is unlikely to have a major effect on the population level impact of circumcision interventions.


Darren C Greenwood
email: d.c.greenwood@leeds.ac.uk

Nisreen Alwan

Sinead Boylan

Janet E Cade

Jim Charvill

Karen C Chipps

Marcus S Cooke

Vivien A Dolby

Alastair W M Hay

Shabira Kassam

Sara F L Kirk

Justin C Konje

Neelam Potdar

Susan Shires

Nigel Simpson

Nicholas Taub

James D Thomas

James Walker

Kay L M White

Christopher P Wild

Á S Boylan

Á A W M Hay

Á S F L Kirk

Á S Shires

Á N Simpson

Á J Walker

Á K L M White

Á K C Chipps

Á M S Cooke

Á S Kassam

Á N Potdar

Á N Taub

Caffeine intake during pregnancy, late miscarriage and stillbirth

Keywords: Caffeine, Miscarriage, Still birth Abbreviations CI Confidence interval IQR Interquartile range OR Odds ratio SD

Caffeine is a commonly consumed drug during pregnancy with the potential to affect the developing fetus. Findings from previous studies have shown inconsistent results. We recruited a cohort of 2,643 pregnant women, aged 18-45 years, attending two UK maternity units between 8 and 12 weeks gestation from September 2003 to June 2006. We used a validated tool to assess caffeine intake at different stages of pregnancy and related this to late miscarriage and stillbirth, adjusting for confounders, including salivary cotinine as a biomarker of smoking status. There was a strong association between caffeine intake in the first trimester and subsequent late miscarriage and stillbirth, adjusting for confounders. Women whose pregnancies resulted in late miscarriage or stillbirth had higher caffeine intakes (geometric mean = 145 mg/day; 95% CI: 85-249) than those with live births (103 mg/day; 95% CI: 98-108). Compared to those consuming \ 100 mg/day, odds ratios increased to 2.2 (95% CI: 0.7-7.1) for 100-199 mg/day, 1.7 (0.4-7.1) for 200-299 mg/day, and 5.1 (1.6-16.4) for 300? mg/day (P trend = 0.004). Greater caffeine intake is associated with increases in late miscarriage and stillbirth. Despite remaining uncertainty in the strength of association, our study strengthens the observational evidence base on which current guidance is founded.

Introduction

Caffeine is a commonly consumed drug during pregnancy with the potential to affect the developing foetus [START_REF] Goldstein | Passage of caffeine into human gonadal and fetal tissue[END_REF][START_REF] Morris | Caffeine and the foetus: is trouble brewing?[END_REF][START_REF] Kirkinen | The effect of caffeine on placental and fetal blood flow in human pregnancy[END_REF]. Cohort studies have shown that maternal caffeine intake may be associated with decreased birth weight and fetal growth restriction, [START_REF] Martin | The association between low birth weight and caffeine consumption during pregnancy[END_REF][START_REF] Boylan | Maternal caffeine intake during pregnancy, risk of fetal growth restriction: a large prospective observational study[END_REF] though not consistently [START_REF] Bracken | Association of Maternal Caffeine Consumption with Decrements in Fetal Growth[END_REF]. The only trial conducted suggested no substantial effect of caffeine reduction in the second half of pregnancy [START_REF] Bech | Effect of reducing caffeine intake on birth weight and length of gestation: randomised controlled trial[END_REF]. The evidence for caffeine consumption earlier in pregnancy to be a risk factor for miscarriage may be stronger, [START_REF] Wen | The associations of maternal caffeine consumption and nausea with spontaneous abortion[END_REF][START_REF] Tolstrup | Does caffeine and alcohol intake before pregnancy predict the occurrence of spontaneous abortion?[END_REF][START_REF] Khoury | Consequences of smoking and caffeine consumption during pregnancy in women with type 1 diabetes[END_REF][START_REF] Weng | Maternal caffeine consumption during pregnancy and the risk of miscarriage: a prospective cohort study[END_REF] but again there is still uncertainty [START_REF] Mills | Moderate caffeine use and the risk of spontaneous abortion and intrauterine growth retardation[END_REF][START_REF] Klebanoff | Maternal serum paraxanthine, a caffeine metabolite, and the risk of spontaneous abortion[END_REF][START_REF] Savitz | Caffeine and Miscarriage Risk[END_REF] and no large pragmatic effectiveness trials have been conducted. Most guidelines recommend limiting caffeine intake immediately before and during pregnancy as a precaution, [START_REF]Consumer products and the environment. Statement on the review of caffeine[END_REF][START_REF]Statement on the reproductive effects of caffeine[END_REF][START_REF]Statement on the reproductive effects of caffeine[END_REF][START_REF]Organisation of teratology information specialists[END_REF] whilst recognising that the strength of any association is still unclear [START_REF]Education pamphlet AP001-nutrition during pregnancy[END_REF].

The inconsistency in results may be attributable to one or more sources of bias, including imprecise estimation of caffeine intake, assuming that coffee or tea are the only sources of caffeine, retrospective measurement of assessment of caffeine consumption, or inadequate control for important confounding factors, especially smoking and alcohol consumption. To contribute to understanding the role of caffeine during pregnancy, using improved methodology, we examined the association of maternal caffeine intake with late miscarriage and intrauterine fetal death (stillbirth), using a highly detailed caffeine assessment tool, previously validated against both a food and drink diary and repeated biomarkers of caffeine intake (salivary caffeine and paraxanthine), to prospectively quantify total caffeine intake, from all possible sources, in the first trimester and throughout pregnancy [START_REF] Boylan | Assessing caffeine exposure in pregnant women[END_REF]. We aimed to adequately correct for potentially important confounding from alcohol intake and most importantly from smoking using cotinine as a biomarker in a large cohort of pregnant women.

Methods

Subjects

A cohort of 2,643 pregnant women aged 18-45 years with singleton pregnancies attending two large UK teaching hospital maternity units was recruited between September 2003 and June 2006 between 8 and 12 weeks gestation, as described in detail elsewhere [START_REF] Boylan | Maternal caffeine intake during pregnancy, risk of fetal growth restriction: a large prospective observational study[END_REF]. Women with prior chronic disease, psychiatric illness, HIV or hepatitis B infection, were excluded. The demographic details of age, parity, maternal height, weight, socioeconomic status and gestational age were recorded by a questionnaire. Local ethics committee approval was obtained at both centres and written (signed) informed consent was obtained from each volunteer before recruitment.

Outcomes and measurements

Information regarding antenatal pregnancy complications and delivery details were obtained from the electronic maternity databases. The primary outcome of fetal growth restriction has been reported elsewhere [START_REF] Boylan | Maternal caffeine intake during pregnancy, risk of fetal growth restriction: a large prospective observational study[END_REF]. This paper presents our secondary outcome measure of late miscarriage (spontaneous pregnancy loss between 12 and 24 weeks) and stillbirth (delivery C24 weeks with no signs of life at birth).

A validated caffeine assessment tool completed during initial hospital registration at 8-12 weeks recorded habitual caffeine intake, divided into the 4 weeks before pregnancy, weeks 1-4, and weeks 5-12, repeated again for weeks 13-28, and again for weeks 28-36 weeks or end of pregnancy [START_REF] Boylan | Assessing caffeine exposure in pregnant women[END_REF]. The questionnaire covered all potential dietary sources of caffeine, both food and drink, and over the counter medications. Specific brand names, portion sizes, methods of preparation, quantity and frequency of intake were collected for different gestational periods. Precise caffeine intakes were then estimated based on the published caffeine content for each item, [START_REF]Survey of caffeine and other methylxanthines in energy drinks and other caffeinecontaining products (updated)[END_REF] and information provided by manufacturers and coffee houses. Alcohol intake was self-reported using a tool developed for the UK Women's Cohort study [START_REF] Cade | The UK Women's Cohort Study: comparison of vegetarians, fish-eaters and meat-eaters[END_REF]. Self-reported smoking, nausea and vomiting were also collected.

Women provided saliva samples at recruitment for determining nicotine exposure, collected using a Salivette Ò (Sarstedt, Aktiengesellschaft & Co., Loughborough, UK) kept in the mouth for 5-10 min. Saliva was isolated from the salivettes by centrifugation then stored at -80°C. All samples were analysed at the University of Leeds. A commercial ELISA kit (Cozart Bioscience Ltd, Oxfordshire, UK) was used to measure salivary cotinine concentrations, according to the manufacturer's instructions.

Statistical methods

Derived caffeine intake was expressed in mg/day averaged over the first trimester (defined as the first 12 weeks of pregnancy) and divided into the 4 weeks before pregnancy, weeks 1-4, and weeks 5-12 of the first trimester. Numbers were too small to investigate associations with intake in the second and third trimesters.

Geometric mean caffeine intakes were compared between those pregnancies resulting in late miscarriage or stillbirth and those resulting in a live birth, based on log-transformed caffeine intake in the first trimester.

Unconditional logistic regression modelling was used to predict late miscarriage or stillbirth, with stratification for the two centres, using Stata version 10 survey facilities, [START_REF] Statacorp | Stata statistical software[END_REF] adjusting for mother's age, previous history of miscarriage, salivary cotinine concentrations (as a continuous variable measured in ng/ml) and self-reported alcohol consumption, converted into units/day (one unit of alcohol is 10 ml by volume, or 8 g by weight, of pure alcohol) and categorised as 0, \0.25, 0.25-0.5, 0.5-1, and over 1 unit/day.

Tests for linear trend over the continuous measure of caffeine exposure were conducted by inclusion of the uncategorised continuous measure in the models. The robustness of the results to adjustment for nausea was assessed by adding this term to the model. The robustness of the results to adjustment for Index of Multiple Deprivation was also assessed. In addition to adjusting for previous history of miscarriage, the robustness of results to excluding women with previous deliveries was assessed. As a further exploration of possible residual confounding by smoking, the models were repeated using self-reported smoking status and amount smoked added to the model. The models were also repeated using caffeine from each of the two main sources, tea and coffee.

Results

A total of 13,071 eligible women were invited to participate and 2,643 (20%) consented. Women who had terminations were no longer at risk of miscarriage or stillbirth, and were therefore excluded (n = 8). Of the remaining 2,635 pregnancies, 19 resulted in late miscarriage and 9 in stillbirth. The mean (SD) maternal age was 30.0 (6.6) years. The mean pre-pregnant body mass index (SD) was 24.5 (4.5) kg/m 2 . 1,228 (47%) of women were primiparous, and 1,324 (51%) gave birth to boys. In the cohort, 106 (4%) women went into pre-term labour (\37 weeks gestation) and 67 (3%) had gestational hypertension (systolic blood pressure over 140 mmHg, diastolic blood pressure over 90 mmHg, on at least two occasions 30 min apart) or pre-eclampsia (systolic blood pressure over 140 mmHg, diastolic blood pressure over 90 mmHg, with proteinuria). The demographic and clinical characteristics are shown in Table 1 according to outcome.

The median (IQR) caffeine intake was 217 (104-366) mg/day, 183 (74-321) mg/day, and 92 (22-217) mg/day in the 4 weeks before pregnancy, weeks 1-4, and weeks 5-12, respectively. The median (IQR) caffeine intake over the whole first trimester was 132 (58-241) mg/day. The median (IQR) caffeine intake over the whole pregnancy was 134 (63-232) mg/day. In our study, tea contributed more than half of all caffeine consumption during pregnancy [START_REF] Boylan | Maternal caffeine intake during pregnancy, risk of fetal growth restriction: a large prospective observational study[END_REF]. The median (IQR) alcohol intake during the first trimester was 0.2 (0-0.6) units/day, with the highest consumption occurring, as might be expected, pre-pregnancy, and dropping rapidly in the first few weeks.

The geometric mean caffeine intake during the first trimester for women whose pregnancies resulted in a live birth was 103 (95% CI: 98-108) mg/day. The geometric mean caffeine intake during the first trimester for women whose pregnancies resulted in late miscarriage or stillbirth was 145 (85-249) mg/day.

There was a strong association between caffeine intake in the first trimester and subsequent late miscarriage between 12 and 24 weeks or stillbirth after 24 weeks adjusting for maternal age, parity, alcohol intake and amount smoked measured by salivary cotinine (Table 2). Compared to those consuming\100 mg/day, the odds ratio for late miscarriage or stillbirth increased to 2.2 (95% CI: 0.7-7.1) for intakes between 100 and 199 mg/day, to 1.7 (0.4-7.1) for those taking between 200 and 299 mg/day and to 5.1 (1.6-16.4) for those consuming over 300 mg/day (P trend = 0.004). There were insufficient events to present these separately for late miscarriage and stillbirth. This relationship appeared strongest in relation to caffeine consumption in the first few weeks of pregnancy, though confidence intervals are wide.

The results were unchanged by further adjustment for nausea: Compared to those consuming \100 mg/day of caffeine during the first trimester, the odds ratio for late miscarriage or stillbirth increased to 2.2 (95% CI: 0.7-7.1) for intakes between 100 and 199 mg/day, to 1.7 (0.4-7.1) for those taking between 200 and 299 mg/day and to 5.2 (1.6-16.6) for those consuming over 300 mg/day (P trend = 0.004). When adjusting for Index of Multiple Deprivation, again the results were unchanged: Compared to those consuming \100 mg/day of caffeine during the first trimester, the odds ratio for late miscarriage or stillbirth increased to 2.2 (95% CI: 0.7-7.0) for intakes between 100 and 199 mg/day, to 1.2 (0.2-6.0) for those taking between 200 and 299 mg/day and to 5.3 (1.6-to 17.1) for those consuming over 300 mg/day (P trend = 0.005). Conclusions were also unchanged when women with previous deliveries were excluded, though confidence intervals were wider than before because of the reduced numbers of cases (P trend = 0.001). When the model was repeated using selfreported smoking status and amount smoked added to the model, estimates were again very similar: Compared to those consuming \100 mg/day of caffeine during the first trimester, the odds ratio for late miscarriage or stillbirth increased to 2.3 (95% CI: 0.7-7.4) for intakes between 100 and 199 mg/day, to 1.7 (0.4-7.2) for those taking between 200 and 299 mg/day and to 5.3 (1.6-17.4) for those consuming over 300 mg/day (P trend = 0.003). Tea and coffee were the dominant sources of caffeine during pregnancy. Results for these sources are presented in Table 3 with caffeine intake categories combined to compare \200 mg/day with 200? mg/day. Wide confidence intervals reflect the small numbers of cases in the higher intake categories for these sources. The number of women with [200 mg/day from other sources of caffeine was too low to investigate those sources.

Discussion

Though the outcomes are rare, we have shown that there is a potentially strong association between caffeine and late miscarriage or stillbirth. Our results of increased late miscarriage and stillbirth with greater caffeine intake add further support to claims of an association between caffeine and negative birth outcomes [START_REF] Wen | The associations of maternal caffeine consumption and nausea with spontaneous abortion[END_REF][START_REF] Tolstrup | Does caffeine and alcohol intake before pregnancy predict the occurrence of spontaneous abortion?[END_REF][START_REF] Khoury | Consequences of smoking and caffeine consumption during pregnancy in women with type 1 diabetes[END_REF][START_REF] Weng | Maternal caffeine consumption during pregnancy and the risk of miscarriage: a prospective cohort study[END_REF][START_REF] Mills | Moderate caffeine use and the risk of spontaneous abortion and intrauterine growth retardation[END_REF][START_REF] Savitz | Caffeine and Miscarriage Risk[END_REF]. Many previous studies suffered from the potential for measurement error in the exposure. Our methods minimised the potential for exposure misclassification by including all sources of caffeine, including all caffeine-containing foods, beverages and medications, in a detailed validated questionnaire. Similarly, misclassification of important confounders can lead to bias through residual confounding. In our study, alcohol intake was recorded through the same detailed tool used for caffeine, and the important covariate of smoking status was recorded objectively using salivary cotinine as a biomarker of exposure. Many previous studies, particularly case-control studies, have collected exposure information retrospectively, with the potential for recall bias. However, in our study the exposure was assessed prospectively.

Whilst our study identified only small numbers of late miscarriages and stillbirths, the proportion of pregnancies resulting in these outcomes (1.1%) is similar to that found in a large survey of late miscarriage in Canada (0.6%) [START_REF] Wyatt | Age-specific risk of fetal loss observed in a second trimester serum screening population[END_REF]. However, the small number of events limited our power to * Adjusted odds ratios take account of maternal age, parity, amount smoked (cotinine concentration as a continuous variable, ng/ml) and alcohol intake (categorised as 0, \0.25, 0.25-0.5, 0.5-1 and 1? units/day). Where numbers do not total to 2,635, this is due to a small proportion of missing data. A total of 271 women were excluded from adjusted analyses with missing covariate information investigate subgroups, and resulted in very wide confidence intervals for the estimates. Our findings therefore need further replication in larger prospective population-based studies. Our response rate (20%) was disappointing and reflects the demanding requirements of the study. For ethical reasons responders could not be compared to nonresponders. However, mean birth weights for gestational age, rates of late miscarriage and stillbirth, and caffeine intakes were comparable to those found in other studies [START_REF] Wyatt | Age-specific risk of fetal loss observed in a second trimester serum screening population[END_REF][START_REF] Gardosi | Customized fetal growth standards: rationale and clinical application[END_REF][START_REF] Rosenberg | Selected birth defects in relation to caffeine-containing beverages[END_REF]. In addition, there is no reason to assume that caffeine metabolism phenotype should differ between responders and non-responders, other than in those confounders that we have adjusted for, e.g. smoking status.

Whilst it has been suggested that the relationship between caffeine and miscarriage or stillbirth is confounded by nausea, [START_REF] Wen | The associations of maternal caffeine consumption and nausea with spontaneous abortion[END_REF][START_REF] Fenster | Caffeine consumption during pregnancy and spontaneous abortion[END_REF] which may lead to both a reduction in caffeine intake and risk to the foetus, we consider that nausea is more likely to be either a marker for a healthy pregnancy, rather than causative, or that caffeine intake may lead to nausea in some. [START_REF] Wen | The associations of maternal caffeine consumption and nausea with spontaneous abortion[END_REF][START_REF] Fenster | Caffeine consumption during pregnancy and spontaneous abortion[END_REF][START_REF] Stein | Miscarriage, caffeine, and the epiphenomena of pregnancy: the causal model[END_REF] Wherever nausea lies on the causal pathway, analyses both with and without adjustment for nausea gave very similar results that did not alter the conclusions. In addition, we found an association with caffeine intake before pregnancy, which could not be influenced by physiological changes and processes during pregnancy (such as slowing caffeine metabolism and increasing the half life of serum caffeine).

It may be that higher caffeine intake increases risk of late miscarriage in susceptible women. A case-control study of women with recurrent pregnancy loss found a statistically significantly increased risk of pregnancy loss only in women with the homozygous (A/A) genotype representing the high inducibility genotype for CYP1A2, which increases the rate of caffeine metabolism [START_REF] Sata | Caffeine intake, CYP1A2 polymorphism and the risk of recurrent pregnancy loss[END_REF]. Studies in monkeys have shown a high rate of stillbirths and miscarriage with maternal caffeine intakes of 10-15 mg/kg body weight per day given via the drinking water, though experimental research such as this tends to provide evidence at doses of exposure that are not relevant to most human populations. For example, this is equivalent to a 60 kg woman consuming 8-12 average cups of coffee per day (assuming 75 mg caffeine/cup). Interestingly, the main serum metabolite of caffeine in monkeys is theophylline, whereas in humans it is paraxanthine. The comparative toxicities of these two metabolites is unknown [START_REF]Statement on the reproductive effects of caffeine[END_REF]. Klebanoff et al. using paraxanthine as a marker for caffeine intake, found an increased risk of late miscarriage only for very high intakes, [START_REF] Klebanoff | Maternal serum paraxanthine, a caffeine metabolite, and the risk of spontaneous abortion[END_REF] though it is unclear whether it is caffeine, paraxanthine, or another caffeine metabolite that is the potentially causative agent.

Conclusion

Our study identified only small numbers of late miscarriages and stillbirths, limiting power to detect small associations, and resulting in very wide confidence intervals. Therefore, although reaching statistical significance, there is considerable uncertainty in the size of the association. However, given the careful methodology employed, and the advances over some earlier studies, our findings make a valuable contribution to the literature in this field. Given also the paucity of randomised controlled trials in this area, and the practical and ethical difficulties in conducting them, our study strengthens the observational evidence base on which current guidance is founded.

Table 1

 1 Demographic and clinical characteristics of 2,635 subjects, according to outcome

	Characteristic	Late miscarriage	Live birth	All
		or stillbirth (n = 28)	(n = 2,607)	(n = 2,635)
	Mother's mean age (years) (SD)	30.2 (5.0)	30.0 (6.6)	30.0 (6.6)
	Mother's mean pre-pregnant weight (kg) (SD)	66.6 (13.4)	66.8 (13.1)	66.8 (13.1)
	Mother's pre-pregnant BMI (kg/m 2 ) (SD)	24.9 (4.3)	24.5 (4.5)	24.5 (4.5)
	History of previous miscarriage (%)	4 (14%)	611 (23%)	615 (23%)
	Gestational hypertension/pre-eclampsia (%)	0 (0%)	67 (3%)	67 (3%)
	Mean alcohol intake in 1st trimester (units/day) (SD)	0.5 (0.7)	0.5 (0.8)	0.5 (0.8)
	Salivary cotinine concentration (ng/ml) (%)
	\1 ng/ml	19 (70%)	1,816 (73%)	1,835 (73%)
	1-5 ng/ml	5 (19%)	302 (12%)	307 (12%)
	[5 ng/ml	3 (11%)	364 (15%)	367 (15%)
	Child's gender (male) (%)	6 (32%)*	1,318 (52%)	1,324 (51%)
	Mean birth weight (g) (SD)	1,880 (1,300)	3,460 (540)	3,450 (550)

* Child's gender not recorded for 9 late miscarriages Caffeine, late miscarriage and stillbirth

Table 2

 2 Odds ratios (95% CI) of late miscarriage and stillbirth by caffeine intake categories

		Caffeine	No. of	Unadjusted	(95% CI)	P trend	Adjusted	(95% CI)	P trend
		(mg/day)	cases/	OR			OR*
			total
	Average over 1st trimester	\100	6/998	1	-	0.04	1	-	0.004
		100-199	7/656	1.8	(0.6, 5.3)		2.2	(0.7, 7.1)
		200-299	3/402	1.2	(0.3, 5.0)		1.7	(0.4, 7.1)
		300?	9/426	3.6	(1.3, 10.1)		5.1	(1.6, 16.4)
	4 weeks before pregnancy	\100	3/604	1	-	0.002	1	-	\0.001
		100-199	5/570	1.8	(0.4, 7.5)		1.4	(0.3, 6.3)
		200-299	5/460	2.2	(0.5, 9.3)		2.2	(0.5, 9.4)
		300?	12/870	2.8	(0.8, 10.0)		3.0	(0.8, 10.9)
	Weeks 1-4	\100	3/781	1	-	0.001	1	-	\0.001
		100-199	5/572	2.3	(0.5, 9.6)		1.8	(0.4, 8.2)
		200-299	6/441	3.6	(0.9, 14.4)		3.8	(0.9, 16.7)
		300?	11/706	4.1	(1.1, 14.8)		4.7	(1.2, 18.7)
	Weeks 5-12	\100	12/1,302	1	-	0.6	1	-	0.2
		100-199	3/497	0.7	(0.2, 2.3)		0.8	(0.2, 3.0)
		200-299	6/325	2.0	(0.8, 5.4)		2.5	(0.9, 7.0)
		300?	4/373	1.2	(0.4, 3.6)		1.6	(0.5, 5.5)

Acknowledgments This research was funded by the Food Standards Agency (Contract T01032/33). We thank Stuart Creton, Clifton Gay, Gordon Gibson, David Gott, Fred Kadlubar, Mark Klebanoff Caroline Tahourdin, and Natalie Thatcher for their useful comments during the study., Heather Ong, Vilas Misty, Clare Lawrence, Chetan Ghandi, Bhavin Daudia, Sue Chell, and the Department of Chemical Pathology, University Hospitals of Leicester NHS Trust, for sample handling and processing.

Appendix

Contribution to authorship

Darren C. Greenwood conceived the paper, performed and interpreted statistical analyses, and wrote the manuscript. Principal Investigators were Janet E. Cade (Leeds) and Marcus S. Cooke (Leicester). Co-Investigators were Nigel Simpson, James Walker, Christopher P. Wild (Leeds) and * Adjusted odds ratios take account of maternal age, parity, amount smoked (cotinine concentration as a continuous variable, ng/ml) and alcohol intake (categorised as 0, \0.25, 0.25-0.5, 0.5-1 and 1? units/day). Where numbers do not total to 2,635, this is due to a small proportion of missing data. A total of 271 women were excluded from adjusted analyses with missing covariate information Justin C. Konje (Leicester). All authors were given the opportunity to comment on and approve the paper.

Members of the CARE Study Group are:

Clinicians: Nisreen Alwan, Nigel Simpson (Co-Investigator), James Walker (Co-Investigator) (Leeds); Justin C. Konje (Co-Investigator), Neelam Potdar (Leicester).

Nutritionists: Sinead Boylan (PhD student), Janet E. Cade (Principal Investigator), Sara F.L. Kirk (Co-Investigator) (Leeds).

Laboratory methods: Alastair W.M. Hay, Susan Shires, Kay L.M. White, Christopher P. Wild (Co-Investigator) (Leeds); Marcus S. Cooke (Principal Investigator) (Leicester).

Statisticians: Darren C. Greenwood (Co-Investigator)(Leeds); Nicholas Taub (Leicester).

Research midwives: Vivien A. Dolby (Leeds), Karen C. Chipps, Shabira Kassam (Leicester).

Database managers: James D. Thomas (Leeds); Jim Charvill (Leicester).


Alan D Fleming

Keith A Goatman

Sam Philip

Gordon J Prescott

Peter F Sharp

Dr John A Olson
email: john.olson@nhs.net.

MD Frcpedin Frcophth

Automated grading for diabetic retinopathy: a large-scale audit using arbitration by clinical experts

Keywords: diabetic retinopathy, screening, computer-assisted image analysis, imaging, telemedicine

come

INTRODUCTION

Systematic screening for diabetic retinopathy using retinal photography has been shown to reduce the incidence of blindness among people with diabetes. [START_REF] Hutchinson | Effectiveness of screening and monitoring tests for diabetic retinopathy -a systematic review[END_REF][START_REF] Arun | Long-term impact of retinal screening on significant diabetes-related visual impairment in the working age population[END_REF][START_REF] Stefansson | Screening and prevention of diabetic blindness[END_REF][START_REF] Zoega | Screening compliance and visual outcome in diabetes[END_REF] Diabetic retinopathy screening programmes are challenged by the rising prevalence of diabetes, [START_REF] Wild | Global prevalence of diabetes: estimates for the year 2000 and projections for 2030[END_REF] the costs of implementation, [START_REF] James | Cost effectiveness analysis of screening for sight threatening diabetic eye disease[END_REF][START_REF] Scotland | Cost-effectiveness of implementing automated grading within the national screening programme for diabetic retinopathy in Scotland[END_REF] the maintenance of an effective quality assurance system, [START_REF] Garvican | A pilot quality assurance scheme for diabetic retinopathy risk reduction programmes[END_REF] and the repetitious nature of grading. As a result, means for improving the efficiency of screening are being sought such as optimisation of the screening interval, [START_REF] Olafsdottir | Biennial eye screening in patients with diabetes without retinopathy: 10-year experience[END_REF] reducing the number of photographic fields used, [START_REF] Williams | Single-field fundus photography for diabetic retinopathy screening -A report by the American Academy of Ophthalmology[END_REF] and automation of image grading, [START_REF] Sharp | The value of digital imaging in diabetic retinopathy[END_REF] the topic of this paper.

The performance of automated grading software is often measured in terms of sensitivity and specificity for detecting retinopathy. [START_REF] Larsen | Automated detection of fundus photographic red lesions in diabetic retinopathy[END_REF][START_REF] Usher | Automated detection of diabetic retinopathy in digital retinal images: a tool for diabetic retinopathy screening[END_REF] However, absolute values for these may not be immediately useful when deciding whether an automated grading system would be effective at improving screening programme efficiency. It is also important to know the relative sensitivities and costs between alternative grading systems, one of which is performed only by human graders while the other includes automated grading software. It is also important to know whether any missed cases have equivocal or unequivocal retinopathy.

In a previous study, involving [START_REF] James | Cost effectiveness analysis of screening for sight threatening diabetic eye disease[END_REF],722 patients from Grampian, we compared automated and manual grading systems to a reference standard grading. [START_REF] Scotland | Cost-effectiveness of implementing automated grading within the national screening programme for diabetic retinopathy in Scotland[END_REF][START_REF] Philip | The efficacy of automated "disease/no disease" grading for diabetic retinopathy in a systematic screening programme[END_REF] The automated grading software had a better detection rate, 90.5% (2283/2523), than manual grading, 86.5% (2182/2523), for any retinopathy (mild or more severe retinopathy) and there was no significant difference between the detection rates, 97.9% (323/330) by automated grading and 99.1% (327/330) by manual grading, for observable/referable retinopathy (more severe than mild retinopathy). With 45.7% of cases not requiring manual grading, an annual saving of £200,000 would be made for the 160,000 people with diabetes in Scotland, estimated for the financial year 2005/2006. However, before the software could be implemented into the Scottish national diabetic retinopathy screening programme, it was necessary to demonstrate that its performance was maintained when used in other Scottish screening centres.

Table 1:

The Scottish grading scheme. [START_REF]Scottish Diabetic Retinopathy Grading Scheme[END_REF] Inferior and superior hemi-fields are delineated by a line passing through the centres of the fovea and the optic disc. Designations in the Scottish grading scheme are given (R0…R6, M0…M2).

Grade

Features Location Outcome

Retinopathy

No visible retinopathy (R0,M0)

MATERIALS AND METHODS

The study outline is shown in figure 1.

Images were obtained from two Scottish screening centres, Glasgow and Fife, working in accordance with the requirements of National Health Quality Improvement Scotland. Caldicott Guardian approval was obtained for the study.

Photography and grading were performed by graders employed within the Scottish screening programmes according to the recommendations of Scotland's National Diabetic Retinopathy Screening Collaborative. [START_REF] Leese | A national retinal screening programme for diabetes in Scotland[END_REF] A single 45° macula-centred photograph was taken of each eye, with additional photographs obtained as required, for example, to obtain better image quality or a better view of pathology. Mydriasis was used where small pupils prevented the capture of adequate quality images; a process known as "staged mydriasis". Images were obtained with 7 fundus cameras in Glasgow (Canon EOS D30 and EOS 20D digital cameras with Canon CR6-45NM and CR-DGi non-mydriatic retinal cameras) and 3 in Fife (Canon EOS 20D digital cameras with Canon CR-DGi non-mydriatic retinal cameras). Image sizes were 2336x3504 (10,950 images), 1696x2544 (25,378 images) and 1440x2160 (42,273 images) pixels.

Manual grading was performed by the screening centres following the Scottish grading scheme illustrated in table 1. This table defines the lesions present at each grade of retinopathy, explains the terms observable and referable maculopathy/retinopathy and illustrates the associated patient outcomes. The outcome for the majority patients is 12month recall.

There are three more urgent possible outcomes; referral to ophthalmology, 6-month recall or referral for slit-lamp examination. 84,658 consecutive images and grading results from 34,785 anonymised patient screening episodes attending screening centres in Glasgow or Fife between 1st January 2007 and 31st January 2008. In the case of patients with repeat photographs, only the earliest screening episode was used. This set contained 78,601 images from 33,535 patients.

Automated grading software

All images were processed by the automated grading software which assessed image quality and disease using previously described algorithms [START_REF] Fleming | Automated assessment of retinal image quality based on clarity and field definition[END_REF][START_REF] Fleming | Automated microaneurysm detection using local contrast normalization and local vessel detection[END_REF], summarised here. The location of the optic disc and the fovea were determined to check that the image shows a macula-centred view and to determine whether the image is of a left or a right eye. Image clarity was assessed by checking that sufficient small vessels were visible in the macula region. If an adequate quality macula-centred image is present for each eye then disease is assessed in these images by performing microaneurysm/dot haemorrhage detection. The first stage in the detection of these was performed on the green plane of the image, after correction for uneven illumination, using a non-specific morphological filter that determines candidate lesions by separating dot-like dark objects from the linear vasculature . [START_REF] Fleming | Automated microaneurysm detection using local contrast normalization and local vessel detection[END_REF] The second stage performs more detailed analysis of the candidate lesions, measuring such features as their area, contrast and likelihood of lying on a vessel. An automated classifier, which had been previously trained using a set of 35 images containing 198 individually annotated microaneurysms/dot haemorrhages, separates true lesions from background objects based on the values of these features.

Artefacts caused by dirt on camera internal surfaces or by faulty photosensors can have a very similar appearance to microaneurysms/dot haemorrhages and hence cause frequent false positive detections. Therefore, an automated artefact masking algorithm was developed that noted the location of potential microaneurysms/dot haemorrhages in images taken with the same camera. Repeated detection of a potential microaneurysm/dot haemorrhage at the same location and in the same camera was judged to indicate the presence of an artefact.

The automated grading software produces a binary output: a patient is software positive if the software assessed quality to be inadequate or if it detected microaneurysms/dot haemorrhages. The automated grading software performance was evaluated by comparing this binary output against the screening programme manual grading results.

Arbitration

The most important discrepancies between the binary output of the automated grading software and the screening programme were those involving patients who were not detected by the automated grading software but who the screening programme referred to ophthalmology, recalled at 6 months or referred for slit-lamp examination. Arbitration grading was performed on these discrepancies.

The arbitration grading was performed by seven senior ophthalmologists working at screening centres in Scotland. The discrepancies were mixed with 40 control images and presented to each grader in random order. As the control images had been graded by the arbitrators at least 6 months earlier, during a quality assurance test, an intra-grader comparison was made.

If 5 or more arbitrators allocated a grade requiring referral to ophthalmology, recall at 6 months or slit-lamp examination, then this was defined as consensus for an outcome more urgent than the standard 12-month recall. In these cases, a consensus grade was allocated. This was necessary in order to evaluate the automated grading software performance for each grade. The grade assigned by the greatest number of arbitrators was used as the consensus grade. Where there was a choice between possible grades, due to more than one grade being associated with the same maximum number of arbitrators, the most severe of these grades was chosen as the consensus grade.

If 3 or 4 arbitrators allocated a grade which would imply that either referral to ophthalmology, recall at 6 months or slit-lamp examination was required, then this was defined as no consensus. The patient was deemed to be indeterminate regarding whether or not the outcome was more urgent than the standard 12-month recall. In this case the consensus grade was indeterminate.

If 2 or fewer arbitrators allocated a grade which would imply that referral to ophthalmology, recall at 6 months or slit-lamp examination was required, then this was defined as consensus that the patient required 12-month recall. The consensus grade was mild retinopathy or no retinopathy.

Statistical Methods

The intergrader agreement between the assigned grades of the 7 senior ophthalmologists were summarised using a Kappa statistic. A Kappa statistic was also used to assess intragrader agreement, comparing the earlier and later gradings of the 40 control images. The detection rates within different ethnic groups were compared using Pearson chi-squared tests or Fisher's exact tests if required due to small numbers not being detected. This was done separately for those patients recalled at 12 months and those having a more urgent outcome. Stata version 10, StataCorp, Texas was used for these analyses.

RESULTS

The diabetic retinopathy screening centres referred 5.0% of patients to ophthalmology, recalled 1.8% at 6 months and referred 5.5% for slit-lamp examination.

Arbitration grading was performed on the images of 127 patients referred to ophthalmology, recalled at 6-months or referred for slit-lamp examination but who were not detected by the automated grading software. The Kappa statistic for inter-grader agreement was κ=0.37. Mild retinopathy [START_REF] Fleming | Automated assessment of retinal image quality based on clarity and field definition[END_REF] No retinopathy (46)

Table 2 summarises the consensus grades resulting from the arbitration grading and shows the number of arbitrators who considered referral to ophthalmology, 6-month recall or slitlamp examination was required for each case. In 63 cases the output of the automated grading software was appropriate and in 37 cases (31 with referable maculopathy, 3 with observable maculopathy and 3 ungradable) it was inappropriate according to the consensus of the arbitrators. 27 cases were indeterminate due to lack of consensus.

Considering the assigned grades, all 7 arbitrators gave identical grades in only 20 cases (5 cases of no retinopathy, 1 case of mild background retinopathy, 1 case of observable maculopathy, 12 cases of referable maculopathy, 1 case ungradable).

For the 40 control images, there was good intra-grader agreement with κ=0.77. 83% of the grades allocated to these images were identical to those allocated during a quality assurance test over 6 months earlier, 10.7% were higher than the previous ones and 6.3% were lower.

DISCUSSION

This audit assessed the performance of an automated system based on microaneurysm/dot haemorrhage detection and image quality assessment operating on a large, unselected population of people with diabetes participating in Scotland's systematic diabetic retinopathy screening programme. Out of 33,535 patients, all 697 patients with observable or referable retinopathy, other than maculopathy, were detected. In addition, the detection rates for observable and referable maculopathy and technical failures are compatible with published standards and quality assurance protocols and are very high compared to the sensitivity of retinal photography. [START_REF] Diabetic | Retinal Photography Screening for Diabetic Eye Disease[END_REF][START_REF]Diabetic Retinopathy Screening[END_REF][START_REF] Williams | Single-field fundus photography for diabetic retinopathy screening: a report by the American Academy of Ophthalmology[END_REF] They are also higher than previously reported manual grading rates for detection of referable retinopathy which were 96% by Olson et al [START_REF] Olson | A comparative evaluation of digital imaging, retinal photography and optometrist examination in screening for diabetic retinopathy[END_REF] and a maximum of 85% by Abramoff et al [START_REF] Abràmoff | Evaluation of a System for Automatic Detection of Diabetic Retinopathy From Color Fundus Photographs in a Large Population of Patients With Diabetes[END_REF].

The scale of the audit means that there are sufficient arbitration results to warrant detailed consideration. It has been shown that for cases where the automated grading software and the diabetic retinopathy screening programme results are conflicting, the grading software result was correct more frequently than incorrect; only 29% of the 127 arbitrated discrepancies had 5 or more arbitrators (defined as consensus) assigning an outcome requiring more urgent action than the standard 12-month recall. There was much disagreement between the senior ophthalmologists; full agreement occurred in only 16% of arbitrated discrepancies. The good concordance in the grading of the control images suggest that there was no tendency by the arbitrators to give the images a less or more severe grade than they would have done outside this evaluation.

The automated system performed equally well for images from the three ethnic categories tested; Caucasian, Asian and Afro-Caribbean.

While the design of our earlier study made possible a direct comparison between manual and automated grading, this audit did not look for cases of referable retinopathy/maculopathy that were missed by manual grading. It is therefore not known if there were patients who were disadvantaged by manual grading.

All cases that required referral to an ophthalmologist but were not detected by the automated grading software were of referable maculopathy. This may be explained by these cases having clear exudates but no obvious microaneurysms. However, in a study, also involving people attending the Scottish diabetic retinopathy screening programme, only 13.2% of patients with signs of referable maculopathy received laser treatment. [START_REF] Schofield | Macular oedema is not predicted by perifoveal single blot haemorrhages[END_REF] This implies that only a few of the cases of referable maculopathy not detected by the automated grading software would be considered for laser treatment and some of these may be detected on their next screening attendance.

It has been confirmed that the automated system achieves similar performance to that found in our previous study, [START_REF] Philip | The efficacy of automated "disease/no disease" grading for diabetic retinopathy in a systematic screening programme[END_REF] when applied to a larger multi-centre dataset. The current audit and our previous study achieved, respectively, detection rates of 87.0% and 87.9% for any retinopathy, 98.5% and 97.9% for observable/referable retinopathy/maculopathy and 99.8% and 99.5% for technical failures. While the earlier study indicated that 45.7% (3,070/6,722) of patients would be removed from the manual grading workload, in this audit 36.3% (12,185/33,535) would have been removed. This is may be due to the presence of artefacts, which affected 8 out of the 10 cameras used in the audit.

In this audit the automated grading software achieved 100% detection of proliferative, referable background and observable background retinopathy. It has been shown that the performance of an automated screening system for diabetic retinopathy is maintained at safe and effective levels over a large unselected set of images obtained from regional grading centres. Following this audit, Scotland's National Diabetic Retinopathy Screening Collaborative has concluded that automated grading should be introduced into Scotland's diabetic retinopathy screening programme.

Figure 1 :

 1 Figure 1: Diabetic retinopathy screening programme and study procedures.

Table 2 :

 2 Results of the arbitration grading. The second column groups the cases according to the definition used for consensus.

	Number

of arbitrators, out of 7, who assigned a grade implying referral to ophthalmology, 6-month recall or slit-lamp examination (number of cases) Consensus grade (number of cases)

	7 arbitrators (19)	Referable maculopathy (31)
	6 arbitrators (8)	Observable maculopathy (3)
	5 arbitrators (10)	Ungradable (3)
	4 arbitrators (11) 3 arbitrators (16)	Indeterminate (27)
	2 arbitrators (13)
	1 arbitrator (22)
	0 arbitrators (28)

Table 3

 3

	displays the automated system results for each grade of retinopathy and for each
	possible patient outcome. The software was positive in 100% (193/193) of cases of
	proliferative retinopathy, 100% (324/324) of cases of referable background retinopathy and
	100% (180/180) of cases of observable background retinopathy. Other software positive
	rates were, 97.3% (1099/1130) for referable maculopathy, 99.2% (384/387) for observable
	maculopathy and 99.8% (1824/1827) for patients classified as ungradable. The error rate
	for cases requiring referral to ophthalmology, 6-month recall or slit-lamp examination was
	0.11% (37/33535).

Table 3 .

 3 Proportions of patients, with confidence intervals (CI), for which the automated grading software output was positive. 27 cases have been omitted since they had indeterminate grade.

	Grade	Software positive % 95% CI
	No visible retinopathy	49.6 (10668/21503)	48.9-50.3
	Mild background retinopathy	83.9 (6678/7964)	83.0-84.6
	Observable maculopathy	99.2 (384/387)	97.8-99.7
	Observable background retinopathy 100 (180/180)	97.9-100
	Referable maculopathy	97.3 (1099/1130)	96.1-98.1
	Referable background retinopathy	100 (324/324)	98.8-100
	Proliferative referable retinopathy	100 (193/193)	98.1-100
	Ungradable	99.8 (1824/1827)	99.5-99.9
	Outcome
	12-month recall	58.9 (17346/29467)	58.3-59.4
	6-month recall	99.5 (564/567)	98.5-99.8
	Slit-lamp examination	98.1 (1616/1647)	97.3-98.7
	Refer to ophthalmology	99.8 (1824/1827)	99.5-99.9
	51.7% (17346/33535) of all patients required 12-month recall but were software positive.
	Out of the 33,535 patients screened, 12,185 would not require manual grading since they
	were assessed by the automated grading software as having adequate quality and no
	microaneurysms. Automated grading would therefore reduce the number of patients
	requiring manual grading by 36.3%.
	Ethnicity was recorded for 12,100 patients; 10,068 (83%) were Caucasian, 1,915 (16%)
	were Asian and 117 (1%) were Afro-Caribbean. Mean age of all patients was 62 years and
	54.6% of all patients were male.
	Table 4 displays the proportions which were software positive for the three ethnic categories
	for each patient outcome. No statistically significant differences were found between the
	ethnic groups in detection rates either for patients with non-referable retinopathy (Pearson
	chi-squared test p=0.38) or for patients requiring referral to ophthalmology, 6-month recall or
	slit-lamp examination (Fisher's exact test p=0.82).

Table 4 :

 4 Proportions of patients, with confidence intervals (CI), for which the automated grading software gave a positive result for each possible patient outcome, split by ethnicity.

		Caucasian		Asian		Afro-Caribbean
	Outcome	Software positive %	95% CI	Software positive %	95% CI	Software positive %	95% CI
	12-month recall	65.0 (5560/8555)	64.0-66.0	63.3 (989/1563)	60.9-65.6	66.3 (67/101)	56.7-74.8
	6-month recall 99.6 (257/258)	97.8-99.9	100 (27/27)	87.5-100	100 (1/1)	20.7-100
	Slit-lamp examination	100 (542/542)	99.3-100	99.3 (136/137)	96.0-99.9	100 (6/6)	61.0-100
	Refer to ophthal-mology	98.3 (695/707)	97.1-99.0	98.4 (183/186)	95.4-99.4	100 (9/9)	70.1-100

ACKNOWLEDGMENTS

We thank Dr Alison Bow, Dr Brian Power, Dr Cynthia Santiago, Dr Anne Sinclair, Dr William Wykes, Dr Sonia Zachariah and Dr Usha Zamvar for voluntarily performing the arbitration grading. We thank staff of NHS Greater Glasgow & Clyde and NHS Fife retinal screening programmes, and their directors Dr William Wykes and Dr Caroline Styles, for their support. We thank Dr Nigel McLean of Scottish Health Innovations Limited and Scotland's National Caldicott Guardian, Dr Adam Bryson, for their support.

COMPETING INTERESTS

Implementation in Scotland and elsewhere is being considered. If this occurs it is likely that there will be some remuneration for the University of Aberdeen, NHS Grampian, Alan Fleming, John Olson and Peter Sharp. Authors Keith Goatman, Sam Philip and Gordon Prescott have no financial conflict of interest other than by association with the institutions mentioned above.

FUNDING

Funding for Alan Fleming was provided by Medalytix Ltd under an agreement with Scottish Health Innovations Limited.

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